Representation and Policy Responsiveness:
The Median Voter, Election Rules, and
Redistributive Welfare Spending
Shin-Goo Kang Korea University
G. Bingham Powell Jr. University of Rochester
Many economic and social conditions shape public welfare spending. We are able to show, however, that after taking
account of these conditions, the expressed left-right preferences of the median voters significantly affect comparative
welfare spending. These new findings support the representational claims of liberal democracy and the theoretical
expectations of the literature on ideological congruence. However, we also show that insofar as the preferences of
citizens and the promises of governing parties (which are highly correlated,) can be disentangled, it is the former that
affect the long-term redistributive welfare spending equilibrium, while the latter have small, but significant shortterm effects. Surprisingly, despite greater representational correspondence between positions of voters and governments under PR than SMD, the impact of the median voter preferences is quite similar under the two systems.
W
e are accustomed to identifying liberal
democracy by a process: competitive elections systematically producing representative governments. A great deal of research in political
science has been devoted to illuminating elements in
that process. Much progress has been made in showing
how voters, party competition, election rules, and
government formation shape political representation.
But liberal democracy promises substance as well
as process. As both a normative and an empirical
theory, the process is supposed to deliver public
policies that the citizens want. Under a democratic
government ‘‘a majority of citizens can induce the
government to do what they most want it to do and
to avoid doing what they most want it not to do’’
(Dahl 1989, 95.) Analyses of representation and
responsiveness have seldom attempted to bridge this
final gap between citizens and substantive policies.
In this paper we show that in at least one major
policy dimension, redistributive welfare policy, the
preferences of the citizens did significantly shape
government spending in the Western industrialized
democracies between 1960 and the early 1990s, as we
would expect from the literature on ideological
congruence. The impact of the median voter emerges
even controlling for social and economic context and
changing conditions that also affect redistributive
welfare spending in major ways. These results are
encouraging for the basic promises of liberal
democracy.1
Why the Median Voter Should
Matter for Policy Outcomes
Studies of ideological representation have clearly
articulated a theoretical structure connecting the
ideological position of the median voter and the
committed ideological position of the policymaking
government in developed parliamentary democracies.
Early versions of this theoretical construct can be
found in Huber and Powell (1994) and Cox (1997,
chap. 12). The connections are argued to work
through somewhat different mechanisms of party competition and government formation in democracies
1
An online appendix for this article available at http://journals.cambridge.org/JOP contains detailed discussion of variable construction
and statistical analyses. Data and supporting material necessary to reproduce the numerical results in the paper are available at http://
mail.rochester.edu/collegefaculty/gpowell/webpage_210.htm.
The Journal of Politics, Vol. 72, No. 4, October 2010, Pp. 1014–1028
Ó Southern Political Science Association, 2010
1014
doi:10.1017/S0022381610000502
ISSN 0022-3816
representation and policy responsiveness
1015
with single-member district election rules than in
democracies with proportional representation rules
and their associated party systems. But both mechanisms theoretically link the median voter to the
campaign promises articulated by the party or parties
in government. The governing parties are then expected to carry out policies consistent with these
promises. At least in the time period of our study
(1960–91) redistributive welfare policies constituted
an important part of the discourse of left-right
ideological competition.
In the single-member district systems the number
of parties is reduced (Duverger’s Law) and Downs’s
(1957) theory of two-party convergence to the median
voter leads to the expectation that a party close to the
median voter will receive a parliamentary majority.
This party then forms a government and, under close
scrutiny from the electorate, which can clearly assess
its accountability in keeping its promises, carries out
its promised policies, which are favored by the median
voter. In the PR systems more parties compete and,
given the accurate vote-seat correspondence generated
under PR in developed democracies (Lijphart et al.
1994; Rae 1967) the ideological range of the citizens is
reflected into the legislature. The median legislative
party is close to the position of the median voter. As
no party has received a majority, postelection bargaining results in coalition (or minority) governments. The
legislative median party is advantaged in coalition
bargaining and is usually included in the coalition
government, (Laver and Schofield 1990, 113) thus also
linking the new government and its policies to the
median voter. Thus, through the processes of party
competition, representation and government formation we expect:
The representation literature is divided on
whether majoritarian or PR systems should be expected to generate governments closer to the median
voter, or whether each should be equally effective.
(See, e.g., Powell 2000 versus Blais and Bodet 2006 or
Golder and Stramski 2007.) Cox (1997) argues that it
depends on where in the competition and formation
stages we are more likely to find ‘‘coordination
failures.’’ Empirical studies are also divided. In the
time period that we are examining, at least, the PR
systems seem to generate governments closer to the
median (Huber and Powell 1994; McDonald,
Mendes, and Budge 2004; McDonald and Budge
2005; Powell 2000, 2006.)
This greater government ideological correspondence under PR leads us to expect, prima facie, that the
effect of the median voter on welfare spending would
be greater under PR. Governments are expected to
carry out promised levels of spending, so that leftist
governments would spend more on redistributive
welfare than more conservative governments. Therefore, these literatures showing closer connections
between voters and governments under PR provide
good reason to expect that PR election rules would
also induce greater effect of the median voter position
on redistributive welfare spending. That is, we would
expect either that (1) a unit change in the position of
the median voter would lead in equilibrium to a
greater change in redistributive welfare spending
under PR (a larger slope coefficient,) or (2) that
there would be more random error in the process
under SMD, leading to a similar equilibrium effect,
but larger standard errors and less statistical significance. Thus, we expect:
H1: Controlling for stable and changing social and
economic conditions, the left-right preferences of
the median voter significantly affect the equilibrium level of redistributive welfare spending.
H4a: Ideological preferences of the median voters are
more accurately reflected in legislatures and
governments under PR than under SMD.
H2: Controlling for stable and changing social and
economic conditions, the left-right preferences of
the government significantly affect the short-term
and equilibrium levels of redistributive welfare
spending.
H3: Controlling for stable and changing social and
economic conditions, introducing a control for the
left-right position of the government will account
for the median voter-spending connection.
Hypothesis 3 reflects the status in this literature of the
government positions as the mediating causal mechanism through which the median voter preferences
exert their impact.
H4: Election Rules, Voters and Welfare Spending.
H4b: Because of the more accurate ideological correspondence of governments, the preferences of the
median voter will have greater impact on redistributive welfare spending under PR than under
SMD.
