UCL DEPARTMENT OF POLITICAL SCIENCE
SCHOOL OF PUBLIC POLICY
School of Public Policy Working Paper Series: ISSN 1479-9472
Working Paper 17
Judicial Independence in Transition:
Revisiting the Determinants of Judicial Activism in the
Constitutional Courts of Post-Communist States
Jonathan C. Bond
J.D. Candidate, The George Washington University Law School (Washington, DC)
The support of UCL Friends Programme is gratefully acknowledged.
School of Public Policy
University College London
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Date: February 2006
Judicial Independence in Transition:
Revisiting the Determinants of Judicial Activism in the
Constitutional Courts of Post-Communist States
Jonathan C. Bond 1
J.D. Candidate, The George Washington University Law School (Washington, DC)
06 February 2006
ABSTRACT:
This study investigates the relationship between the dimensions of judicial independence
and judicial review in the jurisprudence constitutional courts of ten countries in Central
and Eastern Europe and the Former Soviet Union. The study is in part a modified
replication of prior works examining the issue, using newly collected data from a panel of
ten countries. It examines the relationship of judicial review with: (1) judicial
independence (using both measures employed by the prior works, corrected versions of
those measures, and measures original to this study); (2) political and social contextual
factors examined by the previous works; and (3) the receptiveness of post-communist
countries to the importation of transplanted legal institutions. Improvements on the
conceptualisation of judicial independence, inclusion of the dimensions of receptiveness,
and a more appropriate panel of countries and dataset, inter alia, enable this study to
present a more complete and accurate portrait of constitutional judicature in transition
contexts.
The results show that while corrections to prior measures of judicial independence
improve the results at the margin, the entirely new measures of the concept represent a
greater step forward. Several dimensions of judicial independence are positively related
to judicial review, as are the measures of countries’ receptiveness to legal transplants.
Other key factors positively related to judicial review in transition include legislative
fragmentation at the time of each court decision, the scope of rights guarantees in a bill of
rights, and popular trust in courts. Presidential power is negatively related to judicial
review. The findings further indicate that aside from judicial independence, the prior
works do present correct portrayals of most of the contextual influences they investigate.
1
The support of the US-UK Fulbright Programme is gratefully acknowledged.
1. INTRODUCTION
The collapse of communist regimes in the Former Soviet Union (FSU) and Central
and Eastern Europe (CEE) brought with it the dramatic transformation of entire
systems of constitutional justice across Europe’s eastern frontier (e.g. Schwartz 2000;
Procházka 2003; Přibáň et al. 2003). At the same time that research exploring the
roles of courts in shaping public policy at nearly every level of governance was
accelerating rapidly (Tate and Vallinder 1995; Stone Sweet 2000), many states where
the ‘rule of law’ had existed in name only for nearly half a century were freed
suddenly to design and attempt to build (or import) new institutional frameworks for
the administration of law and justice (Sadurski 2001, 8ff.). Of the structural features
designed and built in transitioning states, the new courts “were by far the most
important of the new institutions” (Dupré 2003, 26).
These new tribunals stand out for at least two reasons. First, while the process
and products of transition differed in many ways across the region, the presence of a
constitutional court was one of a few important commonalities. Though to 1989 only
two countries throughout CEE possessed some (nominally) independent constitutional
court, by the mid-1990s every state in the region had established a constitutional
adjudicatory body, the vast majority patterned after the German Federal
Constitutional Court (FCC) model (Sadurski 2001, 1; cf. Tanchev 2000; Dupré 2003;
Howard 2001; Schiemann 2001). 2 The result is a sort of natural experiment—the
same basic court model, transplanted into different contexts, yielding noticeably
different results—which has not gone unnoticed by students of courts as policyshapers (e.g. Epstein, Knight, and Shvetsova 2001; Boulanger 2003a; Scheppele
2003b; Sólyom 2003).
Second, the courts themselves often played a very active part in shaping the
process of democratic transition (Dupré 2003, 26), and many became formidable
policy players in a very brief span of time. The Hungarian Constitutional Court, for
instance, struck down nearly one-third of all primary legislation (273 of 905 national
laws) brought before it for constitutional review during its first six years alone
(Scheppele 2003b, 224). 3 The weight of the issues in which that Court played a
pivotal role is striking: retroactive criminal legislation, 4 restitution for expropriation
2
Indeed, establishing a constitutional court became the distinguishing “trade mark” and the “proof of
the democratic character” of post-communist states eager to demonstrate that a genuine transformation
to democracy was underway (Sólyom 2003, 134). Procházka (2003) insightfully emphasises the
importance attached in transitioning states – especially the Visegrad Four (Hungary, Czech Republic,
Slovakia, and Poland) – to being perceived in CEE and Western Europe as genuinely committed,
indeed at the forefront, of transformation from authoritarian socialism to democracy.
3
This is particularly significant given the scope and range of the Hungarian Constitutional Court’s
exercise of its review powers in its first decade of operation. As Sólyom and Brunner (2000, 81) note,
nearly all of Hungary’s legislative framework was reviewed by the constitutional court (largely as a
consequence of the ease of access for petitioning review by the court; specifically, via the actio
popularis, any individual could petition for constitutional review of any law, regardless of whether that
individual was in any way affected by the law itself; some 3,170 cases of this type were initiated
between 1990 and 1996, inclusive).
4
See Decisions 11/1992 and 53/1993; available (in English) in Sólyom and Brunner (2000). For a
fuller discussion of the retroactive punishment issue, see Nalepa (2002) and (2003).
by the communist regime, 5 lustration laws, 6 the IMF-directed economic austerity
programme, 7 abortion, 8 the death penalty, 9 same-sex partnerships, 10 and many others
(Sólyom and Brunner 2000). More recently, in July 2005, on the basis of a petition
by thirteen citizens, the Slovakian Constitutional Court nullified the legislature’s
ratification the EU Constitution and (temporarily) barred the country’s president from
signing it (Jurinová 2005; Forbes 2005).
Given both the natural experiment generated by their widespread creation and
the political significance of their activity, the new courts deserve the attention of those
who study constitutional adjudication in other contexts. (Epstein, Knight, and
Shvetsova 2001; Boulanger 2003a; Scheppele 2003b; Sólyom 2003). Importantly,
however, the development of the new courts is significant not only as an opportunity
to apply existing models of judicial independence and judicial activism, but also
because of how lessons from these unconventional contexts can improve and expand
existing theories of judicial politics. As Scheppele observes (2003b, 220), the bright
line drawn in the democratisation literature (e.g. Linz and Stepan 1996) between
‘consolidated’ democracies from states still struggling through (or waiting to begin)
the process of transformation to democracy may prove counterproductive in the study
of judicial institutions. Instead of treating the new constitutional courts of the FSU
and CEE as categorically separate, all but quarantined from conventional contexts, the
new courts
are good to study not because they are different in kind from what
becomes normal politics, but precisely because they reveal in sharper
relief the problems buried in what passes for normal in ‘consolidated’
democracies (2003b, 220).
In short, the experiences, contexts, and truncated histories of the new constitutional
courts may be pivotal in identifying gaps in existing theory: they may indeed bring to
light forces which are invisible in ‘conventional’ contexts, but which ought to be
incorporated into any general theory of constitutional adjudication.
Several noteworthy obstacles make the study of these courts especially
problematic, however. First, at the theoretical level, despite the myriad volumes of
research on judicial politics produced in the last two decades, a general theory of
judicial independence – even a common definition – is still lacking (Russell 2001;
Herron and Randazzo 2003). Second, the methodologies developed for studying
constitutional politics in stable democracies with long traditions of constitutional
review may be inadequate in states where the courts are brand new, have no recent
tradition or legacy to build upon, and where the rules of the game are constantly in
flux (Boulanger 2003a). For instance, rational choice theorists have been able to
5
Decisions 21/1990 and 16/1991 (Ibid.).
6
Decision 60/1994 (Ibid.).
7
Decision 43/1995 (Ibid.).
8
Decision 64/1991 (Ibid.).
9
Decision 23/1990 (Ibid.).
10
See Decision 14/1995 (Ibid.).
2
develop relatively few truly dynamic models (i.e. those which can incorporate
changes in the ‘rules of the game’ themselves) relevant to post-communist transition,
and are instead limited to repetitive, non-progressive games (e.g. Epstein, Knight and
Shvetsova 2001). 11 This problem, which is troublesome enough in studying
conventional, consolidated democracies (Stone Sweet 1998; Vanberg 1998b;
Boulanger 2003a), is all but fatal in understanding transition contexts in which the
existence, content, and effectiveness of formal and informal rules of law are never
known with certainty.
Finally, the transition context presents problems concerning the actual data
needed for study. Not only is there no standardised, transparent, and widely used
database on court activity available—as there is for the world’s most-studied
court 12 —but the lack of familiarity with the new courts makes it hard for scholars
even to know what data to collect from each of the cases courts decide. Add to these
problems the tremendous language barriers, 13 numerous inconsistencies between the
courts in what documents and statistics are publicly available, and the fact that most
in-depth research continues to come from single-country studies where the authors
frequently are sitting or former jurists on the court they describe, and the challenges
facing researchers of the new courts become quite daunting.
Keeping these limitations in mind, this study aims to identify factors which
determine the level of judicial review 14 asserted by the new constitutional courts,
focusing on the relationship of judicial review with judicial independence in postcommunist transitioning states. Prior investigations of this question have already
attempted to identify the factors which determine or predict judicial activism in
transition contexts, explicitly seeking the relationship of judicial review and judicial
11
This particular model displays a number of crucial deficiencies, however, discussed in Part 2.
12
The “Original United States Supreme Court Judicial Database,” compiled by Prof. Harold Spaeth, is
“certainly the most important and influential” – as well as the most comprehensive and transprent–
dataset available for the study of the USA Supreme Court (Epstein, Knight, and Martin 2003, 807).
“Most empirical articles on judicial behavior, published over the last decade in the leading political
science journals, avail themselves of one of the Spaeth databases,” and those which did not either drew
from public survey data or examined courts other than the Supreme Court (Id., 812).
13
Schwartz, whose book on the new constitutional courts in CEE is perhaps the most frequently cited
major work on the topic, acknowledges: “I do not, however, know the languages of the many countries
covered in this book, I certainly am not very familiar with either their legal systems or national
cultures, and I have translations of only some of the decisions and actions I discuss” (2000, xix).
14
Judicial review is used as a proxy for activism by many prior studies of judicial behaviour, and
especially those examining constitutional courts in transition, and thus is used here. The two concepts
are not identical, however, as even those who employ this measure appreciate (Herron and Randazzo
2003). Scholars have developed an applied a wide variety of definitions of “judicial activism” (for
discussions of several definitions in comparison, see Kmiec 2004; see also Epstein, Knight, and Martin
2003; Edelman 2005; Calabresi 2004). For instance, judicial activism has been defined variously as:
any invalidation by a court of executive or legislative branch actions (e.g. Sunstein 2003), decisions
which in actual substance expand the scope of the judiciary’s authority (e.g. Kerr 2003), substantive
departures from prior court precedents (see Kmiec 2004, 1466ff.), rulings which do not comport with
the ‘original meaning’ of the secondary rule (e.g. constitution) by the authority of which they invalidate
a law or executive action (e.g. Barnett 2002). For purposes of this study, however, the first definition
(equating activism with invalidation) is used, despite conceptual inferiority, because existing studies of
constitutional justice in CEE and FSU have employed it (Herron and Randazzo 2003; Smithey and
Ishiyama 2002). Subsequent analysis of judicial independence and judicial activism should explore
alternative measures.
3
independence. However, several important factors (discussed below) give rise to
doubts about the reliability of the findings of these prior works. These doubts in turn
point to the need for re-examination to replicate, correct, and expand upon these
works’ methods and conclusions.
To this end, this study explores whether the degree of institutionally inscribed
formal independence of the courts themselves affect the degree of judicial review
observed in CEE and the FSU. Additionally, the study examines other factors
influencing the level of judicial review, some of which were studied in prior
investigations of this question, while others are original to this study. What is learned
about the determinants of judicial review by constitutional courts in CEE and the FSU
will contribute not only to an understanding of these courts, but can also help to both
test and reshape general theories of both constitutional adjudication and the
transplantation of legal institutions.
To this end, the study proceeds as follows: Part 2 reviews existing literature
addressing constitutional judicial politics in transition, focusing on two particular
investigations (Smithey and Ishiyama 2002; Herron and Randazzo 2003) which
studied the relationship between judicial independence and judicial review in CEE
and the FSU, and which form the basis for replication. Part 3 describes the data and
methodology employed in this study, and Part 4 analyses the results. Part 5 offers
conclusions regarding methodological improvements and substantive findings.
2. REVIEW AND CRITIQUE OF EXISTING LITERATURE
2.1. BACKGROUND
As Sadurski (2001, 8) astutely observes, the new constitutional courts of CEE and the
FSU remain the most “under-theorized” of any of the aspects of transition. While the
new courts have received attention from around the world and from a variety of
disciplines (Dupré 2003, 3-4), much of this attention has been descriptive and
exploratory in nature (e.g. Schwartz 2000; Přibáň, Roberts, and Young 2003; Krygier
and Csarnota 1999; Sadurski 2002; Harutyunyan and Mavčič 1999). While those
studying the new courts have brought with them established theories and models of
judicial independence developed in other contexts (Howard 2001; Burbank and
Friedman 2002), comparatively little new theoretical ground has been broken in this
area. 15 This is surprising not only because of the ubiquity, activist tendencies, and
importance of these courts to the transformation to democracy, but also because
constitutional judicature in transition lies at the nexus of two rich fields of the
legal/political science and transition literatures.
First, the study of constitutional courts in transition is benefited by the
established theoretical and empirical literature on constitutional judicial politics,
which has produced a multiplicity of methodologies and models for investigating
established constitutional courts (e.g. Shapiro 1981; Stone Sweet 2000; Vanberg
2005; Garrett, Kelemen, and Schulz 1998; Epstein and Knight 1998; Epstein, Knight,
and Martin 2003). 16 Intertwined with these advances have been efforts, incomplete as
15
There are exceptions, of course: the work of Procházka (e.g. 2003) and Scheppele (e.g. 2002, 2003b)
reflect efforts to approach the area with notable originality.
16
As Boulanger (2003, 7ff.) recounts, three major threads of have taken shape in the literature on
courts and their role in governance: formalist (i.e. rational choice) models, “behavioural/empiricist
accounts” employing several tools of statistical analysis, and the more abstract “triadic dispute
resolution” models such as that elaborated by Stone Sweet (2000). As scholars have begun to turn their
4
yet, to uncover the nature, causes, and effects of judicial independence (Russell and
O’Brien 2001; Ackner 1997; Burbank and Friedman 1997; Shetreet and Deschenes
1985; Ramseyer and Rasmussen 2003). Several applications of these theories and
methods to transition contexts are discussed in the following section.
Second, the literature on the “transplantation” (Watson 1993; 1995) or
“importation” (Ajani 1995) of legal systems, and the explicit extensions of this line of
investigation to post-communist contexts (Dupré 2003), have offered a number of
critical insights pertinent to constitutional jurisprudence in transition countries. Most
importantly, such explorations have identified several strategic and contextual factors
– including familiarity with and adaptation of the legal system transferred (Pistor,
Raiser, and Gelfer 2000; Berkowitz, Pistor and Richard 2003b) – which can influence
the successful transfer of legal norms and institutions from one state to another.
When such factors are absent, the result is exhibit a “transplant effect” – that is, they
will not function as effectively as the same institutions did in their original context
(Berkowitz, Pistor, and Richard 2003a).
These insights have been specifically
extended to constitutional law in CEE and the FSU with promising results (Pistor
2002; Hendley 1999).
2.2. SUMMARY OF PRIOR RESEARCH: AIMS & FINDINGS
2.2.1. Rational Choice Models
Rational choice models have become one of the most prevalent and provocative tools
for modelling constitutional courts in a number of contexts (e.g. Ferejohn and
Weingast 1992; Knight 2001; Vanberg 2001). 17 Given the youth of the transition
literature, however, comparatively few attempts to systematically apply rational
choice models to new constitutional courts have appeared regarding the transition
context. The most pertinent example is that of Epstein, Knight, and Shvetsova
(2001), in which they construct a formal model to predict the interaction of
constitutional courts with legislative and executive branches. Grounding their model
in the assumption that courts do not (at least necessarily) have the final say in
determining the outcome of a given constitutional controversy, they both hypothesise
and confirm that constitutional courts can only survive, let alone increase their
position in the political order, when they issue decisions which all of the other
branches find it too costly to contest. For empirical support, they apply their model to
the decisions of the Russian Constitutional Court in the decade following transition,
though they assert that their model is also well suited to many other constitutional
courts in post-communist contexts.
The virtues of this type of analysis are indeed several, but most the important
is the most obvious: it offers a view inside the ‘black box’ of judicial decision
making vis-à-vis political influences on the courts as strategic actors. However, there
are several problems which must be dealt with. First, much of these courts’ most
important cases concern conflicts between other branches of government. These,
attention toward CEE and FSU constitutional courts, they have made more use of the first two of these
approaches, while the third has been partly subsumed in historical-contextual accounts (e.g. Procházka
2003; Boulanger 2003a; 2003b).
17
Applications of rational choice models to constitutional courts have been wide ranging, including
inter alia the USA (Epstein and Knight 1998), Germany (Vanberg 1998; 2000; 2001; 2005), the
Netherlands (Steunenberg 1997), and the European Court of Justice (Garrett 1995; 1998).
5
however, are precisely the cases which the Epstein, Knight, and Shvetsova (2001)
model is least able to explain: their model excludes the possibility of the court
deciding in favour of one branch of government, over and against another, and if
necessary relying on the branch it supported to protect it from attacks. Yet the courts
of Hungary (Scheppele 2003b; Schwartz 2000; Halmai 2002) and Poland (Brzezinski
2000; Schwartz 2000) did precisely this. 18 Second, as is implicit in Boulanger’s
(2003, 12ff.) critique, models which do accurately portray the nature of political
decisions faced by courts in a given country are prone ipso facto to become too
country-specific to be of general value. Third, as Stone Sweet (1998) has observed,
rational choice models—and game theory in particular—is unable to explain
situations where the ‘rules of the game’ are in flux, especially when such changes are
the result of interaction among the players. 19
2.2.2. Behavioural-Empiricist Investigations
In post-communist transition contexts, the main alternative to formal rational choice
models has been statistical analysis of case outcomes (viz. the “behaviouralempiricist” approach; Boulanger 2003a, 7). Research of this type has examined both
the creation of constitutional courts in CEE and the FSU (e.g. Smithey and Ishiyama
2000) and the actual behaviour of the courts once established (e.g. Smithey and
Ishiyama 2002; Herron and Randazzo 2003). Regarding the former, Smithey and
Ishiyama (2000) discovered that the level of judicial independence chosen for the new
institutions did not depend on economics, political history, ethnic factors, or legal
institutions under the prior regime, but rather on the political conditions in the form of
party structure at the outset of transition. 20 Specifically, they found a strong negative
relationship (contrary to their hypothesis) between the number of political parties and
the degree of judicial independence with which their new constitutional courts were
endowed. 21 To measure judicial independence, these authors devised what has since
been dubbed the “Smithey- Ishiyama Index” (SII). However, while in some cases it is
used as a measure of “judicial independence” (Smithey and Ishiyama 2000; Herron
and Randazzo 2003; Boulanger 2003a), in other instances it is termed a measure of
“judicial power” (Smithey and Ishiyama 2002).
Building on the work of Smithey and Ishiyama (2000), and utilising the same
synthetic SII, both Smithey and Ishiyama themselves (2002) and Herron and
Randazzo (2003) examined what relationship, if any, existed between the level of
judicial independence and the incidence of judicial review. Except for differences in
time span examined, 22 the studies are markedly similar. Both projects were tailored
18
Additionally, the Epstein, Knight, and Shvetsova (2001) model ignores the role of parties/coalitions
within legislatures or cabinets, which ought not and need not be excluded from rational-choice-type
models, as evidenced by Steunenberg (1997).
19
The third analytical framework in the broader literature on courts – the ‘triadic dispute resolution’
model (Shapiro 1981; Stone Sweet 2000) – sets out to solve precisely this problem.
20
See also Schiemann (2001).
21
Importantly, as will be discussed below, this contrasts with some of the expectations suggested by
the ‘demand for law’ literature and scholarship on legal importation and transplantation.
22
Smithey and Ishiyama (2002) studied only the first three years of each court’s operation, while
Herron and Randazzo (2003) examined all available years.
6
to capture the influence of various country- (and for Herron and Randazzo, decisionlevel) attributes on the level of judicial review. Both pairs examined a (weighted)
cross-sectional pool of constitutional court decisions from similar 23 panels of
countries, employ the same SII originally constructed by Smithey and Ishiyama
(2000), and both employ similar logit and/or probit 24 regression models to isolate the
effects of macro and micro attributes.
While their studies did still differ in important respects, both yielded similar
and some surprising results. First, though both pairs of authors had expected to find a
positive relationship between judicial independence and review, Herron and Randazzo
could find no significant relationship, and Smithey and Ishiyama identified a
significant negative relationship. 25 Each study, however, highlighted several
contextual factors—political, social, and even economic—which were more strongly
associated with judicial review. In particular, Smithey and Ishiyama (2002) identified
significant positive relationships between judicial review and the degree of legislative
fragmentation, the number of elected layers of government, and popular trust in the
courts.
Herron and Randazzo (2003) identified significant negative relationships
between judicial review and change in the country’s GDP growth as well as the level
of presidential power. At the level of individual decisions, Herron and Randazzo also
found that cases where the president or an ordinary individual citizen was the
appellant were more likely to result in judicial review, as were cases concerning
economic issues. Consequently, each pair of authors concluded that these aspects of
context, rather than judicial independence embedded in institutional design, must be
the dominating factor in explaining the experience of constitutional courts in CEE and
the FSU.
2.3. GROUNDS FOR SCEPTICISM
These works by Smithey and Ishiyama (2002) and Herron and Randazzo (2003) bear
directly on the question posed at the outset: namely, what factors affect and/or predict
the level of judicial review in transitioning states? If one accepts at face value the key
conclusion reached by both pairs of researchers—that judicial independence is not a
key determinant of judicial review in transitioning states (or is negatively correlated to
review)—then part of the puzzle has been solved, and some other explanation of the
degree of review in these states must be sought.
A closer look at the design and execution of both Smithey and Ishiyama’s and
Herron and Randazzo’s research, however, indicates that taking these investigations’
23
Each includes the Czech Republic, Estonia, Georgia, Lithuania, Moldova, and Russia; to this
common set of six, Smithey and Ishiyama add Latvia and Slovakia, and Herron and Randazzo add
Slovenia.
24
Smithey and Ishiyama employ logistic (logit) regression, while Herron and Randazzo employ probit;
as the underlying dependent variable is nominal, however, logit appears a more appropriate
specification, and is thus used here. See Pampel (2000).
25
In addition, Smithey and Ishiyama (2002) observed, but did not test statistically, that judicial
activism (defined as the frequency of judicial review) rose with judicial independence (measured by
their judicial independence index scale) up to around 0.55 on the judicial – approximately the mean and
median of the countries in their sample as well as the sample utilised in their earlier (2000) work which
first employed the index – and then judicial activism begins decreasing, thus presumably yielding a
unimodal peak. This possibility is tested and verified in the replication study herein.
7
conclusions at face value would be unwise. The most important grounds for
scepticism fall primarily into two categories: data and cases selected for their studies,
and the conceptualisation of the independent variables. After reviewing these key
deficiencies, it will become apparent that revisiting the ground these authors covered,
via a modified and expanded replication study, is essential.
2.3.1.1. Data and Case Selection
Three attributes of Herron and Randazzo’s (2003) and Smithey and Ishiyama’s (2002)
concerning data and case selection should immediately stand out to those familiar
with the work of these courts. First, both studies exclude several of the most visible
and most intensely investigated courts in the region. Both, for instance, exclude
Hungary, which numerous scholars hold to have been, during the period studied by
these authors, “perhaps the most activist constitutional court not only in the CEE but
also in the world” (Sadurski 2001, 3; Osiatynski 1994, 151ff.; Brunner 1992, 539-40;
see also Scheppele 2003; Schwartz 2000; Sólyom 2003; Procházka 2003; Halmai
2002). Both likewise exclude Poland, a court which began as one of the weakest in
the region, but began a steady rise to power years before its authority was formally
expanded in the 1997 Constitution (Brzezinski 2000; Schwartz 2000). 26
Second, the studies exclude or appear to mishandle one of the most important
areas of the courts’ activity, that of abstract review. The new constitutional courts in
post-communist states, and especially those of CEE, were modelled not on AngloAmerican templates such as the USA Supreme Court; instead, nearly all imported
most of their structure from Western European constitutional courts, specifically that
of Germany’s FCC (Tanchev 2000; Sólyom and Brunner 2000; Procházka 2003).27
For courts based on this model, the abstract review competence—which enables them
to adjudicate the constitutionality (or legality, for sub-statutory acts) outside of the
context of a concrete dispute 28 —is central to their activity and their power (Stone
Sweet 45ff.), especially in transition countries (Procházka 2003; Sólyom 2003). It is
surprising, then, that Herron and Randazzo (2003) exclude cases involving the
26
The authors of both studies indicated (in response to queries) data availability as the grounds for
excluding Hungary, Poland, and other countries, both pairs of researchers indicated that data
availability was a concern. However, it would appear that such data limitations result from those
authors’ decision to use decisions published on court websites as the source of their data, which is
commented on below. Cases from Hungarian Court, however, were available in non-electronic form in
Hungarian from the first case forward (in the Hungarian Official Gazette), began to be published in
German from 1995 forward (Dupré 2003, 7), and are now available in print in a number of languages
including English.
27
The model adopted in CEE represents the third generation of Kelsenian constitutional courts, after
the pattern designed by Hans Kelsen for Austria in 1920 and later imported into post-war Germany and
Italy (Sólyom 2003). The new courts, importantly, are linked to the Kelsenian model both in that they
imported much of their structure (and in some cases, substantive jurisprudence – see Dupré 2003) from
Germany, and also through more direct ties to the original Kelsenian design (e.g. As Schwartz [2000,
270-71, note 9] recounts, quoting an unpublished manuscript by Scheppele, “the idea for the court
originated with the Socialist Justice Minister Kalman Kulczar, who was with Hans Kelsen at the
University of California at Berkeley in the early 1960s”; citing: Scheppele, American University,
Washington D.C., Sept. 27, 1996, transcript on file with author.)
28
There remains, surprisingly, non-trivial disagreement over the border between concrete and abstract
review; e.g. Procházka (2003, 79-80) distinguishes his own view from that of Stone Sweet (2000) in
regard to review by constitutional courts of lower court decisions, which Procházka believes can be
seen as abstract but which Stone Sweet classifies as concrete only.
8
exercise of abstract review from their sample altogether. 29 Smithey and Ishiyama
(2002), meanwhile, seem to conflate a priori review (which is abstract by nature) with
a posteriori abstract review. 30 This ostensible confusion—explicit in their earlier
work (2000, 167-68) as well—makes it difficult to decipher how Smithey and
Ishiyama conceptualise judicial review as well as to precisely assess the scope of their
research.
One final data concern relates to the actual source of the studies’ data. Both
works used cases available from the websites of the respective courts, using a
combination of original language publications and English translations. 31 This is
problematic for two reasons. First, the selectivity of website publication of decisions
by courts may mean the range of those decisions used by the study are not
representative. 32 While most of the new constitutional courts are required by law to
publish their decisions in an official state publication, publishing any (let alone all) of
those decisions on websites is neither required nor practiced. In Hungary, for
instance, a total of 11,092 proceedings were initiated between 1990 and 1996 (Sólyom
and Brunner 2000, 72), but the Court’s website currently lists only 3,837 decisions or
case descriptions in Hungarian, and fewer than 50 in English.
Some degree of selectivity is unavoidable, of course. This is to be expected
with the output of courts, as many courts publish only their more important decisions.
The difficulty, however, lies in the differences between the courts’ procedures and
standards for choosing cases for electronic publication. Whereas some courts offer
electronic versions of every published decision (e.g. Slovakia), others offer only a
selection of the most important cases (e.g. Hungary, Czech Republic). Because these
decisions about which decisions to make available online are taken by different actors
according to different standards in each country (whereas publication in official state
bulletins is more uniform), the potential result is that the sample used by these two
studies from one country will be more representative of its entire jurisprudence, while
the sample taken from another country will reflect only the more important cases.
This differential in bias between countries could be partially mitigated if a single body
utilising a single set of standards selected the cases from each jurisdiction.
29
Herron and Randazzo (2003, 429) give some reasons for their exclusion of cases of what they term
“abstract review”, but their description applies only to instances where legislators petition for review.
The courts of CEE and the FSU do not limit this power to legislators, however, and thus their
arguments, as stated, are unpersuasive.
30
The difference between the two categories is actually quite important. For purpose of
disambiguation, a priori review – or preventive norm control – enables the court to review statutes
prior to either passage by the legislature, promulgation by the executive, or application. A posteriori
review – or repressive norm control – can include both abstract and concrete review of statutes after
they have been passed, promulgated, and brought to application (Schwartz 2000; Sólyom and Brunner
2000; Procházka 2003).
31
Herron and Randazzo explicitly state this (2003, 428), and both pairs of authors confirmed this in
response to queries by the author.
32
Herron and Randazzo (2003, 428) do note that their study is limited only to published decisions, and
thus generalizable only to the universe of published decisions; however, it is their conflation of
publishing of decisions with publishing electronically, and in some cases in an accessible language
which creates the difficulty.
9
2.3.1.2. Operationalising the Independent Variable
As noted above, both Herron and Randazzo (2003) and Smithey and Ishiyama (2002)
employ the Smithey and Ishiyama Index (SII) of “judicial independence” first
developed and presented by Smithey and Ishiyama (2000, 167-69). The SII is a scalar
measure, ranging from 0 to 1, higher scores reflecting greater independence. The
score is comprised of the average of six individual scores which also range from 0 to
1, each of which is designed to reflect a single aspect of independence. 33 However,
due to deficiencies in both the operationalisation and design of the SII, the substantive
results generated by both studies which rely upon it must be called into question. As
both studies use the SII (and the original scores) without modification, errors at this
initial stage are pertinent.
First, several instances of errors in the coding of countries in the original SII
give cause for concern. 34 Specifically, coding errors apparent in measuring countries’
a priori review powers, judges’ relative term length, and number of parties involved
in appointment of judges. Moreover, there is at least one important arithmetical error
in their tabulations, which yields the strikingly counter-intuitive result that Georgia’s
court is nearly equal to Hungary’s in independence. Thus, even if the SII were a valid
measure by construction, there are reasons to doubt its accuracy.
Beyond these important (but remediable) flaws in the use of the SII, there are
reasons to doubt the conceptual soundness and validity of the measure in its current
form. The six components are intended to be representative of several vaguely
defined dimensions of judicial independence, and this is precisely where they are
weakest. 35 For instance, though the SII is used by its designers as a measure of
“judicial power” (Smithey and Ishiyama 2002, 731), only two of its six components
directly relate to the competences of the court, while the rest deal with elements of the
judges’ protection from manipulation by other branches. While both dimensions are
clearly important to judicial independence, the balance or relationship between them
is not specified by the authors, yet the composition of their index hinges on this
balance. In sum, the SII does not reflect a well-grounded conception of the factors
which comprise judicial independence, nor of the interactive relationships between
those component factors. Consequently, the measure’s validity in measuring these
concepts is suspect. In short, there is serious doubt that the SII is capable of
meaningfully measuring the level of judicial independence enjoyed by a court.