Of course, the theoretically posited mechanisms may
fail under either type of system. The presence or
threat of additional parties may inhibit convergence
to the median voter in majoritarian systems, or other
Downsian assumptions may be unrealized (Grofman
2004). Or, party activists may systematically resist
convergence to the median, as proposed in analyses
of intraparty decision making (Kitschelt 1994; May
1973; Norris 1995.) In systems relying on multiparty
1016
coalition formation, the need to build legislative
majorities may pull the coalition away from median-oriented campaign pledges. Or party leaders in
both types of systems may respond to informed
opinion leaders, rather than the average citizen
(Adams and Ezrow 2009).
Moreover, the ideological congruence and representation literature has focused on party campaign
commitments or expert party placements, not on the
outcomes of policies actually chosen. The putatively
greater accountability of the government in the
majoritarian systems (Duch and Stevenson 2008)
might offer a reason to expect its policies to show
to greater advantage than its promises. Governments,
anticipating retribution for unpalatable policies, may
in policymaking practice pay more attention to the
voters’ preferences than to their own campaign
promises, especially in the majoritarian systems. Or,
the time lags involved in policy outcomes might lead
to off-setting ideological distortions.2 It seems particularly important, therefore, to investigate the expected relationship between the median voter and
policy outcomes associated with left-right ideology.
Measurement and specification problems have previously made this very difficult.
Measurement and Specification:
Drawing on Previous Research
Traditions
There are at least two very good reasons why most
representation studies do not include substantive
policy analysis. One problem is the difficulty in
ascertaining what policies the voters want. A second
problem is the complexity of policies themselves,
which are shaped by many social and economic
conditions, very often making it hard to determine
the role of the political elements. Here we take
advantage of two distinguished bodies of research.
To identify the position of the median voter in 500
country-years across 17 countries over 30 years we
draw on the work of Kim and Fording (1998, 2002),
themselves building on the important research program coding party manifestos in terms of the space
devoted to different political issues (Budge, et al.
2001,) transformable into left-right placement (Laver
and Budge 1992, 24–27). It is critical for our purposes
2
Also see the alternative hypotheses and inferences in the threecountry responsiveness comparison of Hobolt and Klemmensen
(2008).
shin-goo kang and g. bingham powell jr.
that these estimated voter placements be substantively comparable across countries and time, not
simply reflecting a party’s relative position in local
campaign discourse. It is also highly relevant that
issues of redistributive welfare constitute a part of the
left-right discourse. The approach of the manifesto
coding meets these needs.3 Studies relying on citizen
surveys tend to find that citizen in most countries
place themselves in the center of what are obviously
varying local discourses, reflecting different balances
of issues and different understanding of ‘‘center,’’
hence not useful for predicting substantive policies.
To specify the role of political factors in the midst
of social and economic ones we draw on the work of
Robert Franzese (2002, chap. 2). Franzese’s ‘‘tax and
transfer systems’’ are measured by ‘‘the sum of socialsecurity benefits, social-assistance grants, and unfunded welfare and pension payments by general
government’’ (2002, 62). These tax and transfer
systems (also called ‘‘social insurance’’ or ‘‘redistributive welfare’’) consumed a large and increasing part
of the share of GDP in the Western countries from
the 1960s onward, from about 8% in 1962 and
reaching over 20% of the GDP in a number of
countries by 1995.4
Franzese develops a carefully specified theory and
econometric model of redistributive welfare transfers
as a percentage of GDP. By building on Franzese’s
work we can avoid simplistic attribution of these
redistributive welfare expenditures to voter preferences
when they are really responding to such powerful
factors as change in age distribution, unemployment
levels, national income fluctuations, and inflation.
Indeed, one problem of an analysis of redistributive welfare expenditure is that welfare spending is
embedded in large bureaucracies and a web of social
policy, which directly connect it to the age level of the
population and the level of unemployment, two large
3
For a complete description of how to compute median voter and
government ideological position, see Kim and Fording (1998,
2002) and the online appendix. We recognize not only that these
estimations are not without controversy, but that the revealed
position of the median voter is not quite the same as the
unmediated preference of the median citizen. We are not able
to take account of differences between voters and nonvoters.
Moreover, median voter positions may be affected by party
offerings. One might consider these ‘‘induced’’ preferences,
rather than authentically prepolitical ones. However, any theory
of the effect of elections must work through these induced
preferences.
4
Another approach to measuring redistributive welfare programs
is offered by Mares’s careful coding of social policy legislation in
over 130 countries (2005.) However, her summary country
characterizations are purely cross-sectional and thus not useful
for our dynamic analysis.
representation and policy responsiveness
components of spending, as well as to other social
needs. On one hand, it will change as they change; on
the other hand, such bureaucratized spending inevitably has a large inertial component, limiting programmatic change in a given year. Moreover, as
welfare spending is measured as a percentage of
GDP, necessary to compare spending figures across
countries, it is also sensitive to fluctuations in the
GDP itself, and to inflation changes.
Comparative policy studies, more generally, reflect a substantial debate over what social and
economic factors should be controlled and whether
political factors matter at all in social welfare spending. For example, Imbeau, Petry and Lamari (2001)
report a meta-analysis of 43 studies of the relationship between the left-right composition of the government and public policy outputs, involving 164
estimates of the relationship with social welfare
spending. Only 25% of those estimates yielded
statistically significant (at .10) estimates in the correct
direction (2001, 18.) See also, for example, the small
partisan effects reported in the cross-national analysis
of Blais, Blake, and Dion (1993) and the varying
cross-national findings of Klingemann, Hofferbert,
and Budge (1994). Both specification disputes and
lack of substantively comparable ideological measures
are involved in these varying results.
Franzese’s own theoretical and empirical analyses
(2002, 86), show the power of the specification
variables and also examine openness of the economy,
income skew, and other features proposed by various
political economists. Moreover, his error correction
model enables us to ascertain both short-term effects
and long-term equilibrium relationships, which will
be essential in connecting voter preferences to policies. Thus, it provides a very valuable starting ground
for our endeavor to explore the relationship between
the redistributive welfare spending and the preference
of the citizens in these countries.
Redistributive social welfare policies and their
expansions and contractions have been at the heart of
left-right policy discourse in the Western democracies for many years, and especially in the 30 years
between 1962 and 1991 which are our focus here.5 At
the same time many election studies have shown that
voters in these countries during this time period
5
Although the Franzese data that we use include the 1950s, we
begin our analysis in 1961. On one hand the 1950s include
economic and social rebuilding after WWII; probably for this
reason the social and economic model does not fit well in those
years. On the other hand, the Australian and New Zealand data
are not available until after 1960; even Italy only begins in 1955;
and France switches to SMD in 1958, so our comparison of PR
and SMD systems cannot be carried out in the 1950s.