2.3.2. The Need for Replication
The conclusions reached by both Smithey and Ishiyama (2002) and Herron and
Randazzo (2003) must prompt one of two responses: (1) accept that judicial
independence neither significantly nor directly influences judicial review in transition
33
See Table 2 (and notes accompanying it) for original and corrected scores (corrections coded by the
author) on each dimension of the index. See Smithey and Ishiyama (2000, 167-169) for a detailed
discussion of the construction and coding of the index.
34
A full account of the errors described summarily here is provided in the notes to Table 2.
35
Moreover, even if the balance of areas represented by the six components was appropriate, the
particular measures used to reflect each of the six components are markedly deficient proxies for the
dimensions they are used to represent.
10
states; or (2) due to problems in the design and execution of these works, approach
these findings critically until the evidence can be re-examined. The foregoing
discussion points to the second response. 36 It may be that their original findings were
correct, but before those findings can either be relied upon substantively or used to
guide further research, they must be tested more rigorously. In the process of
replication, the opportunity also arises to incorporate additional variables—both to
test alternative causal stories and to control for important factors not accounted for in
the original studies. It is precisely this endeavour which is undertaken here.
3. DATA AND METHODOLOGY
3.1. HYPOTHESES
The main hypotheses of this study concern the effect on judicial review of judicial
independence, legislative fragmentation, 37 and several new variables introduced to
measure the ‘receptiveness’ of the transition countries to the new legal institutions. 38
Regarding judicial independence, both pairs of authors—drawing on substantial
support from the literature—anticipate that judicial independence and/or power is
positively related to the degree of judicial review, which both define as the frequency
of judicial review. This expectation is premised on the view that courts enjoying
“greater guarantees of independence” are “freer to exercise their own will” without
fear of censure or retribution by the other branches (Herron and Randazzo 2003, 425).
The primary hypothesis for judicial independence is therefore:
H1-A: Judicial independence is positively related to the frequency of
judicial review (exercised by the constitutional court).
Because the prior authors (Smithey and Ishiyama 2002; Herron and Randazzo 2003)
use the exact same measure (the SII) to capture both judicial independence and
judicial power, and because the present work aims to disambiguate these concepts, an
alternative formulation of this hypothesis is:
H1-B: Judicial power is positively related to the frequency of judicial
review (exercised by the constitutional court).
It is specifically anticipated that with each improvement in the execution of the study
relative to the works being replicated, this relationship between judicial independence
and judicial review should become gradually more apparent.
Intriguingly, the two pairs of authors harboured different expectations
toward the effect of legislative fragmentation on judicial review. Herron and
B
36
There is debate, of course, over the proper definition of “replication” (Herrndon 1995; see also King
1995). The present study satisfies Hernndon’s definition, whereby a replication study must gather new
data. Incidentally, both codebooks and datasets for replication were requested from the original
authors, but in both cases data was unavailable (due to ongoing research from the same dataset and
codebook).
37
As both the expectations and findings of Smithey and Ishiyama (2002) and Herron and Randazzo
(2003) diverged on this issue, a clear hypothesis is necessary.
38
The other variables included by Smithey and Ishiyama (2002) and Herron and Randazzo (2003) are
likewise tested here for purposes of replication, but separate hypotheses are not offered for these.
Expectations for each variable are presented in the following section.
11
Randazzo (2003), building on a familiar argument by Stone Sweet (2000, 54ff.), argue
that a more divided legislature is likely to pass laws which are “generally less
contentious than those produced by one dominant party” (Herron and Randazzo 2003,
427). The legislation thus produced will be less prone to challenge, as presumably
more parties were involved in its design and thus have less reason to petition for its
nullification.
However, the established literature more strongly supports the expectations
and findings of Smithey and Ishiyama (2002) that legislative fragmentation is
positively associated with judicial review. As a number of scholars observe, less
unified legislatures and/or governing coalitions invite challenges by the judiciary,
which (the judiciary expects) the legislature and/or coalition will be unable to override
(Vanberg 2001; see also Tate and Vallinder 1995, 31ff.; Holland 1991, 9ff). 39 The
primary hypothesis for legislative fragmentation follows Smithey and Ishiyama’s
expectations and results:
H2-A: Legislative fragmentation is positively related to the frequency of
judicial review.
The view of Herron and Randazzo (2003) that diverse participation may yield
less contentious legislative output should not be discarded entirely, however.
Specifically, Herron and Randazzo’s expectation might be true of fragmentation
within governing coalitions, for the following (conjectural) reasons. First, while
overall fragmentation would not seem to guarantee that a party’s preferences are
incorporated into the legislative programme, membership in the coalition might offer
more assurance of this. Second, while overall fragmentation would serve to weaken
the legislature’s ability to override a court which struck down the legislature’s bills (in
that overriding a court may require, as it does in many transition countries, a supermajority vote), fragmentation within the coalition seems less likely to generate this
weakness: despite its internal divisions, the members of a coalition would
presumably be willing to defend the coalition’s policies against interference by the
courts. A secondary hypothesis for legislative fragmentation is therefore:
H2-B: Legislative fragmentation within the governing coalition is positively
related to the frequency of judicial review.
For the remainder of the variables included for purposes of replication,
separate hypotheses are not stated, as the original authors’ hypotheses and findings
serve as the propositions to be verified or rejected (King 1995; Herrndon 1995).
However, a few final hypotheses are called for in regard to several new independent
variables, not examined by either Smithey and Ishiyama (2002) or Herron and
Randazzo (2003). These variables reflect aspects of the “demand for constitutional
law” (Pistor 2002): familiarity with the imported legal traditions, adaptation of
imported institutions, and political participation in the design of new institutions.
These variables aim to capture the receptiveness of each country to the
‘transplantation’ (Watson 1993) or ‘importation’ (Ajani 1995; Dupré 2003; Pistor
2002) of legal institutions, including constitutional adjudicatory systems.
According to the typology developed by Pistor, Raiser, and Gelfer (2000,
15ff.), receptiveness is comprised of adaptation and familiarity, and transplants of law
to receptive countries are more successful than those to unreceptive countries. Those
B
39
A possible exception pertinent to some transitioning states is noted by Procházka (2003, 117), who
suggests that legislatures divided to the point of fragility might receive extra deference from a
constitutional court concerned more with stability of the new state than with conformity of certain legal
provisions with the constitution.
12
countries “without previous exposure to the modern formal legal order before the
collapse of the socialist system” are termed “new transplants.”
The
unifying
argument across the legal importation literature in this vein is that in countries which
are ‘receptive’ to the institutions they import, the import/transplant will be successful,
whereas in unreceptive countries the institutions grafted in are more likely to operate
dysfunctionally, exhibiting what scholars have termed a legal “transplant effect”
(Berkowitz, Pistor, and Richard 2003a; 2003b; Pistor 2002; Dupré 2003; Ajani
1995). 40 However, while those advancing this framework (Pistor, Raiser, and Gelfer
2000) expect and find “new transplants” to function similarly to unreceptive
transplants, there is reason to think this may be inaccurate in the case of constitutional
judicature. 41
What dysfunctionality means in the case of constitutional review, however, is
not immediately clear, nor addressed directly in the literature. One conjectural answer
is that dysfunctional courts do not develop in the spiral-shaped pattern, based on the
judicialization-politicization cycle (Stone Sweet 2000). Since this pattern is
associated with continually (if subtly) increasing judicial power and purview, it is
speculated here that judicial review is positively associated with the aspects of the
demand for law indicated above. Thus, the final hypotheses are as follows:
H3:
The more a country adapts the legal institutions it imports, the
higher the level of judicial review its constitutional court exercises.
H4:
The greater a country’s familiarity institutions it imports, the
higher the level of judicial review.
H5:
Judicial review is higher in countries ‘receptive’ to legal
transplants than those that are ‘unreceptive’.
H6:
Judicial review is higher in countries which are ‘new transplants’
than those which are ‘unreceptive’.
H7:
The greater the diversity of political participation in the decisions
to design and/or import legal institutions at the beginning of
transition, the higher the level of judicial review.
This last hypothesis connects also to the omnipresent principal-agent theories of
courts—especially constitutional judiciaries—pervasive in the literature. Specifically,
more ‘principals’ engaged in the design of a new constitutional court implies 1) more
legitimacy and support it can presumably command (Pistor 2002, 84) and 2) less unity
among the principals, meaning a) the court can become powerful by adjudicating
disputes between principals, and/or b) the principals may find it more costly to
constrain the court.
3.2. VARIABLES AND DATA
The study examines the decisions of constitutional courts of ten countries of CEE and
the FSU: the Czech Republic, Estonia, Georgia, Hungary, Lithuania, Moldova,
40
Pistor, Raiser, and Gelfer (2000) (and works which build on this framework) identify countries
classed either as exhibiting adaptation or familiarity as receptive; intuition suggests altering this
definition to limit it to those countries exhibiting both.
41
Namely, a country with no exposure to the system might be thought of as having a ‘clean slate’
compared to those which are distinguishably “unreceptive” for substantive reasons. For example, in
the case of Poland, the Constitutional Tribunal began operations several years prior to the formal
transition to democracy, but this brief heritage proved all but fatal to the Tribunal as a potent political
actor in its early years after transition. See Schwartz (2000, 264, at note 10).
13
Poland, Russia, Slovakia, and Slovenia. This includes all but one of the countries
(Latvia 42 ) examined by both Herron and Randazzo (2003) and Smithey and Ishiyama
(2002). All cases (n = 915) from these ten countries where judicial review was
requested by the petition or referral to the court are included, from the earliest
(available) year of each court’s operation through the end of 2003. Table 1 presents a
comparison, by country, of the samples examined by Smithey and Ishiyama (2002),
Herron and Randazzo (2003), and the modified replication study.
TABLE 1. SAMPLE COMPARISONS—PRIOR RESEARCH VS. REPLICATION STUDY
Herron and
Randazzo (2003)
Smithey and
Ishiyama (2002)
Replication Study
Country
Czech
Republic
Estonia
Georgia
Hungary
Latvia
Lithuania
Moldova
Poland
Russia
Slovakia
Slovenia
N
11
Years
1992—1996
N
10
Years
1993—1995
N
61
Years
1993—2003
42
11
0
0
103
228
0
86
0
93
1993—2000
1996—1997
n/a
n/a
1993—2000
1995—2000
n/a
1995—1998
n/a
1993—1995
19
11
0
13
41
74
0
59
96
0
1993—1995
1997—1999
n/a
1993—1995
1993—1995
1995—1997
n/a
1995—1997
1993—1995
n/a
58
17
118
1994—2003
1996—2003
1991—2003
n/a
1993—2003
1998—2003
1991—2003
1995—2003
1994—2003
1992—2003
TOTAL
574
323
136
33
250
60
47
135
915
Decisions involving abstract review (both a priori and a posteriori) of legal
norms were included along with concrete review. The source of the data is the
CODICES Database published by the Venice Commission of the Council of Europe
(see Appendix, part A), which provides full-text decisions and case descriptions from
constitutional courts around the world. The CODICES Database does not entirely
remove the problem of selection bias, but it significantly reduces the problem of
differential selection bias in that cases are selected according to a single set of
standards by a central agent.
3.2.1. Dependent Variable: Judicial Review
To replicate the prior works’ methods, both bivariate correlation analyses and logistic
regression analysis (logit) are employed. For the logit analysis, which forms the main
part of the study, the dependent variable is the likelihood of a constitutional court
exercising its powers of judicial review. Following Herron and Randazzo (2003) and
42
Excluded to avoid overrepresentation of the Baltic region; both Estonia and Lithuania were examined
by both sets of authors, and thus were the better of the three candidates for inclusion.
14
Smithey and Ishiyama (2002), cases were coded 1 if the courts declared
unconstitutional or otherwise annulled a statute, explicit government action,
substatutory legislation or regulation, a treaty, or other similar legal norms, and 0 if
the courts did not do so.
For the correlation analysis, following Smithey and Ishiyama (2002), static,
country-level attributes are compared with an aggregate measure of the frequency of
judicial review: the number of cases where a court did exercise its review powers
across all years divided by the number of cases where it was petitioned (or otherwise
empowered) to do so. Three measures of this were used: (1) the scores originally
calculated by Smithey and Ishiyama (2002) covering the first three years of each
court’s operation, (2) the same scores supplemented by third-party data on Hungary
and Poland, 43 and (3) scores calculated from the replication dataset. In a second set of
correlations presented side-by-side with counterparts of the first set, each of these
three measures of the dependent variable is adjusted to reflect disparate levels of
accessibility of the courts in each country.
3.2.2. Independent Variables
3.2.2.1. Judicial Independence: Power, Access, and Insularity
The prior works operationalise judicial power in the form of the SII described above,
a synthetic measure designed to encompass “both the extent to which the
constitutional courts possess judicial review powers and the extent to which the
constitution extends independence of action to the constitutional court or supreme
court from other institutional actors” (Smithey and Ishiyama 2000, 167). The
measure is computed as the average of six component variables, each ranging from 0
(not independent) to 1 (independent), the final score thus also ranging from 0 to 1.
The present study begins with this measure judicial independence, but only as a point
of departure. First, certain arithmetical flaws in the original computations are fixed
(which were not corrected in either Smithey and Ishiyama [2002] nor Herron and
Randazzo [2003]). Second, apparent errors in actually coding the countries were
corrected. Table 2A presents the Smithey and Ishiyama Index (SII) of judicial
independence as originally calculated. Table 2B presents the SII when corrected for
coding and calculation errors.
43
Data from Scheppele (2003b) and Brzezinski (2000), respectively; see the Appendix, Part A for full
references and details of all primary source data.
15
TABLE 2A. SMITHEY AND ISHIYAMA INDEX (SII) OF JUDICIAL INDEPENDENCE
(2000): ORIGINAL INDEX
Country:
(A) Can
judicial
decision be
overturned?
(B)
Presence
of a
priori
review?
(C) Judges
term
relative
legislative
session
(D) No. of
actors
involved in
selecting
judges
(E) Who
establishes
court
procedures
(F)
Conditions
for judicial
removal
Judicial
power
score:
A+B+C+
D+E+F/6
Czech
Republic
Estonia
1.00
1.00
0.00
0.00
0.33
0.33
0.50
0.50
1.00
0.00
0.50
0.50
1.00
1.00
1.00
1.00
0.00
0.00
0.00
0.00
0.33
1.00
0.67
0.33
1.00
0.50
1.00
1.00
0.00
0.00
1.00
1.00
0.00
1.00
0.50
1.00
0.56
0.39
0.56
(0.39)†
0.58
0.70
0.72
0.00
0.00
0.00
0.00
0.00
0.00
0.00
0.00
1.00
1.00
1.00
0.00
0.00
0.00
0.00
0.00
1.00
0.33
0.33
0.00
0.50
0.50
0.50
0.00
0.00
0.00
1.00
0.00
0.00
0.00
0.50
0.00
0.42
0.31
0.56
Georgia
Hungary
Lithuania
Moldova
Poland
(19891997/9)
Poland
(1997/9)
Russia
Slovakia
Slovenia
16
TABLE 2B. SMITHEY AND ISHIYAMA INDEX (SII) OF JUDICIAL INDEPENDENCE
(2000): DATA-CORRECTED INDEX
Country:
(A) Can
judicial
decision be
overturned?
44
(B)
Presence
of a
priori
review?
45
(C)
Judges
term
relative
legislativ
e session
(D) No.
of actors
involved
in
selecting
judges 47
(E) Who
establishes
court
procedures
(F)
Conditio
ns for
judicial
removal
Judicial
power
score:
A+B+C+
D+E+F/6
46
Czech
Republic
Estonia
Georgia
Hungary
Lithuania
Moldova
Poland
(198997/9)
Poland
(1997/9)
Russia
Slovakia
Slovenia
Net Difference
1.00
1.00
1.00
1.00
1.00
1.00
0.00
1.00
0.00
1.00
0.00
0.00
0.67
0.33
0.67
1.00
0.67
0.33
0.50
1.00
1.00
0.25
1.00
1.00
1.00
0.00
0.00
0.00
1.00
1.00
0.50
0.50
0.00
1.00
0.50
1.00
0.61
0.64
0.45
0.71
0.70
0.72
0.00
0.00
0.67
0.25
0.00
0.00
0.15
0.15
1.00
1.00
1.00
0.00
1.00
0.00
0.67
1.00
0.67
0.25
0.50
0.50
0.00
0.00
0.00
0.00
0.00
0.00
0.32
0.50
0.36
0.32
0.08
0.05
1.00
0.50
0.67
0.50
1.00
0.50
0.70
0.14
0.05
0.25
-0.11 (0.06)
0.13
0.00
0.00
N.B.: Scores in bold underline indicate a change from the original Smithey and Ishiyama (2000) score.
44
Poland's Constitutional Tribunal gained the ability to issue binding decisions which the Sejm cannot
override in the 1997 Constitution (which took effect in October 1997), and the provisions concerning
binding decisions (not open to overruling by the Sejm) took effect in October 1999. Smithey and
Ishiyama (2000) acknowledge this but ignore the change in computing Poland's index score.
45
From the text of their respective constitutions and laws empowering their constitutional courts, it is
clear that the Supreme Court (Constitutional Chamber) in Estonia and the Constitutional Court of
Hungary can each consider constitutional challenges prior to the final promulgation (and thus
application) of legislative acts. Also, though Smithey and Ishiyama specifically consider the possibility
that a court could have a priori jurisdiction in a narrow area of cases, such as the review of not-yetratified treaties, they do not count Russia nor Slovenia in this group, in contrast to the text of those
courts' respective empowering statutes.
46
Smithey and Ishiyama (2000, 168) define the values of this variable as follows: the variable "was
coded as 0 when the term of the constitutional court judge was less than or equal to one term of the
actor with the longest constitutional term; 0.33 when it was less than or equal to two parliamentary
sessions; 0.67 when the judges term was more than two parliamentary sessions (but had
constitutionally specified limit in the number of terms) and 1 when the term was life or until voluntary
retirement". However, the maximum term length of parliamentarians in all of these countries is four
years, and the term lengths of judges are as follows: Czech Republic, 10 years; Georgia, 10 years;
Poland, 9 years; Slovakia, 12 years; and Slovenia, 9 years.
47
In Estonia, the chairman of the Supreme Court is proposed by president, and adopted by national
assembly; others are proposed by the chairman, and adopted by national assembly; the five members of
constitutional panel elected by General Assembly of the Supreme Court. In Hungary, the court is
chosen by a special committee of the (unicameral) parliament comprised of one member from each
political party represented in parliament. In Poland, the panel of judges is chosen by the Sejm, but the
President can appoint any of these to be the President and Vice-President, without the approval of the
Sejm. Thus, for Estonia, a score of "1" seems more appropriate, and ".25" for Hungary and Poland.
48
In their original table, Smithey and Ishiyama report this total as .56, when in fact the data in their
table generate a value of .39.
17
48
Additionally, Smithey and Ishiyama (2002) observe casually, without testing
systematically or attempting to explain, that the level of judicial review generally rises
with judicial independence up to 0.55, a point immediately between the mean (0.54)
and median (0.56) and mode (0.56) of the original distribution of judicial
independence scores (Smithey and Ishiyama 2000), but that judicial review declines
thereafter. This possibility of a central peak is tested by comparing the absolute value
of the distance of a country’s SII score from this central ‘peak’ point (0.55) with its
level of judicial review. If 0.55 is indeed at the apex of a curve, then the distance
between a country’s score and the peak will be negatively correlated to judicial
review.
The limits of the SII itself can only be pushed so far before a wholly new
measure is needed. At the very least, judicial independence and judicial power –
nebulous concepts which the literature has failed to define (Burbank and Friedman
2002; Shetreet and Deschenes 1985) – should not be treated as coterminous. An
improved measure of judicial independence must conceptually identify, separate, and
thoughtfully recombine various dimensions of judicial independence, and should be
designed from theoretical premises rather than an unbalanced, unrepresentative
amalgamation of attributes of transition courts. 49 The present study undertakes this
by identifying three separate conceptual components of judicial independence—
Access, Power, and Insularity—which are tested both separately and interactively for
their influence on judicial review. 50 Table 3A details the composition of the
individual elements used to construct new indices of judicial independence, while
Table 3B lists the computations used to generate these indices.
49
An alternative and/or supplement to designing such an index deductively is factor analysis, a data
reduction method aimed to identify underlying forces in a range of independent variables. However,
when employed in the present case, no single component derived accounted for even a majority of
variance, and the main components identified were not associated with groupings of independent
variables which were intuitively meaningful.
50
See Tables 3A and 3B for a complete specification of all elements and computations of these new
independence measures.
18
TABLE 3A: NEW MEASURES OF JUDICIAL INDEPENDENCE: ELEMENTS OF NEW
INDICES
Power
A priori
A posteriori abstract
A posteriori concrete
Unconstitutional
Omission
Access
Elements of the New Indices
Average of binary scores for each of the following powers (1: court has this power; 0:
court lacks this power):
Review of: Constitution itself; International Agreements; Statutes; Regulations; Acts
of the President; Acts of local/regional units; Other General Acts
Review of: Constitution; International Agreements; Statutes; Parliamentary
Resolutions; Regulations; Acts of the President of the State; Local Government
Statutes; General Acts-Exercise of Public Powers; Other General Acts; National
Norms versus Treaties; Regioal Agreements
Review of constitutional complaints
Declare instance of unconstitutionality by omission (i.e. legislature has failed to act
where it is obligated to do so)
Average of binary scores for each of the following actors who have this right of access
(1: this actor can invoke court for this power; 0: this actor cannot invoke court for this
power):
A priori
Initiation by: President; Parliamentary leadership/cabinet; Individual members of
parliament; Second legislative chamber's leadership; Individual members of second
legislative chamber; Court itself; Other/lower courts; Administrative or legal official;
Any citizen.
A posteriori abstract
Initiation by: President; Parliamentary leadership/cabinet; Individual members of
parliament; Second legislative chamber's leadership; Individual members of second
legislative chamber; Court itself; Other/lower courts; Administrative or legal official;
Any citizen.
A posteriori concrete
Initiation by: Court itself; Lower court; Administrative/legal official; Local/regional
government; Individuals with specific vested interest; Any individual citizen
regardless of interest
Unconstitutional
Omission
Initiation by: Court itself; Any individual citizen regardless of interest
Immunity
Values: 0: Judges have no special immunity; 0.5: Judges can be removed only with
legislature's consent; 1.0: Judges can be removed only with the Court's own consent.
Number of Effective
Appointers
= 1 / ∑ (ai2)
Where ai is the proportion of control over Court appointments allotted
to each actor i; same basic modification of Laakso-Taagepera (based on Herfindahl)
formula for effective number of political parties. N.B. where two parties share
authority for the same appointment, the proportion they share is split between them
(e.g. if Legislature and President each must confirm every appointee, they are each
coded 0.50).
Relative Term Length
= T / L Where T is the term length of a single judge in years and L is the length of
a legislative session
Control of Court's
Procedures
Smithey and Ishiyama Index, component E; Values: Court sets own procedures = 1;
Else = 0
Removal Score
Smithey and Ishiyama Index, component F; Values: Constitutional bar on removing
judges = 1; Else 0
19
TABLE 3B: NEW MEASURES OF JUDICIAL INDEPENDENCE:
COMPUTATION OF NEW INDICES
New Indices of Dimensions of Judicial Independence
Power
= [(a priori power) + (a posteriori abstract power) +
a posteriori concrete power) + (omission power)] / 4
Access
= [(a priori access) + (a posteriori abstract access) +
(a posteriori concrete access) + (omission access)] / 4
Insularity
= [(Immunity score) + (number of effective appointers)/4 +
(relative term length)/4 + (removal score) + (control of procedure)] / 5
Composite measure of
power and access
= [(a priori power)*(a priori access) + (a posteriori abstract power)*
(a posteriori abstract access) + (a posteriori concrete power)*(a
posteriori concrete access) + (omission power)*(omission access)] / 4
Power * Access
= [(a priori power) + (a posteriori abstract power) +
(a posteriori concrete power) + (omission power)] / 4 * [(a priori access)
+ (a posteriori abstract access) + (a posteriori concrete access) +
(omission access)] / 4
Power * Insularity
= [(a priori power) + (a posteriori abstract power) +
(a posteriori concrete power) + (omission power)] / 4 *
[(Immunity score) + (number of effective appointers)/4 +
(relative term length)/4 + (removal score) + (control of procedure)] / 5
= { [(a priori power)*(a priori access) + (a posteriori abstract power)*
(a posteriori abstract access) + (a posteriori concrete power)*(a
posteriori concrete access) + (omission power)*(omission access)] / 4 } *
{[(Immunity score) + (number of effective appointers)/4 +
(relative term length)/4 + (removal score) + (control of procedure)] / 5}
Composite measure of
power and access *
Insularity
‘Access’ refers to the range of actors authorised to invoke the court’s
jurisdiction and request judicial review. It is comprised of the average access score of
four types of constitutional review cases, each weighted equally: a priori review, a
posteriori abstract review, a posteriori concrete review, and unconstitutional
omissions. The access score for each of these case types is determined by the number
of categories of petitioner authorised to initiate a case of that type.
‘Power’ refers to the range of the court’s substantive competences, reflecting
the scope and depth of its judicial review authority. This measure is built in the same
way as Access, constituted of four average power scores (reflecting the proportional
number of powers within each of the four case types which the court possesses) which
are also in turn averaged together.
Finally, the measure ‘Insularity’ reflects the degree to which constitutional
courts are shielded from political attacks or censure for decisions unfavourable to
other political actors. It results from the average of five component scores (each
ranging from 0 to 1) reflecting different aspects of judges’ protection from political
retribution: judges’ immunity from prosecution, 51 the number of effective appointing
51
Values: 0: Judges have no special immunity; 0.5: Judges can be removed only with legislature's
consent; 1.0: Judges can be removed only with the Court's own consent.
20
actors (deflated by a factor of 4 to yield an index with limits at 0 and 1), 52 the term
length of the judges relative to a legislative session (likewise deflated by 4), 53 the
original SII component for removal, 54 and the original SII component for who
controls court procedure. 55
These basic elements are tested both by themselves and in certain specific
combinations to test interactive relationships. Thus, the measure Power * Insularity
captures the interactive relationship expected between competences and protection
from censure, testing the inference that competences on paper only empower the court
to the degree that judges are shielded from retribution. Power * Access aims to adjust
the measure of the court’s competences on paper to reflect the ease with which those
competences are invoked (since even the court with the greatest competences on
paper can achieve nothing if not case can be brought). Also, a Composite measure of
courts’ power and accessibility is computed by averaging the products of a priori
power and access, a posteriori abstract power and access, a posteriori concrete power
and access, and omission power and access. This Composite measure was also
incorporated in an interactive measure: Composite * Insularity.
All the foregoing measures of dimensions of judicial independence have
values ranging from 0 to 1 (1 reflecting greater independence). Each reflects one of
the many possible elements of judicial independence. It is expected that all of these
elements, taken separately and when combined interactively, will relate positively to
judicial review.
3.2.2.2. Additional Replication Variables
For all independent variables included for replication purposes, variable specifications
computation (when necessary) followed the formulae used by the original authors
(data sources for all variables are given in the Appendix, part A). Thus, following
Herron and Randazzo (2003), legislative fragmentation (both at time of decision and
in the first democratic election) was computed using the widely-used 56 LaaksoTaagepera measure of effective political parties. 57 Both of these measures are
expected to relate positively to judicial activism. Data on the number of parties
52
= 1 / ∑ (ai2), where ai is the proportion of control over Court appointments allotted to each actor i;
same basic modification of Laakso-Taagepera (based on Herfindahl) formula for effective number of
political parties.
53
= T / L, where T is the term length of a single judge in years and L is the length of a legislative
session
54
Values: Constitutional bar on removing judges = 1; Else 0
55
Values: Court sets own procedures = 1; Else = 0
56
The measure, incidentally just the inverse of the Herfindahl ownership dispersion index from the
economics literature, is given by: EPP = 1 / ∑ (pi2), where pi is the proportion of seats allocated to each
party i. The index was first presented by Laakso and Taagepera (1979) and later by Taagepera and
Shugart (1989); despite numerous calls for modification (e.g. Dumont and Caulier 2003), some by its
own creators (e.g. Taagepera 1999), it remains the standard measure.
57
Data for political parties’ proportions of seats came from the University of Essex Project on Political
Transformation and the Electoral Process in Post-Communist Europe. All computations of effective
political parties are original to the present author.
21
within the governing coalition is taken from Frye (1997), which is expected to relate
negatively to judicial review for reasons given above.
The measure of and data for presidential power used, following Herron and
Randazzo (2003), also comes from Frye (1997). The measure incorporates twentyseven distinct competences potentially possess by presidents, adjusted for cases where
powers are shared or if the president is indirectly elected. This is expected to relate
negatively to judicial review, in line with both the hypothesis and findings of Herron
and Randazzo (2003).
Both Herron and Randazzo (2003) and Smithey and Ishiyama (2002) include a
measure of rights guarantees in each country, both of which are tested. Herron and
Randazzo use each country’s “Civil Liberties” score from annual Freedom House
surveys. As lower scores on this scale (which ranges from 1 to 7) are associated with
greater civil rights protection, a negative relationship is expected here, which Herron
and Randazzo (2003) expected but were unable to find. Smithey and Ishiyama (2002)
generate their own index reflecting the scope of rights guaranteed in each country’s
bill of rights, which they expected but did not find to be positively related to judicial
review.
Smithey and Ishiyama’s (2002) measure of popular trust in the courts is also
included here, and it is expected that higher values of public confidence in courts
should correspond to higher incidence of judicial review (cf. Vanberg 2005; 1998a),
an expectation which Smithey and Ishiyama (2002) confirmed. Additionally, the
degree of federalism in each country – manifested in the number of elected
subnational tiers of government – was also included in the Smithey and Ishiyama
(Ibid.) models and thus in the replication. The expectation, which Smithey and
Ishiyama’s data supported, is that greater divisions of power between the centre and
regions should result in higher judicial review frequency. 58
The final replication variable, which Herron and Randazzo (2003) include, is
the identity of the petitioner requesting the court to exercise judicial review.
Following their scheme, dummy variables for individual citizen petitioners and the
president as petitioner are included. In addition, the replication study includes
dummies for legislators as well as administrative or cabinet officials (e.g. ministers,
ombudsman, etc.) as petitioners. 59 Each of these dummy variables is expected to
relate positively to judicial review, following Herron and Randazzo’s hypotheses and
findings (2003, 431).
3.2.2.3. New Independent Variables
Aside from legislative fragmentation in the first democratic election (already
discussed), variable specifications and data for measures of countries’ receptiveness
to legal transplantation are taken from Pistor, Raiser, and Gelfer (2000) and
Berkowitz, Pistor, and Richard (2003a). Both sets of authors present the same coding
scheme, used with modifications here, which identifies countries as possessing
Familiarity, Adaptation, both, or neither. Those “without previous exposure to the
modern formal legal order before the collapse of the socialist system” are coded as
58
Smithey and Ishiyama (2002) cite the World Bank’s annual World Development Report as their
source, though there are discrepancies between the edition they cite and the numbers they report. In
these cases, the original World Bank numbers are preferred.