1017
make extensive use of the perceived positions of the
parties on a left-right dimension in making their
voting decisions. Thus, we have good reason to hope
that if welfare policies are responsive to what voters
want, we shall be able to estimate the degree of
responsiveness building on these two research
programs.
Does Redistributive Welfare
Spending Respond to the Median
Voter?
Because of the problem of estimating the median
voter position, few studies have attempted to link
these positions to the redistributive welfare spending
levels. One exception is McDonald and Budge (2005),
who also use manifestos to estimate the positions of
median voter, legislature, and government. They do
find, encouragingly, reoccurring associations between
average median voter positions and average government spending in their analysis of 21 countries
between 1973 and 1995 (216 ff.) However, they only
control for a few socioeconomic conditions, openness
of the economy, and government centralization, in
predicting central government spending (217.) Nor
do they take account of any fixed country effects,
presumably because their analyses focus on average
cross-sectional comparisons (222, 224.)
Figure 1 shows the severity of the analytic
problem linking changing voter preferences and
redistributive welfare spending. Panel (A) in Figure
1 shows average redistributive welfare spending each
year from 1962 through 1991. The three lines show
spending in countries with proportional representation electoral systems (PR, marked by squares),
spending in countries with Single-Member District
electoral systems (SMD, marked by diamonds), and
(in the middle) the average for all countries (marked
by black circles).6 We shall return below to the
distinction between spending under the different
electoral systems. At the moment the important point
is that redistributive welfare spending as a percent of
GDP increased more or less continuously from the
mid 1960s until the early-1980s, leveled off, and then
ticked up again at the end of the period. We can
6
The PR countries are Austria, Belgium, Denmark, Finland,
Ireland, Italy, Netherlands, Norway, Sweden, Switzerland, and
West Germany; the SMD countries are Australia, Canada, France,
New Zealand, United Kingdom, and United States. (France 1986,
although held under PR rules for that one election, is still coded
as SMD.)
1018
shin-goo kang and g. bingham powell jr.
F IGURE 1 Redistributive Welfare Spending and Median Voter Ideological Position
(A) Redistributive Welfare Spending
compare this to the median voter ideological position
(on a 0–100 scale from right to left,) which is shown
in Panel (B). Again, postponing the discussion of the
different tracks for different electoral systems, we see
that voters were moving left in the 1960s, at least in
the PR systems, but shifting towards the right in the
1970s and early 1980s—contrary to the increased tax
and transfer spending. A simple comparison of the
trends between median voter changes and tax and
transfer spending changes suggests little correspondence between the two, perhaps even a negative
association, which is completely contrary to our usual
assumptions about ideology and spending.
But social and economic conditions, such as
growth, unemployment, and inflation must be taken
into account, as they have direct effects on social
insurance as a percent of GDP. We also take into
account the effect of spending level itself, which
should ease demands and run into resource constraints, such that we expect that the higher the current
level of expenditures, the slower the increase. Voter
preference effects must operate against that background, so we must turn to the fully specified multivariate error correction models. In an error-correction
model, a set of variables have a long-run equilibrium
relationship. When a certain shock occurs due to
changes in independent variables or some unexplained
(B) Median Voter Ideological Position
factors, the dependent variable changes—or adjusts to
a new value—in a way to maintain the equilibrium
relationship.
Model 1 in Table 1 replicates the full Franzese
2002 model of changing tax and transfers for the 17
countries in our analysis for the 30-year period from
1962 through 1991.7 The analysis includes all the
economic and social factors suggested in his analysis,
including lagged spending, average spending change
in all the countries to capture a possible time trend,
and political variables and proxies for them, as well as
country dummies to capture fixed effects. We have
deleted his estimate of the government’s ideological
position (about which more later), which was not in
fact statistically significant by usual standards in
Franzese’s analysis. As we see in column 1, which
reports coefficients similar to those in Franzese
(2002), the change in tax and transfers is powerfully
affected by previous spending levels and by such
7
For comparison purposes of over time analysis, we include only
the 17 ‘‘Western’’ countries that are democracies throughout the
period. That is, from Franzese’s countries we exclude the new
democracies of Greece, Portugal, and Spain, which introduce or
reestablish democracy in the 1970s. We also exclude Japan, as we
are less confident in the substantive left-right comparability of
the manifestos. To Franzese’s countries we are able to add New
Zealand, drawing on the similar sources and measures.
representation and policy responsiveness
short-term factors as change in unemployment,
change in inflation, economic growth, change in
economic growth, and national wealth (the natural
log of real GDP per capita), each of which has a
strong theoretical justification in Franzese’s discussion. (Most of the proposed proxies for political
conditions emerge as insignificant, so are not discussed further.) We cannot reliably assess the effect of
the expressed preferences of the median voter without controlling for the key social and economic
conditions.
In Model 2 of Table 1 we add our estimates of the
position of the median voter in the previous year
(Median voter t-1) and change in position of the
median voter (DMedian voter t) to the fully specified
equation. Following on previous research on welfare
spending in general (e.g., Wilensky 2002, Part II,) we
also add a variable for percent of the population who
are Catholic (Catholic), as Catholic countries and
party systems are argued to additionally favor such
policies.8
We can draw four inferences from the coefficients
in Model 2. First, we see that most of the control
variables are largely unchanged. Second, the number
of significant country dummies has declined from
eight (at the .10 level) or six (at the .05 level), to only
two, which is a pleasing improvement in theoretically
interesting explanatory power. In these specifications
including voter preferences only Belgium and the
Netherlands retain statistical significance as country
dummies. Third, we see that contemporaneous
change in position of the median voter is insignificantly related to changes in tax and transfer. Given
the lags and complexities of tax and transfer spending, we are not surprised by this, but it is still notable
that short-term change in median voter position does
not induce change in redistributive welfare spending.
Finally, we see that the coefficient for the lagged
ideological position of the median voter is significantly related to change in tax and transfer spending
at the .05 level. Given the conservative nature of the
fixed effect specification, it is normatively pleasing, as
well as interesting, to find that expressed median
voter preferences do affect tax and transfer spending,
through a long-term equilibrium process, consistent
with Hypothesis 1.
8
Since Catholic is measured as a time-invariant variable, an
additional dummy variable for country fixed effects—in our
case, Ireland—should be eliminated from the specification to
avoid perfect collinearity. Note that there is no identification
problem between the two excluded countries—the United States
and Ireland—since they are different in the level of Catholic
population.