59
Herron and Randazzo identify cases where the legislature is the respondent, not the petitioner.
22
“new transplants” (Pistor, Raiser, and Gelfer, 15). In the present study, the presence
of Familiarity, Adaptation, or both, as well as being coded as “new transplants”, is
expected to relate positively to judicial review, for reasons given above.
3.2.2.4. Control Variables
Both as a possible explanatory independent variable and as a necessary macro-level
control, Herron and Randazzo (2003) incorporate the annual change in GDP growth
in their model. They suggest several reasons why such economic conditions may lead
to either higher or lower levels of judicial review (Ibid., 426), none of which are
compelling, but their findings are uniformly negative and statistically significant. The
present study thus incorporates change in GDP growth as well (base year = 1990),
expecting negative results in line with Herron and Randazzo’s (2003) finding.
Additionally, in the correlation analysis only, each set of correlations were run
a second time controlling for ease of access to each court (viz. the judicial review
aggregate statistic was multiplied by the country’s Access score). This control is
incorporated to account for the possibility that easier access to the courts yields not
only an increase in case volume but also a decrease in the proportion of cases with
strong legal merits. 60
Consequently, courts such as Hungary where access is wide open to all (under
the actio popularis, any individual can petition for abstract review of any statute
without having been affected by the statute herself; See Hungarian Constitution and
Act on the Constitutional Court, Act XXXII of 1989), it makes sense that a smaller
proportion of the 11,092 cases submitted in the court’s first six years covering nearly
every statute the legislature had passed (Sólyom and Brunner 2000, 72) were actually
of sufficient legal merit to warrant judicial review. Correcting for access in
correlations controls for this peculiarity. Thus, while the method of controlling for
this aspect requires greater sophistication, ignoring this aspect would be the greater
mistake.
60
The aim of adjusting the dependent variable—the frequency of judicial review—for access to the
courts is to control for the possibility that easier access permits not only more persons to petition the
court, and thus presumably more cases, but also a higher proportion of cases with weak legal merits.
The rationale for this expectation is complex, and can only be sketched briefly here: Assuming all
parties are rational actors, if access to a court is suddenly opened to more parties and/or in more types
of cases, the costs of petitioning the courts has effectively fallen (from infinity to a non-infinite level
for those previously unable to petition the court at all; for some others, their ability to petition the court
may have been indirect [as is common in CEE]—in that they could only request the ombudsman or an
MP to petition the court on their behalf). Lower costs of petition, ceteris paribus, mean a fall in the
cost-benefit ratio of petitioning the court – in short, a fall in the relative price of petitioning. That a fall
in ‘price’ of petitioning should be met by an increase in the number of petitions, ceteris paribus, is not
surprising. The important inference, though, is that presumably the group of ‘sub-marginal petitioners’
– for whom it was not on balance worthwhile to petition the courts (due to the cost-benefit ratio) before
access was made easier, but for whom ex post petitioning is worthwhile – would come to make
petitions with a lower average likelihood of success (since cases more likely to succeed would already
have a lower cost-benefit ratio). In short, it is most likely, on average, sub-marginal cases (in terms of
legal merits and/or likelihood of success in securing judicial review) which are added to the courts’
docket when access is made easier.
23
4. RESULTS & ANALYSIS
4.1. OVERVIEW
The results of bivariate correlations and selected regression analyses conducted (fortyeight regression models in total) are presented in Tables 4–16, in seriatim, to reflect of
each layer of corrections and expansions upon the original Smithey and Ishiyama
(2002) and Herron and Randazzo (2003) models. 61 The following sections discuss
the key findings from these analyses.
4.2. EFFECTS OF ALTERNATIVE MEASURES OF JUDICIAL INDEPENDENCE
Table 4 presents bivariate correlations of various independent variables with three
aggregate measures of judicial review. Tables 5, 6, and 7 report results of logit
regressions testing the SII (original, arithmetically fixed, and data-corrected) against
both pairs of authors’ original models.
61
All 48 logit regressions run employ a weighting procedure, developed by Herron and Randazzo
(2003) and also used by Smithey and Ishiyama (2002), to ensure that countries which issue
comparatively fewer decisions are not treated as less important by the regression analysis. A similar,
slightly modified method of weighting cases was employed here. The full technical details of Herron
and Randazzo’s weighting procedure are not detailed in their article, and were also not available on
request; however, the procedure detailed above achieves the same purpose very similarly. In this study,
the weight factor w given to each case decided by the court of country x is given by:
(Total number of cases in the dataset)
wx = ———————————————————————————————
(Total number of cases from country x) * (Number of countries in dataset)
24
TABLE 4. CORRELATION RESULTS (MODIFIED REPLICATION OF SMITHEY AND ISHIYAMA [2002])
Independent Variable
Variable
Judicial Independence
Smithey and Ishiyama Index
Original Index (fixed)
Data-corrected Index
Original Index (fixed): Absolute Value of Distance fro m
0.55
Data-corrected Index (fixed )
Absolute Value of Distance
fro m 0.55
New Measures:
Power measure
Insularity Measure
Power * Insularity
Co mposite power and access
measure
Co mposite power and access
measure * Insularity
Leg islative Frag mentation
At first election
Overall average
Popular Trust in Courts
Rights Index
Federalis m
Smithey & Ishiyama (2002)
Statistic
Incidence of
Judicial Review
Incidence of
Judicial Review Adjusted
for Access
Data for Dependent Variab le
Smithey & Ishiyama (2002)
plus Hungary and Poland
Incidence of
Incidence of
Judicial ReJudicial Review Adjusted
view
for Access
Replication
Incidence of
Incidence of
Judicial ReJudicial Review Adjusted
view
for Access
Pearson's r
Sig. (2-tailed)
N
r2
Pearson's r
Sig. (2-tailed)
N
r2
Pearson's r
Sig. (2-tailed)
N
r2
Pearson's r
Sig. (2-tailed)
N
r2
-0.117
0.802
7
0.014
0.297
0.518
7
0.088
-0.421
0.347
7
0.178
-0.655
0.110
7
0.429
-0.276
0.549
7
0.076
-0.074
0.875
7
0.005
-0.668
0.101
7
0.446
-0.907 ***
0.005
7
0.822
-0.021
0.957
9
0.000
0.147
0.706
9
0.022
-0.147
0.706
9
0.022
-0.441
0.235
9
0.194
0.225
0.561
9
0.050
0.389
0.301
9
0.151
-0.595 *
0.091
9
0.354
-0.378
0.317
9
0.143
-0.024
0.947
10
0.001
0.007
0.985
10
0.000
-0.262
0.465
10
0.068
-0.457
0.184
10
0.209
0.181
0.618
10
0.033
0.263
0.462
10
0.069
-0.418
0.229
10
0.175
-0.126
0.729
10
0.016
Pearson's r
Sig. (2-tailed)
N
r2
Pearson's r
Sig. (2-tailed)
N
r2
Pearson's r
Sig. (2-tailed)
N
r2
Pearson's r
Sig. (2-tailed)
N
r2
Pearson's r
Sig. (2-tailed)
N
r2
-0.338
0.458
7
0.114
-0.257
0.578
7
0.066
-0.289
0.530
7
0.083
-0.587
0.166
7
0.344
-0.522
0.229
7
0.273
0.272
0.556
7
0.074
-0.437
0.326
7
0.191
-0.064
0.892
7
0.004
0.183
0.695
7
0.033
0.009
0.985
7
0.000
-0.547
0.127
9
0.300
-0.147
0.705
9
0.022
-0.483
0.188
9
0.233
-0.547
0.128
9
0.299
-0.521
0.150
9
0.272
0.744 **
0.022
9
0.553
-0.107
0.785
9
0.011
0.643 *
0.062
9
0.413
0.815 ***
0.007
9
0.664
0.800 ***
0.010
9
0.639
-0.157
0.665
10
0.025
-0.256
0.475
10
0.066
-0.229
0.524
10
0.053
0.065
0.859
10
0.004
0.024
0.948
10
0.001
0.851 ***
0.002
10
0.724
-0.063
0.863
10
0.004
0.730 **
0.017
10
0.533
0.972 ****
0.000
10
0.945
0.943 ****
0.000
10
0.889
Pearson's r
Sig. (2-tailed)
N
r2
Pearson's r
Sig. (2-tailed)
N
r2
Pearson's r
Sig. (2-tailed)
N
r2
Pearson's r
Sig. (2-tailed)
N
r2
Pearson's r
Sig. (2-tailed)
N
r2
0.350
0.442
7
0.122
0.068
0.884
7
0.005
0.620
0.189
7
0.384
-0.577
0.175
7
0.333
0.033
0.944
7
0.001
0.357
0.346
9
0.127
0.155
0.691
9
0.024
0.620
0.189
9
0.384
-0.577
0.175
9
0.333
-0.170
0.661
9
0.029
-0.126
0.746
9
0.016
0.101
0.797
9
0.010
0.084
0.874
9
0.007
0.178
0.702
9
0.032
0.228
0.554
9
0.052
0.045
0.901
10
0.002
0.363
0.303
10
0.132
0.364
0.478
10
0.132
0.329
0.471
10
0.108
-0.145
0.690
10
0.021
-0.263
0.462
10
0.069
-0.016
0.965
10
0.000
-0.243
0.643
10
0.059
0.792
0.034
10
0.627
0.191
0.597
10
0.036
0.673 *
0.097
7
0.453
0.559
0.192
7
0.313
0.084
0.874
7
0.007
0.178
0.702
7
0.032
0.517
0.235
7
0.267
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001.
25
26
0.516
-0.591
(0.729)
0.124
50.00%
69.50%
39.00%
0.657
12.621
2.899
15.436
414 ___
503.334
24.675
0.000
0.080
-0.924 ****
(0.260)
5.332 *
(3.132)
0.445 ****
(0.113)
0.301
(0.856)
Model 1
Original Index
(Calculations Fixed)
B (S.E.)
Wald
-0.834
(0.966)
0.269
50.00%
69.50%
39.00%
0.745
12.684
2.875
18.255
414 ___
503.188
24.821
0.000
0.081
-0.918 ****
(0.257)
6.184 *
(3.647)
0.438 ****
(0.102)
0.598
(1.152)
Corrected Index (Coding
Fi xed)
B (S.E.)
Wald
50.00%
68.70%
37.40%
2.997
1.041
5.215
4.803
2.252
Wald
397 ___
478.000
14.968
0.005
0.052
-7.694 *
(4.444)
-0.205
(0.201)
30.214 **
(13.23)
0.099 **
(0.045)
5.100
(3.398)
B (S.E.)
Original Index
50.00%
68.70%
37.40%
397 ___
476.610
16.357
0.003
0.057
-10.702 **
(5.191)
-0.437 *
(0.240)
42.516 **
(17.30)
0.091 ***
(0.033)
8.660 *
(4.640)
4.251
3.302
6.036
7.63 6
3.48 2
Corrected Index
(Coding Fi xed)
B (S.E.)
Wald
Model 2
N.B.: Smaller sample sizes (414 and 397, respectively) result fro m unavailability of data for Smithey and Ishiyama’s synthetic rights index fo r Hungary, Poland, or Slovenia, and in
the case of Model 2, popular trust score for Georgia.
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001. DV: Dichotomous measure of judicial Rev iew (1 = legal norm/government action struck down)
50.00%
69.50%
39.00%
Null Model
Predicted Model
Reduction of Error
-0.721
(1.004)
Constant
11.490
414 ___
503.314
24.695
0.000
0.080
-0.955 ****
(0.281)
Nu mber of Sub-tiers
2.924
11.072
0.143
Wald
N
-2LL
Model Chi Square
Prob < ch i square
Pseudo R square
5.568 *
(3.256)
0.463 ****
(0.139)
0.430
(1.135)
B (S.E.)
Original Index
Rights Index
Popular Trust in Courts
Legislat ive Frag mentation
Corrected S-I Index
Index, Georgia Fixed
Judicial Independence
Original S-I Index
Independent Variable
TABLE 5. LOGIT R ESULTS : FACTORS AFFECTING J UDICIAL R EVIEW
(M ODIFIED R EPLICATION OF S MITHEY & ISHIYAMA 2002) — ORIGINAL & CORRECTED INDICES
TABLE 6. LOGIT R ESULTS : FACTORS AFFECTING J UDICIAL ACTIVISM
(M ODIFIED R EPLICATION OF HERRON AND RANDAZZO 2003) — ORIGINAL & CORRECTED I NDICES
Independent Variable
B
(S.E.)
Judicial Independence
Original S-I Index
Original Index
(Calculations Fixed)
Original Index
B
(S.E.)
Wald
-0.146
(0.374)
Wald
B
(S.E.)
Wald
0.151
-0.123
(0.365)
Index, Georgia Fixed
0.113
Corrected S-I Index
Economic Conditions
Change in GDP Growth
(base year = 1990)
Correcte d Index
(Coding Fi xed)
7.309
-0.152 ***
(0.055)
7.785
-0.084 **** 13.586
(0.023)
-0.084 ****
(0.023)
13.275
-0.092 ****
(0.024)
14.949
Legislative Fragmentation
0.193 **** 13.532
(0.052)
0.195 ****
(0.051)
14.307
0.186 ****
(0.051)
13.477
Civil Liberties
0.133
(0.108)
1.503
0.128
(0.106)
1.470
0.140
(0.103)
1.857
0.390
(0.337)
1.000
0.380
(0.338)
1.000
0.714
(0.434)
1.000
Constant
7.308
1.404
-0.147 ***
(0.054)
Conte xtual Influences
Presidential Power
-0.147 ***
(0.054)
-0.552
(0.465)
N
-2LL
Model Chi Square
Prob < ch i square
Pseudo R square
915 ___
1162.729
36.200
0.000
0.053
915 ___
1162.768
36.161
0.000
0.053
915 ___
1161.465
37.464
0.000
0.055
Null Model
Predicted Model
Reduction of Error
50.00%
64.50%
29.00%
50.00%
63.90%
27.80%
50.00%
63.90%
27.80%
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001. DV: Dichotomous measure of judic ial activis m/review (1 = legal
norm/government action struck down)
27
TABLE 7. LOGIT R ESULTS : FACTORS AFFECTING J UDICIAL R EVIEW, CONTINUED
(M ODIFIED R EPLICATION OF HERRON AND RANDAZZO 2003) — ORIGINAL & CORRECTED I NDICES
Independent Variable
Judicial Independence
Original S-I Index
Original Index
(Calculations Fixed)
B
Wald
(S.E.)
Original Index
B
(S.E.)
Wald
0.012
(0.431)
0.001
-0.168
(0.427)
Index, Georgia Fixed
0.155
Corrected S-I Index
Economic Conditions
Change in GDP Growth
(base year = 1990)
Contextual Influences
Presidential Power
Corrected Index
(Coding Fi xed)
B
Wald
(S.E.)
-0.671
(0.551)
1.482
4.219
-0.123 **
(0.061)
4.020
-0.123 **
(0.061)
4.042
-0.126**
(0.061)
-0.077 ***
(0.026)
9.175
-0.081 ***
(0.026)
9.656
-0.088****
(0.027)
11.003
17.517
0.249****
(0.061)
16.681
Legislative Fragmentation
0.262 **** 17.375
(0.063)
0.257 ****
(0.061)
Civil Liberties
0.099
(0.124)
0.644
0.113
(0.121)
0.871
0.118
(0.119)
0.991
-0.188
(0.187)
1.013
-0.195
(0.187)
1.080
-0.209
(0.187)
1.247
Petitioner
Individual
President
1.247 ***
8.162
1.236 ***
(0.437)
Administration or Legal
Official (e.g. Ombudsman)
(0.436)
0.616 **
3.958
Constant
1.218***
7.819
(0.436)
0.586 *
(0.310)
Members of Legislature
8.035
3.600
(0.309)
0.541*
3.110
(0.307)
0.226
(0.237)
0.911
0.243
(0.239)
1.035
0.255
(0.238)
1.156
-0.133
(0.397)
0.113
-0.034
(0.400)
0.007
0.370
(0.520)
0.506
N
-2LL
Model Chi Square
Prob < ch i square
Pseudo R square
764 ___
962.366
53.944
0.000
0.093
764 ___
963.708
54.099
0.000
0.093
764 ___
962.366
55.441
0.000
0.095
Null Model
Predicted Model
Reduction of Error
50.00%
64.90%
29.80%
50.00%
64.40%
28.80%
50.00%
63.90%
27.80%
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001.DV: Dichotomous measure of judicial Review (1 = legal norm/
government action struck down)
N.B. Smaller sample size results fro m missing data on petitioners in some cases in the dataset. Weights have been adjusted.
28
The results concerning the influence of judicial independence on judicial
review are indeed mixed, but a number of important findings deserve mention. First,
though the original SII is clearly a poor predictor of judicial review, substantive
corrections in coding improve it noticeably. Correlations yield weak negative results,
and no significant results were obtained by regression, 62 even when technical and
arithmetical errors are corrected. However, the data-corrected version of the SII
reflects a small but important improvement in the SII’s predictive power: in
correlation analysis it is positively correlated with judicial review (e.g. r = 0.297), a
relation which is stronger when judicial review is adjusted for accessibility of the
courts (r = 0.389 using original authors’ data plus Hungary and Poland; r = .263
using the new replication dataset). It also yields a positive, statistically significant
relationship in one of four regressions where it is tested, and in all others yields no
significant relationship.
Second, the casual observation by Smithey and Ishiyama (2002)—that judicial
review increases up to around the midpoint of the judicial independence scale before
declining—is confirmed. Tables 8 and 9 present results from models testing the
‘peak’ observation, but otherwise still following the authors’ original template. Using
Smithey and Ishiyama’s (2002) own judicial review statistics, the correlation between
judicial review and absolute value of distance from 0.55 on the judicial independence
SII is -0.668 and -0.907 for the original and data-corrected indices, respectively. This
negative relationship is also found, in regressions based on Smithey and Ishiyama’s
(2002) models (where the relationship is statistically significant in every test), though
not those based on Herron and Randazzo’s models.
62
In all statistical procedures the α value is .10, as used by Herron and Randazzo (2003).
29
30
50.00%
69.50%
39.00%
Null Model
Predicted Model
Reduction of Error
0.121
0.572
(0.639)
50.00%
69.50%
39.00%
0.801
16.803
414 ___
499.366
28.644
0.000
0.093
0.411 ****
(0.100)
2.291
13.589
0.831
3.395
1.855
3.555
Wald
-0.629
(0.604)
50.00%
68.70%
37.40%
397 ___
476.488
26.479
0.002
0.057
1.084
0.034 **** 11.582
(0.010)
12.478 **** 11.177
(3.732)
-0.281
(0.206)
-3.524 *
(1.869)
B
(S.E.)
Distance fro m Peak
-0.095
(0.787)
50.00%
68.70%
37.40%
397 ___
475.350
17.617
0.003
0.061
0.032 ***
(0.010)
10.902 ***
(4.017)
-0.282
(0.206)
-2.362
(2.209)
-3.740 **
(1.887)
0.014
9.704
7.366
1.875
1.143
3.927
Distance fro m Peak,
Controlling for Access
B
Wald
(S.E.)
Model 2
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001. DV: Dichotomous measure of judicial Rev iew (1 = legal norm/government action
struck down)
N.B.: Smaller sample sizes (414 and 397, respectively) result fro m unavailability of data for Smithey and Ishiyama’s synthetic rights index fo r
Hungary, Poland, or Slovenia, and in the case of Model 2, popular trust score for Georg ia.
414 ___
500.192
27.817
0.000
0.090
0.155
(0.444)
Constant
18.363
N
-2LL
Model Chi Square
Prob < ch i square
Pseudo R square
0.425 ****
(0.099)
Legislat ive Frag mentation
Popular Trust in Courts
4.935
(3.260)
5.896 *
(3.070)
Rights Index
3.689
-0.955 ****
(0.259)
14.010
-0.968 ****
(0.259)
Nu mber of Sub-tiers
-3.632 *
(1.971)
-2.116
(2.321)
3.059
Accessibility of Courts
Distance fro m 'peak' of Judi- -3.402 *
cial Independence
(1.945)
Model 1
Distance fro m Peak,
Distance fro m Peak
Controlling for Access
B
B
Wald
Wald
(S.E.)
(S.E.)
TABLE 8. LOGIT R ESULTS : EFFECT OF D ISTANCE FROM ‘PEAK’ OF J UDICIAL POWER ON J UDICIAL REVIEW
(FOLLOWING S MITHEY AND ISHIYAMA ’S [2002] M ODELS )
TABLE 9. LOGIT R ESULTS : EFFECT OF D ISTANCE FROM ‘PEAK’ OF J UDICIAL POWER ON
JUDICIAL R EVIEW (FOLLOWING H ERRON AND RANDAZZO ’S [2003] M ODELS )
Without Controlling
for Pet itioner
B
Wald
Distance fro m 'peak' of Judicial Independence
0.352
(0.539)
0.427
Contextual Influences
Legislative Fragmenta0.202 ****
tion
(0.050)
Presidential Power
0.144
(0.109)
Civil Liberties
-0.090 ****
(0.025)
16.064
1.750
Controlling for Petitioner
B
Wald
-0.217
(0.621)
0.122
0.260 ****
(0.061)
18.176
0.085
(0.124)
0.473
12.496
-0.073 ***
(0.028)
6.639
7.422
-0.123 **
(0.061)
3.979
-0.195
(0.188)
1.082
President
1.254 ***
(0.436)
8.255
Administration or Legal
Official
0.639 **
(0.310)
4.245
Members of Legislature
0.225
(0.235)
0.913
-0.090
(0.340)
0.071
Economic Conditions
Change in GDP Growth -0.148 ***
(base year = 1990)
(0.054)
Petitioner
Individual
Constant
0.251
(0.285)
0.776
N
-2LL
Model Chi Square
Prob < ch i square
Pseudo R square
915 ___
1162.453
36.477
0.000
0.054
764 ___
963.742
54.065
0.000
0.093
Null Model
Predicted Model
Reduction of Error
50.00%
64.50%
29.00%
50.00%
65.00%
30.00%
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001. DV: Dichotomous measure of judicial Review (1 = legal norm/government action struck down)
N.B. Smaller sample size results fro m missing data on petitioners in some cases in the
dataset. Weights have been adjusted.
31
Though Smithey and Ishiyama do not offer an explanation for this, at least one
possibility deserves mention. 63 As both Vanberg (2001; 2005) and Stone Sweet
(2000) suggest, more powerful courts can influence policymaking without striking
down statutes, or indeed without hearing a case, if the legislature (or other branch or
body) anticipates that passing a bill which the court might unconstitutional will result
in it being struck down by the court. In short, “the spectre of constitutional censure
hovers over the legislative process” (Stone Sweet 2000, 196). This is true only to the
extent that others anticipate court intervention and only to the extent that the court
possesses the authority to do so. Thus, the degree of the court’s formal authority
determines, as it were, the length of the shadow that the court casts over the
legislature process. If this is true, it may in part explain the finding of the ‘peak’ in
judicial review: the more power courts have, the less they need to exercise it, as
statutes and other legal norms which they are likely to strike down are never passed in
the first place.
Tables 10, 11, 12, and 13, report findings regarding new measures of judicial
independence, tested in the framework of the prior authors’ original models.
Variables measuring countries’ receptiveness to legal transplants are included in
Tables 14 and 15. Finally, Table 16 presents three composite models which
incorporate components from both the Smithey and Ishiyama (2002) or Herron and
Randazzo (2003) template.
63
A second explanation may also account for this, which parallels the argument for adjusting the
dependent variable for ease of access to the courts: the more powerful a court, the greater the demand
for its services by individual appellants, and thus the greater its caseload and potentially the greater the
proportion of cases with weak legal merit.
32
33
50.00%
67.30%
34.60%
Null Model
Predicted Model
Reduction of Error
3.050
0.198
2.211
50.00%
69.50%
39.00%
414 ___
500.175
27.835
0.000
0.090
1.855
(2.091)
-0.538
(0.932)
57.588 *
(34.391)
-11.622
(10.281)
0.334
0.787
1.278
2.804
50.00%
69.50%
39.00%
414 ___
500.165
27.844
0.000
0.090
-0.825
(1.062)
-0.047
(0.905)
19.534 *
(11.203)
(0.122)
-10.231
(9.209)
-0.591 *
(0.306)
0.003
0.604
3.040
3.724
1.234
3
B
Wald
(S.E.)
0.388 *** 10.103
50.00%
67.30%
34.60%
414 ___
501.527
26.482
0.000
0.086
-0.687 *
(0.387)
15.001
(9.314)
(0.100)
-6.788
(7.911)
-0.652 **
(0.295)
3.140
2.594
4.894
0.736
4
B
Wald
(S.E.)
0.444 **** 19.551
50.00%
69.50%
39.00%
414 ___
501.934
26.076
0.000
0.085
-0.786
(0.480)
10.998
(8.942)
(0.109)
0.920
(4.557)
-0.777 ***
(0.275)
2.677
1.513
7.985
0.041
5
B
Wald
(S.E.)
0.477 **** 19.117
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001. DV: Dichotomous measure of judicial Review (1 = legal norm/government action
struck down)
414 ___
500.989
27.020
0.000
0.088
-1.328 *
(0.761)
1.443
(3.243)
8.999
(6.053)
(0.138)
-8.913
1.103
(8.487)
-1.166 **** 14.458
(0.307)
(0.142)
-4.886
0.468
(7.139)
-0.801 *** 8.621
(0.273)
7.219
Wald
2
B
(S.E.)
0.371 ***
1
B
Wald
(S.E.)
0.368 *** 6.739
N
-2LL
Model Chi Square
Prob < ch i square
Pseudo R square
Constant
Insularity
Co mposite power and
access measure
Co mposite power and
access measure * Insularity
Power * Insularity
Power * Access
Judicial Independence
Access Measure
Federalism
Rights Index
Legislat ive Frag mentation
Independent Variable
TABLE 10 LOGIT RESULTS: EFFECT OF NEW MEASURES OF JUDICIAL INDEPENDENCE ON
JUDICIAL REVIEW FOLLOWING SMITHEY AND ISHIYAMA’S (2002) ‘MODEL 1’
34
50.00%
68.70%
37.40%
Null Model
Predicted Model
Reduction of Error
0.149
2.840
0.878
0.389
2.048
1.008
Wald
50.00%
68.70%
37.40%
397 ___
476.415
16.552
0.005
0.057
1.129
(2.929)
-10.636
(11.349)
21.320 *
(12.652)
2
B
(S.E.)
-20.765
(20.685)
-0.399
(0.279)
-0.038
(0.061)
0.149
2.840
0.878
0.389
2.048
1.008
Wald
3.525
(13.754)
24.116
(17.052)
-3.860
(9.910)
B
(S.E.)
-17.336
(40.915)
-0.130
(0.658)
-0.011
(0.132)
50.00%
68.70%
37.40%
397 ___
476.415
16.552
0.005
0.057
3
0.066
0.152
2.000
0.007
0.039
0.180
Wald
50.00%
68.70%
37.40%
397 ___
476.567
16.401
0.003
0.057
-1.825 ***
(0.688)
18.756 *
(10.037)
7.030
3.492
4
B
Wald
(S.E.)
-1.684
0.048
(7.667)
0.106
0.175
(0.254)
0.040 **** 13.697
(0.011)
50.00%
68.70%
37.40%
397 ___
476.427
16.540
0.002
0.057
-3.079 **
(1.196)
25.717 *
(13.544)
6.628
3.605
5
B
Wald
(S.E.)
7.460 *
3.373
(4.062)
0.179
0.419
(0.277)
0.057 **** 12.073
(0.017)
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001. DV: Dichotomous measure of judicial Rev iew (1 = legal norm/government action struck
down)
397 ___
476.415
16.552
0.005
0.057
1.129
(2.929)
-10.636
(11.349)
21.320 *
(12.652)
1
B
(S.E.)
-20.765
(20.685)
-0.399
(0.279)
-0.038
(0.061)
N
-2LL
Model Chi Square
Prob < ch i square
Pseudo R square
Constant
Co mposite power and
access measure * Insularity
Insularity
Co mposite power and
access measure
Access Measure
Judicial Independence
Power * Insularity
Popular Trust in Courts
Federalism
Rights Index
Independent Variable
TABLE 11 LOGIT RESULTS: EFFECT OF NEW MEASURES OF JUDICIAL INDEPENDENCE ON
JUDICIAL REVIEW FOLLOWING SMITHEY AND ISHIYAMA’S (2002) ‘MODEL 2’
35
50.00%
63.90%
27.80%
Null Model
Predicted Model
Reduction of Error
1.677
0.371
0.631
50.00%
65.10%
30.20%
915 ___
1160.982
37.948
0.000
0.056
0.918
(0.943)
0.142
(0.534)
-5.397
(3.988)
2.582
(2.028)
0.071
0.947
1.622
1.832
2
B
Wald
(S.E.)
-0.082 **** 13.375
(0.022)
0.179
2.595
(0.111)
-0.134 **
5.860
(0.055)
0.182 *** 10.207
(0.057)
50.00%
63.90%
27.80%
915 ___
1162.817
36.112
0.000
0.053
-0.131
(0.531)
0.401
(0.499)
0.032
(0.643)
0.647
0.061
0.002
3
B
Wald
(S.E.)
-0.083 **** 13.624
(0.022)
0.123
1.426
(0.103)
-0.146 ***
7.222
(0.054)
0.192 **** 11.344
(0.057)
50.00%
64.50%
29.00%
915 ___
1162.880
36.050
0.000
0.053
0.303
(0.304)
0.036
(0.642)
0.999
0.003
4
B
Wald
(S.E.)
-0.082 **** 13.830
(0.022)
0.117
1.367
(0.100)
-0.146 ***
7.197
(0.054)
0.199 **** 15.715
(0.050)
50.00%
64.50%
29.00%
915 ___
1162.881
36.049
0.000
0.053
0.317
(0.313)
-0.042
(1.150)
1.027
0.001
5
B
Wald
(S.E.)
-0.082 **** 13.714
(0.022)
0.116
1.356
(0.100)
-0.146 ***
7.214
(0.054)
0.199 **** 15.403
(0.051)
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001. DV: Dichotomous measure of judicial Rev iew (1 = legal norm/government action struck
down)
915 ___
1162.238
36.691
0.000
0.054
0.441
(0.341)
0.407
(0.668)
-1.736
(2.186)
1
B
Wald
(S.E.)
-0.083 **** 14.124
(0.022)
0.146
1.842
(0.108)
-0.143 ***
6.896
(0.055)
0.174 ***
8.897
(0.058)
N
-2LL
Model Chi Square
Prob < ch i square
Pseudo R square
Constant
Insularity
Co mposite power and access measure
Co mposite power and access measure * Insularity
Access Measure
Power * Access
Judicial Independence
Power * Insularity
Change in GDP Growth
(1990 = base year)
Legislat ive Frag mentation
Civil Lliberties
Presidential Power
Independent Variable
TABLE 12. LOGIT RESULTS: EFFECT OF NEW MEASURES OF JUDICIAL INDEPENDENCE ON JUDICIAL
REVIEW FOLLOWING MODIFIED HERRON AND RANDAZZO (2003) MODEL (WITHOUT CONTROLS FOR
PETITIONER IDENTITY)
TABLE 13. LOGIT RESULTS: EFFECT OF NEW MEASURES OF JUDICIAL INDEPENDENCE ON JUDICIAL
REVIEW FOLLOWING MODIFIED HERRON AND RANDAZZO (2003) MODEL (WITH CONTROLS FOR
PETITIONER IDENTITY)
Independent Variable
Presidential Power
Civil Lliberties
Change in GDP Growth (1990 =
base year)
Legislat ive Frag mentation
1
B
Wald
(S.E.)