1019
However, we must also report that the effect is
not very large in this specification. As we see in the
equation, many other factors affect short-term tax and
transfer spending as a percent of GDP.9 The expressed
preferences of the voter are only one element. We can
estimate the magnitude of their effect. In this kind
of change (error correction) model, the substantive
impact of permanent change in the level of a simple
independent variable on the expected long-term equilibrium level of the dependent variable is estimated
by dividing its coefficient by the (negative) coefficient
for the lagged dependent variable. (See Franzese 2002,
81–85, notes 40 and 44.) The highly significantly
negative coefficient estimate on the lagged dependent
variable—with a z-value 5.60—justifies this interpretation. Here, we divide the coefficient for lagged median
voter position (.0058) by the lagged spending variable
(2.1029), to derive an expected equilibrium effect
of .0579.
That is, in equilibrium a 10-point ideological
difference in the position of the median voter, which
is roughly standard deviation of the variable, would
result in a difference of .6% in tax and transfer
spending. A 20-point difference in expressed ideological position, which is about the average median voter
difference between the United States and Sweden,
would be expected to result in about a 1.2% spending
difference. (The real average spending difference between these two countries in this 30-year period is
about 5 %.) While it is impressive that this analysis
reverses the negative trends of association in Figure 1,
we could not characterize it as a dominant factor,
despite its substantive, as well as statistical, significance.
However, we have good reason to believe that the
specification in Model 2 is overly conservative. It is a
‘‘fixed effect’’ model that includes dummy variables
for each country (except two, which constitute the
base line.) Thus, it virtually eliminates all the crosssystem effects and reflects only within-country change.
We know that there are systematic differences across
countries in the position of the median voter. The
median voter in Sweden, to use the example mentioned above, is consistently well to the left of the
median voter in the United States. Putting in dummy
variables for each country understates such effects.
(Franzese was well aware of this problem, as he
mentions (84, note 42), but was not concerned with
cross-national differences. In general, see Beck and
Katz (2004, 4–6).) The full fixed-effect specification
9
We assume for purposes of this analysis that these conditions are
largely exogenous with respect to the position of the median
voter, consistent with the stability of their effects when it is
entered into the equation.
1020
shin-goo kang and g. bingham powell jr.
T ABLE 1
Median Voter and Redistributive Welfare Spending (ECM Models)
Variables
Model
1
Constant
Tax-and-Transfer t-1
DMedian voter t
Median voter t-1
Catholic
DUnemployment t
Unemployment t-1
DAge65+ t
Age65+ t-1
DInflation t
Inflation t-1
D(DGDP t)
D GDP t-1
GDP t-2
Openness t-1
Central-tax share t-1
Indirect-tax share t-1
Total-tax share t-1
Union density t-1
DElection year t
Election year t-1
DGovernment hazard rate t
Government hazard rate t-1
Income skew variability t-1
Voter turnout t-1
Voter turnout t-1*Income skew t-1
Income skew t-1
DTax-and-Transfert-1
DWelfare spending trend t
Germany
France
Italy
U.K.
Canada
Austria
Belgium
Denmark
Finland
Ireland
Netherlands
Norway
Sweden
Switzerland
Australia
N.Z.
Number of Cases (d.f)
Adjusted R2
Hausman Test (Model 2 vs. 3)
28.686
20.103
Model
2
(3.176)
(0.018)
0.186
(0.042)
0.004
(0.015)
0.336
(0.179)
0.017
(0.040)
20.028
(0.010)
20.001
(0.008)
212.968
(1.376)
211.689
(2.094)
0.876
(0.329)
20.551
(0.445)
20.331
(0.551)
0.924
(1.086)
0.630
(1.231)
0.009
(0.005)
20.016
(0.063)
0.044
(0.102)
0.027
(0.124)
0.182
(0.127)
0.389
(2.053)
0.440
(1.552)
0.105
(1.649)
0.256
(1.235)
0.031
(0.058)
0.203
(0.063)
0.537
(0.427)
1.215
(0.406)
0.900
(0.436)
0.168
(0.370)
0.266
(0.252)
0.953
(0.494)
1.792
(0.579)
0.384
(0.410)
0.627
(0.323)
1.088
(0.553)
2.196
(0.555)
0.863
(0.416)
0.561
(0.429)
0.255
(0.313)
20.366
(0.364)
0.390
(0.456)
501(459)
0.5584
Model
3
(3.310)
(0.018)
(0.005)
(0.003)
(0.009)
(0.042)
(0.015)
(0.181)
(0.040)
(0.010)
(0.008)
(1.375)
(2.095)
(0.331)
(0.441)
(0.547)
(1.051)
(1.225)
(0.005)
(0.063)
(0.102)
(0.124)
(0.128)
(2.045)
(1.638)
(1.733)
(1.302)
(0.057)
(0.064)
(0.366)
(0.289)
(0.356)
(0.536)
(0.190)
(0.328)
(0.298)
(0.629)
(0.528)
22.810
20.068
0.002
0.010
0.006
0.192
0.000
0.479
0.049
20.021
0.006
213.490
211.516
0.191
20.202
20.314
0.732
0.513
0.008
20.045
20.004
20.025
0.082
2.944
0.224
20.215
20.147
0.038
0.181
(1.792)
(0.013)
(0.005)
(0.003)
(0.001)
(0.042)
(0.013)
(0.166)
(0.018)
(0.010)
(0.008)
(1.417)
(2.118)
(0.144)
(0.175)
(0.281)
(0.658)
(0.857)
(0.003)
(0.064)
(0.103)
(0.118)
(0.093)
(2.049)
(1.274)
(1.436)
(1.215)
(0.056)
(0.066)
0.428
(0.103)
0.395
(0.153)
2.001
(0.520)
1.028
(0.642)
0.765
(0.620)
0.034
(0.279)
20.409
(0.384)
0.538
(0.543)
501(457)
0.5601
x2½32 5 14:18 ð0:9958Þ
1.222
(0.199)
210.246
20.103
0.003
0.006
0.015
0.184
0.011
0.375
0.027
20.026
0.002
213.260
212.221
0.889
20.583
20.326
1.323
0.454
0.009
20.021
0.040
0.026
0.192
0.474
1.306
20.681
0.841
0.027
0.196
0.234
0.410
20.043
0.336
20.086
20.028
0.670
0.625
0.904
501(469)
0.5416
Note: Panel-corrected standard errors are in parentheses. The coefficient estimates in Bold indicate statistical significance at p , .05 in
two-tailed tests. Model 1 includes all country dummy variables except for the US. Model 2 excludes an additional country dummy
(Ireland) from the specification to avoid perfect collinearity due to the inclusion of Catholic, a time invariant variable. Model 3 excludes all
country dummy variables except for three countries. Description of variables is reported in replication data file. D is the difference operator.
representation and policy responsiveness
impedes us from fully assessing the effect of election
rules and other variables that are either time-invariant
or slowly changing—hence their impact is largely
presumed to be due to cross-sectional differences.
The preferred specification is Model 3. We have
observed that three countries, France, Netherlands, and
Belgium, consistently spend more on redistributive
welfare than other countries in many specifications,
suggesting that these three countries may have different
data-generating processes and that it is not appropriate
to pool these countries together with the others. If we
put in dummy variables for just those three countries,
they are highly statistically significant and, more importantly, retain the significance of the lagged spending
variable, and create a statistically acceptable specification.10 We should emphasize that this approach involves a kind of trade-off, so we shall continue to
compare our results with the full fixed-effects models.
In this specification the short-term effect of change
in median voter position remains insignificant, but the
equilibrium effect of the level of left-right position of
the median voter is highly significant. (In fact, it is
robust to many alterations in specifications, as we shall
see.) The size of the median voter level coefficient is
nearly doubled (0.0101), and when divided by the
lagged spending variable (2.0684), leads to a more
than doubled estimated equilibrium effect of .1478,
(which is significant at .001). Similarly, some of the
other variables that are notable for slow trend movement and large cross-sectional differences, such as
Catholic population, age structure, and union density
show substantially increased size or significance levels.
It is reassuring to confirm previous work on the role of
Catholicism and union membership in redistributive
welfare spending. Most of the other major social and
economic conditions, such as change in unemploy10
As we might expect from these considerations, a specification
without any country dummies greatly increases our estimation of
the effect of median voter differences. However, statistical tests
show this specification is impermissible. Among other concerns,
the estimate of the lagged spending variable is small and barely
significant, suggesting an unstable equilibrium relation. Estimation without country indicators may suffer from inconsistency
due to omitted variable bias when the explanatory variables in
the equation are correlated with the excluded atheoretical
country-specific dummy variables. The Hausman specification
test is devised to test for the orthogonality of the regressors and
disturbances in the equation with the null hypothesis of no
correlation (Greene 2000, 576). The Hausman test of Model 3
against Model 2 indicates that the null cannot be rejected. The
result suggests that the coefficient estimates of Model 3 are
consistent and efficient. Moreover, Lagrange-Multiplier residual
correlation test (Greene 2000, 540) fails to reject the null of no
serial correlation of the errors (p , .1739), suggesting that
estimation of OLS with panel corrected standard error is
appropriate.
1021
ment and economic growth, remain largely unchanged, reminding us of their importance in the
specification.
Substantively, a 10-point difference in average
median voter position would lead to 1.478% greater
spending of GDP. The 20-point difference between
the United States and Sweden leads us to expect 3%
greater spending in Sweden. This is a substantively
very important effect, which would account for a
majority, although not all, of the average spending
difference between the two countries. Change in
redistributive welfare spending does respond, albeit
in a lagged equilibrium, to the preferences of the
median voter, consistent with Hypothesis 1.
Median Voters, Governments, and
Redistributive Welfare Spending
Spending policies are not carried out by voters or
election rules. They are carried out by governments.
Indeed, the standard theory of ideological congruence
expects that government left-right positions will be
the causal mechanism that connects the median voter
to redistributive welfare spending, as sketched in
Hypotheses 2 and 3 above.
Table 2 (Model 4) simply replaces the ideological
position of the median voter with the position of the
government, estimated from its parties’ campaign
manifesto promises, using our preferred, three country-dummy specification (from Table 1, Model 3.)
For ease of reading, Tables 2 and 3 show only the
coefficients (and standard errors in parentheses) of
theoretical interest from the fully specified interactive
models.11 Two points are immediately apparent.
First, whereas the short-term change in position of
the voters was seldom significantly related to policy
change, the short-term change in government position
(DGovernmentt) is linked to distributive welfare policy
change. A 10-point change to the left in government
ideology is immediately followed by about a twentieth
of a percent change [0.005 3 10 5 0.05 percent of
GDP] in distributive welfare policy; the effects are
small, but statistically significant. (This is also true in a
specification with all country dummies, as in Model 2,
that is, a complete fixed effects model, and is generally
a very robust effect.) Recall that social and economic conditions, as well as previous spending, are
controlled.
11
The full estimation results are available in the online appendix,
Tables 2 and 3.
1022
T ABLE 2
shin-goo kang and g. bingham powell jr.
Median Voter, Governments, and
Redistributive Welfare Spending
Model 4
Model 5
Constant
21.509 (1.791) 22.731 (1.803)
Median Voter
short-term effects:
DMedian voter
20.001 (0.006)
long-term effects:
2Median Votert-1/
0.128 (0.045)
Tax-and-Transfert-1
Government
short-term effects:
DGovernmentt
0.005 (0.002)
0.003 (0.002)
long-term effects:
2Governmentt-1/
0.074 (0.028)
0.021 (0.025)
Tax-and-Transfert-1
Number of Cases (d.f)
501(469)
501(467)
Adjusted R2
0.5356
0.5417
Note: Only the coefficients of theoretical interest are reported in
the table. The results of full models are reported in Table A2 in
the on-line Appendix. Panel-corrected standard errors are in
parentheses. The coefficient estimates in Bold indicate statistical
significance at p , .05 in two-tailed tests. All models include
only three country dummy variables for France, Belgium, and the
Netherlands.
Second, there are also long-term equilibrium
effects of government ideological position. These,
too, are statistically significant, and much larger, with
a 10-point government policy difference leading to
about .7% of an increase in distributive welfare policy
spending. So, Hypothesis 2 is supported in both
short-term and long-term equilibria.
However, in comparison to the results from
Model 3, we recall that the long-term government
equilibrium effects are not quite as large as those
associated with long-term position of the median
voter. (Again, the larger median voter effects hold at
the comparatively reduced levels of the full fixed
models also.)