-0.083 *** 10.560
(0.026)
0.191
2.270
(0.127)
-0.112 *
3.348
(0.061)
0.191 *** 8.309
(0.066)
Petitioner
Individual
-0.196
(0.194)
President
0.594 *
(0.303)
Administration or Legal Official 0.357
(0.246)
Members of Legislature
1.258 ***
1.024
3.828
2.106
8.206
(0.439)
Judicial Independence
Power * Insularity
-5.442 **
(2.586)
2
B
Wald
(S.E.)
-0.085 *** 10.771
(0.026)
0.233 *
3.170
(0.131)
-0.105 *
2.898
(0.062)
0.216 *** 10.805
(0.066)
3
B
(S.E.)
-0.082 ***
(0.025)
0.117
(0.119)
-0.122 **
-0.114
(0.199)
0.705 **
(0.315)
0.323
(0.247)
1.325 ***
-0.214
(0.196)
0.568 *
(0.305)
0.289
(0.245)
1.223 ***
0.328
5.010
1.713
8.947
(0.443)
4.428
Power * Access
-10.591 **
(4.644)
4.903 **
Wald
10.349
0.961
3.951
(0.061)
0.235 **** 12.620
(0.066)
1.194
3.461
1.389
7.797
(0.438)
4
B
(S.E.)
-0.078 ***
(0.025)
0.100
(0.117)
-0.123 **
Wald
9.780
0.723
4.022
(0.061)
0.261 **** 18.200
(0.061)
-0.185
(0.194)
0.613 **
(0.302)
0.227
(0.235)
1.249 ***
0.913
4.114
0.927
8.158
(0.437)
5
B
(S.E.)
-0.079 ***
(0.025)
0.098
(0.117)
-0.123 **
Wald
9.835
0.707
4.028
(0.061)
0.259 **** 17.506
(0.062)
-0.177
(0.192)
0.607 **
(0.303)
0.227
(0.235)
1.255 ***
0.842
4.028
0.928
8.246
(0.437)
5.201
4.458
(2.322)
Access Measure
1.399 *
(0.809)
2.989
Co mposite power and access
measure
-0.012
(0.749)
0.000
-0.043
(0.748)
0.003
Co mposite power and access
measure * Insularity
Insularity
Constant
0.196
(0.393)
0.249
1.519
(1.140)
-0.259
(0.641)
1.775
0.163
-0.614
(0.651)
0.321
(0.587)
-0.318
(1.329)
0.057
-0.083
(0.371)
0.051
0.890
0.299
-0.118
(0.361)
0.106
N
-2LL
Model Chi Square
Prob < ch i square
Pseudo R square
764 ___
959.401
58.406
0.000
0.100
764 ___
957.678
60.129
0.000
0.103
764 ___
962.969
54.837
0.000
0.094
764 ___
963.860
53.947
0.000
0.093
764 ___
963.806
54.000
0.000
0.093
Null Model
Predicted Model
Reduction of Error
50.00%
64.90%
29.80%
50.00%
65.10%
30.20%
50.00%
64.90%
29.80%
50.00%
65.10%
30.20%
50.00%
65.10%
30.20%
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001. DV: Dichotomous measure of judicial Rev iew (1 = legal norm/government action struck
down)
36
37
1
50.00%
65.93%
31.86%
Null Model
Predicted Model
Reduction of Error
0.010
5.833
2.580
6.897
6.110
0.289
9.605
2.230
0.871
Wald
2
50.00%
64.43%
28.85%
1.284
2.488
7.788
6.187
0.320
13.027
2.722
0.977
Wald
915 ___
1153.229
45.701
0.000
0.067
0.609 ***
(0.218)
0.563
(0.357)
0.509
(0.449)
-0.503
(0.509)
-0.047 *
(0.028)
0.193 ****
(0.053)
-0.116
(0.205)
-0.141 **
(0.057)
B
(S.E.)
3
50.00%
64.73%
29.45%
915 ___
1159.167
39.762
0.000
0.058
1.252 **
(0.525)
-0.364
(0.524)
0.644 ***
(0.221)
0.909 *
(0.503)
-0.876
(0.880)
-0.032
(0.029)
0.151 ***
(0.056)
-0.037
(0.201)
B
(S.E.)
0.483
5.688
3.265
8.470
0.033
7.410
1.253
0.989
Wald
4
50.00%
65.39%
30.78%
0.189
3.755
1.714
10.297
2.119
0.048
8.842
0.000
1.333
Wald
915 ___
1157.037
41.892
0.000
0.061
1.047 *
(0.540)
-0.231
(0.531)
-0.306
(0.210)
0.861 ***
(0.268)
0.691
(0.528)
-1.030
(0.892)
0.000
(0.036)
0.206 ***
(0.069)
-0.044
(0.200)
B
(S.E.)
5
50.00%
61.89%
23.78%
1.559
4.920
2.614
9.450
0.229
13.249
1.267
Wald
915 ___
1160.155
38.774
0.000
0.057
1.130 **
(0.509)
-0.590
(0.473)
0.674 ***
(0.219)
0.529
(0.327)
-0.032
(0.028)
0.179 ****
(0.049)
-0.092
(0.192)
B
(S.E.)
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001. DV: Dichotomous measure of judicial Rev iew (1 = legal norm/government action struck
down)
915 ___
1153.029
45.900
0.000
0.067
1.273 **
(0.527)
-0.053
(0.546)
0.587 ***
(0.224)
0.813
(0.506)
-0.828
(0.887)
-0.044
(0.029)
0.181 ***
(0.059)
-0.110
(0.205)
-0.140 **
(0.057)
B
(S.E.)
N
-2LL
Model Chi Square
Prob < ch i square
Pseudo R square
Constant
New Transplant
Receptive Transplant
Familiarity
Adaptation
Change in GDP Growth
(1990 = base year)
Federalism
Civil Liberties
Legislat ive Frag mentation
Presidential Power
Judicial Independence
Corrected Index
Independent Variable
TABLE 14. INFLUENCE OF LEGAL “TRANSPLANT EFFECT” ON JUDICIAL REVIEW (PANEL A)
TABLE 15. INFLUENCE OF LEGAL “TRANSPLANT EFFECT” ON JUDICIAL
REVIEW (PANEL B)
6
Independent Variable
B
(S.E.)
7
Wald
B
(S.E.)
8
Wald
B
(S.E.)
Wald
Judicial Independence
Po wer measure
Insularity measure
Po wer * Insularity
Access measure
Presidential Power
Legislat ive Frag mentation
At time of decision
21.457 ***
(8.181)
7.220 ***
6.879
8.311
(2.505)
-36.925 ***
(13.170)
-1.430
(1.516)
-0.066 *
(0.034)
19.035 **
(8.276)
9.419 ****
5.291
11.829
(2.739)
-40.481 ***
(13.364)
0.889
0.966
(1.926)
3.731
-0.082 **
(0.036)
7.861
0.163 ***
(0.059)
7.553
0.225 ***
(0.070)
9.176
0.252
5.181
10.342
0.126 **
(0.051)
At time of first election
Civil Liberties
-0.074
(0.203)
0.135
0.035
(0.215)
0.026
Federalism
1.333 ****
(0.315)
0.283
(0.558)
Adaptation
Familiarity
Receptive Transplant
New Transplant
Constant
1.216 ****
(0.311)
0.868 **
(0.352)
-3.738 ***
(1.391)
6.120
-0.242
(0.154)
17.852 0.902 ****
(0.214)
0.257 0.625 *
(0.327)
2.471
17.770
3.647
15.245
6.074
7.226
1.468 **
(0.640)
-5.633 ****
(1.612)
5.255
1.090 ***
(0.336)
-0.770 *
(0.435)
10.548
N
-2LL
Model Chi Square
Prob < ch i square
Pseudo R square
915 ___
1149.981
48.949
0.000
0.071
915 ___
1145.895
53.034
0.000
0.077
915 ___
1177.614
25.948
0.000
0.038
Null Model
Predicted Model
Reduction of Error
50.00%
64.72%
29.44%
50.00%
63.46%
26.92%
50.00%
61.80%
23.60%
3.128
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001. DV: Dichotomous measure of judicial Rev iew
(1 = legal norm/government action struck down)
38
TABLE 16. SELECTED COMPOSITE MODELS OF FACTORS AFFECTING
JUDICIAL REVIEW
1
B
Wald
(S.E.)
Change in GDP Growth
-0.118 **
4.403
(1990 = base year)
(0.056)
Presidential Power
-0.257 **** 20.430
(0.057)
Civil Lliberties
0.436 ***
6.652
(0.169)
Legislati ve Fragmentati on
At time of decision
0.173
2.309
(0.114)
Inside Coalition
-0.693 **** 12.827
(0.193)
Federalism
0.556 ***
7.385
(0.205)
Judicial Independence
Co mposite power and
-4.256 **
6.499
access measure * Insu(1.669)
larity
Power * Insularity
Independent Variable
2
B
Wald
(S.E.)
-0.098 *
3.222
(0.055)
-0.273 **** 24.448
(0.055)
0.368 **
5.318
(0.160)
3
B
Wald
(S.E.)
-0.105 *
3.319
(0.057)
-0.353 **** 11.374
(0.105)
0.641 ***
7.037
(0.242)
-0.628 **** 11.318
(0.187)
0.540 ***
7.186
(0.201)
0.198 *
(0.118)
-0.602 ***
(0.225)
1.332 **
(0.658)
-4.774 ***
9.543
4.103
(1.624)
Insularity
3.061 ***
(0.991)
7.177
8.644
Power * Access
Constant
2.786
3.838 **** 20.574
(0.846)
-8.002
(7.557)
0.026
(3.192)
4.164
(3.518)
0.161
(2.457)
N
-2LL
Model Chi Square
Prob < ch i square
Pseudo R square
915 ___
913.872
39.603
0.000
0.072
915 ___
916.371
37.103
0.000
0.068
915 ___
912.546
40.929
0.000
0.075
Null Model
Predicted Model
Reduction of Error
50.00%
66.50%
33.00%
50.00%
64.60%
29.20%
50.00%
66.20%
32.40%
1.121
0.000
1.401
0.004
α = .10; * p < .10; ** p < .05; *** p < .01; **** p < .001. DV: Dichotomous measure of judicial Review (1 = legal norm/government action struck down)
39
As is evident, the new measures of the dimensions of judicial independence
developed above do reflect some improvement over the SII, but the results are not
consistent. On the one hand, the separate and combined measures of Power and
Access (and the Composite power and access term) yield several positive, statistically
significant relationships. On the other hand, Insularity and its interactive term with
Power both generate some positive and some negative statistically significant results.
Moreover, no statistically significant relationships at all were evident in regressions
based on Herron and Randazzo’s (2003) models.
Taken together, these findings thus warrant caution making generalizations
about the relationships between judicial independence and judicial review. The
central hypothesis, that judicial independence is positively associated with judicial
review in CEE and the FSU, is confirmed in part and rejected in part. However, the
alternative formulation of this hypothesis, that judicial power is positively related to
judicial review (as the prior authors used the same SII to reflect both independence
and power on different occasions), is confirmed. At the very least, the findings
demonstrate that excluding independence from the judicial politics puzzle in transition
countries is premature. Better conceptualisations and operationalisations are certainly
needed, but writing judicial independence out of the equation in CEE and FSU would
be mistaken.
4.3. EFFECTS OF OTHER REPLICATION VARIABLES
Though the prior authors differed in their expectations of the effect of legislative
fragmentation, the replication findings appear dispositive of that question. In all but
two models in which it is included (35 of 37 regressions), legislative fragmentation (at
the time of the decision) is positively related to judicial review, statistically significant
at or below the .01 level.
The primary hypothesis concerning legislative
fragmentation is clearly confirmed, affirming the hypotheses findings of Smithey and
Ishiyama (2002) but rejecting the hypotheses of Herron and Randazzo (2003), who
expected a negative relationship but found no relationship.
As noted earlier, however, Herron and Randazzo’s expectation may make
sense if applied to fragmentation inside governing coalitions. This is confirmed by
the evidence. Party fragmentation is negatively associated with judicial review to a
statistically significant degree in every model where it appears. This confirms the
secondary hypothesis regarding legislative fragmentation inside coalitions (which
neither previous work had addressed).
Findings for presidential power clearly affirm the findings of Herron and
Randazzo (2003). In the overwhelming majority of models where it is included,
presidential power is negatively associated with judicial review to a highly
statistically significant degree. This makes intuitive sense for reasons described
above: a powerful executive is in a better position to constrain the courts (by action
or by explicit or implicit threat).
Findings for the degree of federalism, however, appear counterintuitive at first
blush: in almost every regression, the number of elected sub-tiers of government is
negatively correlated to judicial review, to a statistically significant degree, which is
directly contrary to Smithey and Ishiyama’s (2002) findings. While the negative and
significant relationship is surprising, part of the problem may lie in the original
authors confusing direct with indirect effects of federalism. While it is reasonable to
expect a higher number of central, regional, and local layers of authority to generate
40
jurisdictional conflicts yielding increased judicial involvement, it need not result in
increased level of judicial review.
Smithey and Ishiyama’s (2002) finding concerning popular trust in the courts
is confirmed, as in most models where it appears it is positively associated with
judicial review, to a statistically significant degree. The authors’ own index of written
rights guarantees is also, in most models, positively and significantly related to
judicial review. 64 The measure used for rights guarantees by Herron and Randazzo
(2003), on the other hand, generated conflicting results. As higher degrees of civil
liberties are reflected by lower scores on the index used, a negative relationship was
expected, but both positive and negative relationships appeared at statistically
significant levels in different models. Herron and Randazzo themselves were likewise
unable to find a clear-cut relationship.
The final attribute included primarily for replication purposes was the identity
of the petitioner requesting that the court engage in judicial review in each case.
Herron and Randazzo (2003) included the categories of individual citizens and the
president, and the replication study coded these as well as legislative petitioners
(Herron and Randazzo also coded cases where the legislature was the respondent) and
where the petition was made by a cabinet member or administrative officer, such as
an ombudsman. Cases where the president or a cabinet or administrative official
made the appeal were positively and significantly associated with judicial review in
almost every model where they appeared. Additionally, where the categories of
individual petitioner and legislative petitioner were related significantly to judicial
review, the relationship was positive, confirming Herron and Randazzo’s hypotheses
and affirming their findings.
4.4. EFFECTS OF NEW VARIABLES
Replication also provided the opportunity to test additional causal factors influencing
judicial review, namely those concerning the receptiveness of countries to legal
transplantation. The present study set out five hypotheses, generally grounded in the
existing literature on the ‘demand for law’, and the empirical data confirms all five.
First, it was hypothesised that in countries which showed a tendency for
adapting the entire legal system (i.e. not just the constitutional court) which they had
imported, the constitutional court the country imported would also function more
‘normally’ (i.e. resemble the model in the ‘exporting’ state more directly), meaning a
higher level of judicial review than those which did not show a pattern of adaptation.
Empirical findings clearly support this hypothesis: in all six regressions in which it is
included, Adaptation (using data from Pistor, Raiser, and Gelfer 2000) shows a
positive relationship to judicial review (statistically significant to at least the 0.01
level).