The greater short-term power of the government
effects, but greater long-term power of the median
voter can be seen explicitly in Table 2, Model 5,
where we enter both median voters and governments
into the equation. Of course, as median voters’ and
governments’ left-right positions are fairly highly
associated, (r 5 .69) multicollinearity may be a
problem; the magnitudes of all voter and government
coefficients are reduced, and we cannot be completely
confident about their shared effects. But we do see
clearly that the two outstanding political effects are
long-term voter equilibrium and short-term government change. A 10-point change to the left by the
voters eventually results in a 1.28% increase in
spending, whereas a 10-point change to the left by
the new government immediately induces a .03%
welfare spending increase. There are no significant
effects of either short-term voter change or—having
taken voter ideology into account—of long-term
government ideological differences.12 This result is
certainly contrary to Hypothesis 3, which posited the
government position as the linking causal mechanism
between voter and policy. Contrary to expectation,
the long-term voter equilibrium does not disappear
when we enter the government position into the
equation. In our preferred 3 dummy specification, as
shown in Table 2, it is the long-term government
position that becomes insignificant, while the longterm voter effect remains robust. Most previous
research on the effect of political factors on welfare
spending has paid attention to the government
partisanship. But this result suggests that it might
be the case that the government partisanship primarily
reflects the median voter connection (see McDonald
and Budge 2005).
Is the Impact of Voter Preferences
Greater under PR?
Consistent with Hypothesis 4a and the previous
ideological congruence studies, the better voter-legislature-government left-right correspondences associated with PR election rules can be also found in our
data. The average government in PR systems is an
average of 8 points from their median voter on a 100point scale, (on average slightly to the left,) while the
average government in the SMD systems is an average
of 13 points from its median voter (on average, more
substantially to the right.) This difference is statistically significant. As another indicator, the correlation
between the median voter position and the government position is .74 in the PR systems and only .54 in
the SMD systems. Moreover, as we have just seen, a
government’s ideological position does predict its
spending, as we expected from Hypothesis 2.
12
Model 5 uses the three country dummies. If we enter both
voters and governments into the full fixed-effects model, along
with a dummy for the election rules, short-term government
change remains significant. (Short-term vote change is not.) The
substantive magnitudes of both voter and government long-term
equilibria are nearly identical, but in this specification government is significant, whereas median voter is not. Multicollinearity
is even more of a problem, of course, with all the country
dummies and so much of voter, government, and election rule
difference being cross-national.
representation and policy responsiveness
T ABLE 3
1023
Median Voter and Spending in SMD/PR Electoral Systems
Model 6
SMD
Model 7
PR
SMD
Model 8
PR
SMD
PR
Constant
29.524 (3.395) 28.967 (3.189) 23.903 (1.940) 23.607 (1.826) 23.716 (1.958) 23.435
Median Voter
short-term effects:
DMedian voter
0.008 (0.008) 0.000 (0.006) 0.010 (0.008) 20.001 (0.006) 0.007 (0.009) 20.004
long-term effects:
2Median Votert-1/
0.045 (0.046) 0.064 (0.038) 0.150 (0.062) 0.136 (0.044) 0.139 (0.066) 0.103
Tax-and-Transfert-1
Government
short-term effects:
DGovernmentt
0.003 (0.003) 0.004
long-term effects:
2Governmentt-1/
0.010 (0.031) 0.037
Tax-and-Transfert-1
Number of Cases (d.f)
501(455)
501(466)
501(461)
Adjusted R2
0.5587
0.5443
0.5427
Hausman Test
x2½34 5 7:02 ð1:000Þ
(Model 6 vs. 7)
(1.842)
(0.007)
(0.052)
(0.003)
(0.037)
Note: Only the coefficients of theoretical interest are reported in the table. The results of full models are reported in Table A3 in the online Appendix. Panel-corrected standard errors are in parentheses. The coefficient estimates in Bold indicate statistical significance at
p , .05 in two-tailed tests. Model 6 includes all country dummy variables except for the US (baseline), Ireland, and Norway. These
variables are excluded to avoid perfect collinearity. Model 7 and 8 includes three country dummy variables for France, Belgium, and the
Netherlands.
We are surprised to report, however, that evidence
to support the expectation of greater voter impact
under PR (Hypothesis 4b) is slight and limited at best.
This is seen most clearly in Table 3, which estimates
the interactions between the median voter position
and change in the median voter position and the
election rules under several specifications.
The first two data columns, called Model 6, show
the results from the full fixed-effects model, whereas
next two columns, Model 7, show the results from the
model with dummies for the three outlying ‘‘big
spender’’ countries of Belgium, France, and the
Netherlands. Both models include all the control
variables from Table 1, plus a PR dummy.
Looking first at the short-term effects, we see that
coefficient for change in the median voter position
under SMD systems, is small, positive, and about
same size as the standard error (thus not significant
by usual standards) under SMD. The PR coefficient,
calculated by adding the SMD coefficient and the
interaction term, is virtually zero. The difference is
actually negative, although again not significant. This
pattern is replicated under the other specifications.
The long-term equilibrium effects (dividing the coefficients by the appropriate lagged dependent variable coefficient) are larger. The equilibrium effect
under PR adds about 40% to the effect under SMD
and is itself nearly statistically significant by usual
standards (p , .088), as shown in the last line of the
table. But the difference is not nearly statistically
significant (the z-value is merely 0.35).
In the specifications with the three ‘‘Big Spender’’
country dummies, shown in Model 7, the equilibrium effects under SMD and PR are both much
larger than in Model 6 and are statistically significant.
But the interaction term for PR is actually negative,
so the net effect coefficient for PR is slightly smaller
than for SMD, although (as we see at the bottom of
the table,) slightly more statistically significant because of the smaller standard error.13 However, the
difference is not statistically significant at any conventional level in this specification too (the z-value
is 20.21).
In the last column of Table 3 (Model 8) we take
an even greater risk with multicollinearity by examining
voters and government in SMD and PR systems.
13
Yet another, less statistically preferable, approach is to partition
the data sets into SMD and PR cases. Here, too, the estimated
long-term equilibrium effects of median voter position are quite
similar under SMD and PR.
1024
Here, as usual, the overridingly powerful effects are
the long-term median voter equilibria (slightly
greater under SMD, but similar and statistically
significant under each regime). After taking the
voters into account, long-term government equilibria
are very small and insignificant (although slightly
larger under PR.) Short-term government change is
not significant under either regime, although slightly
larger and closer to significance under PR (contrary
to what we might think.) The most notable difference
between the processes in the SMD and PR systems
seems to be quite positive (but not significant) shortterm voter change effects in SMD, which are negligible under PR. Because we know that median voter
position and government position are strongly correlated under PR, we don’t want to make too much of
the greater power of the voter. But it does seem quite
important that under SMD the governments, after
some limited movement in the direction of their campaign promises, revert to a policy equilibrium closer
to the position of the median voter.
Thus, we are forced to conclude that differences in
the impact of the median voter on redistributive
welfare spending are quite similar under SMD and
PR. In the full fixed-effects model, the long-term net
PR coefficient is larger, but the magnitude of the effect
is small, with a 10-point ideological difference inducing only about .6 % of spending difference, and the
magnitude of the difference between SMD and PR is
only a third of that and, again, not statistically significant. In all the models, the short-term effects are larger
under SMD although, again, not significant by usual
standards. Our expectations of stronger voter effects
on spending under PR are simply not realized, or, at
best, the traces of them are largely obscured by other
factors. The smaller standard error under PR is consistent with the idea of better representation, but the differences are small. Despite its plausibility given
superior government congruence under PR, Hypothesis 4b is not supported by the data.
Can Voter Preferences Account for
Greater Spending under PR?
A very well-known empirical finding in the policy
literature is that government spending and, especially,
spending on welfare projects is greater in political
systems with proportional representation election
rules (PR), than in political systems with singlemember district (SMD) plurality or majority election
rules (e.g., Iverson and Soskice 2006; Milesi-Ferretti,
shin-goo kang and g. bingham powell jr.
Perotti, and Rostagno 2002; Persson & Tabellini
2001, 2003; Powell 2002; Rogowski and Kayser
2002). This relationship appears in our data also, as
we saw in Panel (A) of Figure 1. In the SMD systems
of Australia, Britain, Canada, France, New Zealand,
and the United States, the average T&T as percent of
GDP in the 1961–91 period was 11.1%, while in the
PR-oriented systems it was 14.9%—a third larger.
Interestingly enough, the SMD and PR systems look
on average quite similar in the early 1960s, but
redistributive welfare spending grows much faster
in the PR systems between the late 1960s and mid
1980s, after which spending largely stabilizes in both
systems (see Wilensky 2002 on the equalization by
1960 and subsequent divergence). A simple comparison of change in T&T as a percent of GDP in the two
types of systems throughout the 1961–91 time period
shows the average yearly increase in PR systems is .35,
whereas in the SMD systems it is a third less at .25.
Various explanations focusing on the nature of
electoral competition have been proposed to explain
this relationship (e.g., Milesi-Ferretti, Perotti, and
Rostagno 2002; Iverson and Soskice 2006; Pesson and
Tabellini 2003; Powell 2002; Rodden 2006).
An alternative explanation for the difference
might lie in the different social and economic conditions confronting the SMD and PR countries. The
latter, after all, are located on the European continent
or near to it, while the latter include two countries in
North American and two in the South Pacific. For
example, examination of the individual conditions
shows greater unemployment growth in the SMD
systems, while the PR systems had more rapidly aging
populations. Perhaps the most notable net encouragement of welfare spending appears in the effect
of union density. Union density was substantially
greater in the PR systems, with about 49% of the
labor force unionized on average in this period,
compared to 36% in the SMD systems. The expected
consequence of this 13% difference would be about a
1.1% equilibrium encouragement of redistributive
welfare spending in the PR systems. However, in
Model 1 of Table 1 we saw the full set of social and
economic conditions, including union density, predicting redistributive welfare spending. If we examine
the country dummies in that specification, we find
that five of the six lowest coefficients (counting the
baseline category of the United States) belong to
SMD systems. This pattern of country coefficients
suggests that social and economic conditions do not
account for the full redistributive welfare spending
difference between countries with the two types of
electoral systems. Moreover, the previous section has
representation and policy responsiveness
already shown that the impact of the median voter is
quite similar under the SMD and PR systems.
However, an obvious and simple explanation is
suggested by Panel (B) of Figure 1. That is, voters in
PR countries simply prefer more welfare spending,
either for path-dependent historical reasons or because both welfare spending and PR rules reflect
values of social and political inclusiveness. With our
data we can now test this explanation for the greater
welfare spending in the PR systems. Except in the
early 1960s, and again in the early 1970s, the average
median voter in the PR systems is revealed as being
substantially to the left of the average median voter in
the SMD systems. Across the full time period, the
average median voter in the SMD systems was
revealed as 53 on the 100-point right-left scale, while
her PR counterpart was at about 61. As we know
from the previous analysis that the position of the
median voter does have in equilibrium a significant
and substantive impact on redistributive welfare spending, we expect that, all else equal, it should encourage
greater spending in the PR systems.
H5: The difference in ideological placement of the
median voters accounts for the greater redistributive welfare spending in PR countries.
The models we have examined so far cannot tell us
whether differences in social and economic conditions plus differences in the orientations of the
electorate are responsible for the difference between
the redistributive welfare spending under SMD and
PR. To explore that question we need simultaneously
to take account of election rules, social and economic
conditions, unique country contexts, and voter preferences. We want to find out whether the election
rules continue to have a substantively and statistically
significant impact after taking account of all these
conditions. This is difficult because of the strong
associations between these factors and because the
election rules are effectively fixed during the time of
our analysis.
We tackle this task by relying on ‘‘fixed-effect
vector decomposition,’’ which provides more reliable
estimates for time-invariant variables in time-seriescross-section data with unit effects than any alternative
estimator using three-stage estimation (Plümper and
Troeger 2007). The results are shown in Table 4.
Model 9 shows the analysis without the median voter
variables, which are added in Model 10.14
14
A very brief explanation of the three-stage ‘‘fixed effect vector
decomposition’’ estimation technique is presented in the online
appendix.
1025
These models yield several clear inferences. First,
the long-term equilibrium effect of the position of the
median voter remains statistically significant and at
about the same magnitude as the full fixed-effect
model. It is not reduced by taking additional account
of the election rules in Model 10. Second, the coefficient for PR systems also remains statistically and
substantively significant (about 5% of GDP) even with
the median voter (and catholic population) entered
into the model. The greater redistributive welfare
spending under PR is not just a product of more leftist
median voters (or, as we shall see, governments). The
substantive magnitude of the PR coefficient is reduced,
as we expect, when we take account of the position of
the median voter. But if we compare the three-step
processes without and with the median voter in the
model (Models 9 and 10), the presence of the median
voter only reduces the greater spending under PR from
5.60% to 5.04% in these full fixed-effects models. (Note
that this is about the magnitude of reduction that we
expected from the fixed-effect error correction Model
2, and is a conservative estimate of these effects, given
that all dummy variables are included.) This implies
that a significant portion of the PR-SMD difference still
remains unexplained. Very substantial parts of the effects of the election rule regimes on redistributive
welfare spending must be found elsewhere than in
the social and economic conditions or the median
voter positions. Thus, Hypothesis 5 is only partly true.