It was also hypothesised that a country’s familiarity with the general legal
system it imported during the transition period would be positive related to judicial
review. The results here are less resounding, but still confirm the expectation:
statistically significant results appear in two of six regressions, and in both of these
64
Their data is taken from a variety of popular opinion surveys by the University of Strathclyde;
however, direct access to this data was unavailable for replication, thus Smithey and Ishiyama’s (2002)
reported figures had to be relied upon. Yet again, these figures exclude Hungary, Poland, and
Slovenia, as well as Georgia, warranting considerable caution (as the sample size is more than halved
when these countries are excluded, and the representativeness of the panel is greatly reduced).
41
cases the relationship is positive. In light of these first two, the results of testing the
third hypothesis are therefore unsurprising. It was expected that receptiveness –
meaning the presence of both adaptation and familiarity – would be positively related
to judicial review. In both models where ‘receptiveness’ measured in this way is
included, a highly statistically significant, positive relationship is evident.
The fourth hypothesis departed somewhat from the literature. It stated the
expectation that countries which had no exposure at all to the “modern formal legal
order” (Pistor, Raiser, and Gelfer 2000, 15) would also experience a higher level of
judicial review. In each of eight models where it is included, the dummy variable for
‘new transplants’ is positive, and in seven of these the results are statistically
significant, indicating that new transplants exhibit a higher level of judicial review
than the reference category, ‘unreceptive’ transplants. Thus, the fourth hypothesis is
also confirmed.
Finally, it was hypothesised that the greater the diversity of participation in the
constitutional design and negotiation process, the greater the legitimacy of the court,
and thus the greater the support for an active, powerful tribunal. Thus, it was
expected that more varied participation at the constitutional design stage (measured
by legislatively fragmentation at the first democratic election 65 ) would be associated
with higher levels of judicial review. This hypothesis is also confirmed by the
empirical finding that the only model which includes this measure of participation
yields a positive, statistically significant relationship between participation (i.e.
legislative fragmentation at first election) and judicial review). It is also worth noting
that in the models testing these new independent variables of receptiveness, the most
important attributes of political context detailed above (viz. presidential power,
legislative fragmentation at most recent election) return the expected results (negative
and positive, respectively).
4.5. EFFECTS OF CONTROL VARIABLES
While change in GDP growth was included in Herron and Randazzo’s model as a
control variable, they make a number of arguments concerning its potential
substantive import. For instance, following Tate and Vallinder (1995), that in periods
of poor or declining economic performance,
if citizens and corporations are dissatisfied with the ability of the
political branches of government to efficiently regulate the economy,
then they may turn to legal remedies in order to advance their own
self-interests (Herron and Randazzo 2003, 426).
This argument is not totally implausible, but appears to explain rather increases or
decreases in the public’s resorting to the courts, rather than judicial review.
While access to the courts was included as a component of the new measures
of judicial independence, it was also incorporated as a kind of dependent-variable
control in a second set of bivariate correlations. The purpose of this was to account
for the possibility that courts where the right to petition was widely diffused would as
65
Complete and accurate data on the composition of various transition countries’ Round Table
negotiations was not available, thus a proxy measure was necessary; note, however, that Smithey and
Ishiyama (2000) use this same indicator to measure the same concept.
42
a result receive far more cases, a larger proportion of which would have lower legal
merit and thus not warrant the exercise of judicial review. 66 Adjusting the dependent
variable (overall average of judicial review incidence) for access to the courts yielded
much stronger correlations (e.g. r = 0.389 vs. 0.147; r = 0.263 vs. 0.007) with the
data-corrected SII than non-adjusted measures. More strikingly, several dimensions
of the new measures of judicial independence (Power; Composite power and access;
Composite power and access * Insularity) switch from strong negative correlations (r
= -0.547, -0.547, and -0.521, respectively) to strong positive correlations with judicial
review (using original Smithey and Ishiyama data plus Hungary and Poland, r =
0.744, 0.815, and 0.800, respectively; using replication dataset, r = 0.851, 0.972, and
0.943, respectively). 67 Again, more rigorous and sophisticated methods of adjusting
the dependent variable properly are still needed, but the elementary method employed
here has yielded remarkable results. 68
5. CONCLUSION
5.1. METHODOLOGY
As the foregoing study is designed partly to improve upon methodology and
operationalisation of key variables, one of its most important findings is that the
changes incorporated above to improve the execution, data-accuracy, and
conceptualisation of judicial independence do make a difference. Technical and
substantive corrections to the original SII both (1) yield small but still noteworthy
improvements in the predictive power of judicial independence and also (2) bring the
findings more in line with Smithey and Ishiyama’s (2002), Herron and Randazzo’s
(2003), and this study’s hypotheses.
Moreover, the new, theoretically-grounded measures capturing several
dimensions of judicial independence also mark a methodological advance. A number
of statistically significant results, which are supported by the literature, indicate that
the measures are at least a step in the right direction. Specifically, the new measures
are designed to assess separately the different dimensions of judicial independence
(before combining them via interactive terms), but which the prior measure does not.
5.2. SUBSTANTIVE FINDINGS
Beyond these improvements in the study of judicial independence, the foregoing
investigation yields a number of intriguing findings. The first hypothesis, that judicial
independence positively correlates with judicial review, was partially confirmed and
partially rejected. The Insularity dimension of judicial independence appears
66
See the Note to Table 4, infra.
67
While it may appear that the including measures of access in both of the variables correlated is
unsound, the computation of the measure of access incorporated into the Composite measure renders it
quite different (in algebraic form and in outcome) from the single scalar measure used to adjust judicial
review figures.
68
For instance, various non-logit regression models were tested where the dependent variable (viz.
decision outcome) was adjusted by access, but the resulting synthetic dependent variable was too
highly leptokurtic to remain within regression assumptions.
43
negatively related to judicial review, while the Power and Access of the court are
both positively related. The second form of this hypothesis, that judicial power
correlates with judicial review, is thus confirmed. Additionally, a central ‘peak’ does
exist in the level of judicial review vis-à-vis the judicial independence distribution, for
which there are several possible explanations.
Additionally, the study answers conclusively at least one question left in doubt
by the prior research and sheds at least some light on others. Overall legislative
fragmentation at the time a decision is issued by a court is positively related to
judicial review in the transition states studied. Fragmentation within the governing
coalition, however, is negatively related to judicial review. These findings support
the replication study’s hypotheses, and make sense out of the disparate expectations
and findings of Herron and Randazzo (2003) and Smithey and Ishiyama (2002).
The study further confirms several secondary findings of the prior works:
presidential power is negatively related to judicial review, while popular trust in the
courts, extent of rights guarantees in the bill of rights, and popular trust in the courts
are all positively associated with the frequency of judicial review. Likewise, cases
where the president or an administration or cabinet official is the petitioner are more
likely to result in the striking down of the law named in the petition. Finally, the
empirical data reveals that factors which affect the success of legal transplantations
broadly speaking also appear to influence the functionality of the newly imported
constitutional courts.
All of these findings, taken together, present a clearer and more complete
picture of the contextual influences on judicial review than the prior works. Both the
holistic portrait of the determinants of judicial review in transition and the new
operationalisations of judicial independence offered here are merely first steps. While
many scholars lament (e.g. Russell 2001; Herron and Randazzo 2003; Ramseyer and
Rasmussen 2003) the dearth of theoretically sound, widely applicable conceptions and
measures of judicial independence, for all its faults, Smithey and Ishiyama’s (2000)
work represents one of the few which actually sets out to fill the gap.
More importantly, the integration of better conceptualisations of independence
with measures of receptiveness represents an important step towards understanding
how the process and conditions of transplanting constitutional adjudication
institutions affects their development and behaviour. It is here in part where future
research of constitutional jurisprudence in transition contexts must now focus. If it is
true that the study of post-communist transition is rapidly becoming a matter of
history (Dupré 2003, 3ff.), it is all the more important that the lessons taught by the
experiences of constitutional judiciaries in CEE and the FSU be gleaned before the
opportunity has vanished.
44
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APPENDIX
(N.B. All website references listed were last accessed and verified 01 August 2005.)
A. DATA SOURCES:
Data Source:
Brzezinski (2000)*
Data Utilised from this Source:
Comprehensive statistics on Polish
Constitutional Tribunal’s activity in first six
years of operation (used for purposes of
comparison with Smithey and Ishiyama
[2002] in their correlation analysis)
CODICES Database. European Commission
for Democracy Through Law (Venice
Commission, Council of Europe. Available
online: <http://codices.coe.int>
Full-text and précis of constitutional court
decisions examined; full-text of
constitutions and laws on the constitutional
courts [see listing below for detail by
country]; ‘Court Descriptions’ prepared by
Venice Commission staff. N.B.: Except
where indicated otherwise, all attributes of
constitutional courts and their powers were
coded directly from the respective
constitutions and/or laws on the courts.
Freedom House (1972-2005), Freedom in the
Indices of political rights and civil liberties
World Country Ratings. Data available online: (as used by Herron & Randazzo [2003] and
<http://www.freedomhouse.org/ratings/index.ht others in regression analysis).
m>. For CEE/transition specific data, see also
the 1998, 2000, 2001, 2002, 2003, 2004, and
2005 volumes in the Nations in Transit series,
each available at:
<http://www.freedomhouse.org/research/nattran
sit.htm>
Frye (1997)*
Index (first of two) of presidential power in
countries examined.
Mavčič, A. (2004) A tabular presentation of
constitutional/judicial review around the world.
Data available online:
<http://www.concourts.net/tab/>
Taxonomic and systematic presentation of
competences and access privileges of all
known constitutional courts in the world;
used for construction of alternative judicial
power index.
NCSEER Post-Communist Elections Project.
Index (second of two) of presidential power,
(1990-1996) Research conducted by J. Hellman and index of prime minister’s power, in
and J. Tucker. Data available online:
countries examined.
<http://www.wws.princeton.edu/~jtucker/pcelec
tions.html>
57
Organisation for Economic Co-operation and
Development. (2005) OECD Main Economic
Indicators, ESDS International, University of
Manchester. Accessed via Economic and
Social Data Service through the UK Data
Archive. Data available online [access via
Athens subscription and ESDS login]:
<http://www.esds.ac.uk/aandp/access/esds_logi
n.asp>
All GDP statistics (The present study uses
2005 OECD time-series data, neither lagged
nor adjusted for PPP, using 1990 as the base
year but reported in constant 2000 US
Dollars. Herron and Randazzo (2003) did
not indicate in either their article or in
response to queries the particular
parameters and source of their GDP
statistic.)
Pistor, Raiser, and Gelfer (2000)*
Indices of legal adaptation, familiarity,
transplant type. Also, provides listing of
rule of law, legal effectiveness and legal
enforcement scores for all countries from
European Bank for Reconstruction and
Development data (data which is not
otherwise available to the public).
Scheppele (2003b)
Comprehensive statistics on Hungarian
Constitutional Court’s activity in first six
years of operation (used for purposes of
comparison with Smithey and Ishiyama
[2002] in their correlation anaylsis)
Smithey & Ishiyama (2000)*
Original Smithey-Ishiyama index of judicial
power; measure of legislative fragmentation
at first election; measure of ethnic
fragmentation; dummy for history of
judicial review; measure of statist political
culture.
Smithey & Ishiyama (2002)*
Synthetic rights index; measures of popular
trust in courts, popular trust in courts vs.
trust in parliament, popular trust in courts
vs. trust in president. N.B.: Data on
popular trust reported by Smithey &
Ishiyama was taken from the New
Democracies Barometer II and III; the New
Russia Barometer II, III, and IV; and the
Nationalities in the Baltic States Survey I
and II. The data from these surveys is no
longer available from its original source
(viz. the University of Strathclyde’s Centre
for the Study of Public Policy, directed by
Dr. Richard Rose), hence the data reported
by Smithey & Ishiyama must be utilized.
58
University of Essex Project on Political
Transformation and the Electoral Process in
Post-Communist Europe. (1999-2002)
Research conducted by F. Millard, S. Birch, K.
Williams, and M. Popescu. Data available
online: <http://www.essex.ac.uk/elections/>
Parliamentary election results, utilised for
calculation of legislative fragmentation in
year of decisions (all calculations made by
the author from election results).
World Bank (2000), World Development Report The 1999/2000 edition of the annual World
1999/2000. New York: Oxford University
Development Report (WDR) provides both
Press.
the measure and data Smithey and Ishiyama
(2002) explicitly claim to have used in
For data, see Id., Appendix, Table A.1 (pp. 216- measuring the degree of federalism (i.e.
217) Decentralization. Available online:
number of elected subnational tiers of
<http://wdsbeta.worldbank.org/external/default/ government). However, as there are several
WDSContentServer/IW3P/IB/2000/12/13/0000 discrepancies between Smithey and
94946_99092312334240/Rendered/PDF/multi_ Ishiyama’s reporting of this measure for
page.pdf>
particular countries and that found in the
WDR data, the WDR data is used in all
cases.
N.B.: Sources marked with * are listed in the Reference list.
59
B. RELEVANT LEGAL INSTRUMENTS
(N.B. All of the following are available, in English, via CODICES Database, supra])
Country
Czech Republic
Legal Instruments
Constitution of the Czech Republic
On the Constitutional Court (Act of 16 June 1993, No. 182/1993 Sb.), 16
June 1993.
Estonia
Constitution of the Republic of Estonia
Estonian Law on Constitutional Review Court Procedure
Law on Courts
Georgia
Constitution of Georgia
Law of Georgia on the Constitutional Court of Georgia, 31 Jan. 1996. As
amended by the Law of Georgia on changes and additions in the Law
of Georgia on the Constitutional Court of Georgia, 21 March 1996.
Law of Georgia on Constitutional Proceedings, 21 Mar. 1996.
Law of Georgia on social guarantees to the members of the Constitutional
Court, 25 Jun. 1996.
Hungary
Constitution of the Republic of Hungary
Act on the Constitutional Court (Act XXXII of 1989). As amended by
Act LXXVIII of 1994.
Lithuania
Constitution of the Republic of Lithuania
Law on the Constitutional Court of the Republic of Lithuania, 3 Feb.
1993. As amended by Law I-1475 of 11 Jul. 1996.
Moldova
Constitution of the Republic of Moldova
On the Constitutional Court (Law No. 317-XIII), 13 Dec. 1994.
Poland
Constitution of the Republic of Poland
The Constitutional Tribunal Act (Dziennik Ustaw 1997 No. 102 item
643), 1 Aug. 1997. As amended by: Dziennik Ustaw 2000 No. 48 item
552; Dziennik Ustaw 2000 No. 53 item 638; Dziennik Ustaw 2001 No.
98 item 1070.
Russian Federation
Constitution of the Russian Federation
Federal Constitutional Law on the Constitutional Court of the Russian
Federation, 24 Jun. 1994.
Slovakia
Constitution of the Slovak Republic
Act on the organization of the Constitutional Court of the Slovak
Republic, on the proceedings before the Constitutional Court and the
status of its judges (Act No. 35/193), 20 Jan. 1993.
Slovenia
Constitution of the Republic of Slovenia
Law on the Constitutional Court (Official Gazette RS, no. 15/94).
60
C. CONSTITUTIONAL COURT WEBSITES
(N.B. All website references listed were last accessed and verified 01 August 2005.)
Czech Constitutional Court: http://www.concourt.cz
Estonian Supreme Court: http://www.nc.ee
Georgian Constitutional Court: http://www.constcourt.gov.ge
Hungarian Constitutional Court: http://www.mkab.hu
Lithuanian Constitutional Court: http://www.lrkt.lt
Moldovan Constitutional Court: http://www.constcourt.md
Polish Constitutional Tribunal: http://www.trybunal.gov.pl
Russian Federation Constitutional Court: http://ks.rfnet.ru
Slovakian Constitutional Court: http://www.concourt.sk/index.html
Slovenian Constitutional Court: http://www.us-rs.si
61