Concluding Comments
Scholarly studies of political representation have examined the correspondence between voters and policymakers, and the responsiveness of the representation
process, in a variety of ways. They have drawn on vote
distributions, surveys of left-right placements by voters
and experts, party manifestos. But they have rarely
examined the responsiveness of public policies themselves to differences or changes in positions of the median
voters. We have used a carefully specified, dynamic error
correction model of redistributive welfare spending to
explore such responsiveness.
The position of the median voter is estimated
from the voter’s electoral choices and the manifesto
positions of the parties. The redistributive welfare
spending is estimated from Franzese’s combination
of government spending statistics. There is no sense in
which the same measurement components appear on
both sides of this basic relationship and create an
artificial connection. Establishing a linkage between
1026
T ABLE 4
shin-goo kang and g. bingham powell jr.
Fixed Effects Vector Decomposition Model
Model 9
Constant
Tax-and-Transfer t-1
PR
DMedian voter t
Median voter t-1
Catholic
DUnemployment t
Unemployment t-1
DAge65+ t
Age65+ t-1
DInflation t
Inflation t-1
D(DGDP t)
DGDP t-1
GDP t-2
Openness t-1
Central-tax share t-1
Indirect-tax share t-1
Total-tax share t-1
Union density t-1
DTax-and-Transfert-1
DWelfare spending trend
t
2PR/ Tax-and-Transfer t-1
2Catholic/ Tax-and-Transfer t-1
2Median Votert-1/Tax-and-Transfert-1
Number of Cases (d.f)
Adjusted R2
Model 10
28.675
20.103
0.575
(2.365)
(0.013)
(0.101)
0.008
0.186
0.004
0.336
0.017
20.028
20.001
212.968
211.689
0.876
20.551
20.331
0.924
0.630
0.009
0.031
0.203
(0.001)
(0.039)
(0.011)
(0.148)
(0.016)
(0.010)
(0.007)
(1.176)
(1.750)
(0.179)
(0.170)
(0.282)
(0.523)
(0.987)
(0.003)
(0.041)
(0.070)
5.604
0.074
(1.023)
(0.013)
501(455)
0.5545
29.901
20.103
0.519
0.003
0.006
0.008
0.184
0.011
0.375
0.027
20.026
0.002
213.260
212.221
0.889
20.583
20.326
1.323
0.454
0.009
0.027
0.196
(2.424)
(0.013)
(0.099)
(0.006)
(0.003)
(0.001)
(0.040)
(0.012)
(0.147)
(0.016)
(0.010)
(0.007)
(1.187)
(1.788)
(0.181)
(0.176)
(0.282)
(0.526)
(0.979)
(0.003)
(0.041)
(0.069)
5.040
0.081
0.058
(0.999)
(0.013)
(0.025)
501(453)
0.5562
Note: Fixed effect vector decomposition models are estimated following Plümper and Troeger’s (2007) suggestion. Standard errors are in
parentheses. The coefficient estimates in Bold indicate statistical significance at p , .05 in two-tailed tests. Both models include the
following political variables as suggested by Franzese (2002):DElection year t, Election year t-1, DGovernment hazard ratet, Government
hazard ratet-1, Income skew variability t-1, Voter turnout t-1, Income skew t-1 and Voter turnout t-1*Income skew t-1. The estimates of these
variables are not reported in the table for ease of reading, but none of them are statistically significant. The results are available upon
request.
election behaviors of voters, as shaped by party
alternatives presented to them, and the policy actions
of government bureaucracies is a fundamental contribution of the paper. It remains for future research,
however, to determine whether these connections will
hold under conditions of greater globalization and
economic interdependence, as well as an altered leftright discourse.
In the large picture our results are reassuring for
policy responsiveness in liberal democracies. Ideological
differences in the position of the median voter consistently and significantly induced responsive changes
in redistributive welfare spending, taking account of a
variety of economic and social conditions. The magnitude of the effects is substantively important as well
as statistically significant. Difference in the positions
of their respective median voters could account, for
example, for about 60% of the average spending
difference between the United States and Sweden in
our 30-year time period. Hypothesis 1 is consistently
supported.
However, there are some surprising connections.
First, despite the greater correspondence between the
positions of voters, legislators, and governments in
(these) PR systems compared to (these) SMD systems,
we find little firm evidence of greater responsiveness in
redistributive welfare spending in the PR countries. In
our interactive models, the PR coefficients are not
statistically different from the SMD coefficients. On
closest inspection the SMD systems might have some
advantage in short-term responsiveness; the PR systems might have some advantage in correspondence of
the long-term equilibria. But generally, these two quite
different processes of representation induce responsive
representation and policy responsiveness
redistributive welfare spending to about the same
(significant) degree. Given the previous research on
the process, and the representational correspondence
here also, this lack of difference in median voter effect
seems surprising.
Moreover, the more liberal orientation of citizens
in the PR systems, evident in Panel (B) of Figure 1,
can account for part, although definitely not all, of
the systematically greater welfare spending in the PR
systems. (The proportion explained depends in part
on whether full fixed effects or only the three country
dummies are used.) That spending difference is
remarkably robust. It is consistent with various
strategic explanations of differences in party behavior
under PR and SMD rules.15
It seems likely that this similarity of responsiveness is related to the other surprise in our results: the
apparently greater responsiveness of welfare spending
policies to the median voters than to the manifesto
promises of the party (or parties) in government. To
be sure, short-term change in government position
has an immediate short-term impact on welfare
spending, a type of effect that we did not see from
voter change (despite some hints in the SMD
systems.) But the long-term equilibrium levels of
welfare spending, which are larger, seem more responsive to the current voters than to the government’s own campaign promises. This greater
responsiveness to the voter is apparent in both the
SMD and PR systems.
Acknowledgments
We are glad to acknowledge advice and assistance
from Neal Beck, Robert Franzese, HeeMin Kim, and
three thoughtful reviewers for the JOP.
Manuscript submitted 23 January 2009
Manuscript accepted for publication 1 March 2010
15
As the right-left gap between governments generated by SMD
and PR systems is greater than the right-left gap between their
respective electorates, we might think that the greater redistributive welfare spending under PR could be more fully explained by
bringing governments into the equations. But following the three
step ‘‘fixed effect vector decomposition’’ model discussed in the
equation only reduces the coefficient for PR from 5.60% in
equations with all variables except median voters and governments, to 4.87% in equations with terms for both short-term and
long-term equilibrium voters and governments. So, the additional reduction is not very much greater than we saw for voters
alone.
1027
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Shin-Goo Kang is Research Professor, Peace &
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G. Bingham Powell Jr. is Professor of Political
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