The Effect of S&P Credit Rating Initiation on Emerging Market Firms
Kee-Hong Bae, Lynnette Purda, and Michael Welker*
August 30, 2007
Abstract
In this paper we examine the impact that initiation of credit ratings by Standard and Poor’s (S&P)
has on firms in emerging markets. We argue that scrutiny from a credible external financial
intermediary such as S&P will impact the information environment of rated firms because S&P will
demand credible, conservative financial reporting as an input to their rating process. We also argue
that examining emerging market firms provides a powerful setting to examine the effects of financial
intermediation. Because S&P is primarily concerned with downside risk to unsecured creditors, we
predict that S&P will demand more conservative accounting practices, i.e., accounting that
incorporates bad news more quickly than it incorporates good news. We document that conditional
accounting conservatism significantly increases in the period after S&P credit rating initiation. We
also document a positive equity market reaction to credit rating initiation and find that the magnitude
of the equity market reaction is related to the change in conservatism, consistent with the idea that
conservative financial reporting benefits other investors by reducing information asymmetry.
Similarly, we find that analyst following increases after S&P credit rating initiation, and again find a
relation between the increase in analyst following and the increase in conservatism. Overall,
initiation of ratings by S&P is associated with increased conservatism that is viewed favorably by
investors.
All authors are from Queen's School of Business, Queen’s University, Kingston, Ontario, Canada.
We thank the Social Sciences and Humanities Research Council of Canada (Grant 410-06-0420), the
Bank of Montreal (Bae) and KPMG (Welker) for generous research support. We are grateful to
Zhefeng Liu, Bill Scott and Dan Thornton for comments on an earlier version of the paper.
* - Corresponding author, mwelker@business.queensu.ca
The Effect of S&P Credit Rating Initiation on Emerging Market Firms
Abstract
In this paper we examine the impact that initiation of credit ratings by Standard and Poor’s (S&P)
has on firms in emerging markets. We argue that scrutiny from a credible external financial
intermediary such as S&P will impact the information environment of rated firms because S&P will
demand credible, conservative financial reporting as an input to their rating process. We also argue
that examining emerging market firms provides a powerful setting to examine the effects of financial
intermediation. Because S&P is primarily concerned with downside risk to unsecured creditors, we
predict that S&P will demand more conservative accounting practices, i.e., accounting that
incorporates bad news more quickly than it incorporates good news. We document that conditional
accounting conservatism significantly increases in the period after S&P credit rating initiation. We
also document a positive equity market reaction to credit rating initiation and find that the magnitude
of the equity market reaction is related to the change in conservatism, consistent with the idea that
conservative financial reporting benefits other investors by reducing information asymmetry.
Similarly, we find that analyst following increases after S&P credit rating initiation, and again find a
relation between the increase in analyst following and the increase in conservatism. Overall,
initiation of ratings by S&P is associated with increased conservatism that is viewed favorably by
investors.
Keywords: credit rating, accounting conservatism, information environment, emerging markets
2
1. Introduction
This paper examines the effect of credit rating initiation by Standard and Poor’s (S&P) on a sample of
emerging market firms. We document that emerging market firms experience increases in conditional
conservatism, share prices, and equity analyst following in the period after S&P credit rating initiation.
We provide some evidence of a relation between the observed changes in conservatism and the equity
price reaction to the initiation announcement and the increase in equity analyst following. Our results
are consistent with 1) a link between debt market driven informational demands and conditional
conservatism, and 2) a link between conditional conservatism and reductions in information asymmetry
that benefit equity market participants as well as debt market participants.
The past literature on conditional accounting conservatism and debt market informational demands has
documented relations between conditional conservatism and firm-level debt contracting for firms in the
United States (e.g., Ahmed et al., 2002; Zhang, 2004; Beatty, Weber and Yu, 2007). This literature has
also documented relations between contractual/debt demands and conditional accounting conservatism
across countries (e.g., Ball, Robin and Sadka, 2006; Bushman and Piotroski, 2006). Our study
contributes to this literature by exploring an ex-ante high power setting to examine how debt market
demands for conservative accounting affect firms’ financial reporting. Emerging markets are known to
show low levels of demand for accounting conservatism and a high degree of flexibility in
implementing accounting standards. Thus, our sample firms from emerging markets are expected to
have little demand driven or regulatory incentive to provide conservative financial reporting. In this
setting, the initiation of credit ratings by a U.S.-based financial intermediary such as S&P that puts
particular emphasis on conservative accounting is expected to be very influential in shaping firm’s
financial reporting decisions.
1
We investigate two research questions. First, we examine whether firms change their financial reporting
practices in response to the informational demands associated with S&P credit rating initiation. We find
a significant increase in firm’s conditional conservatism in the period after the initial rating by S&P,
consistent with rated firms responding to demands by S&P for conservative reporting that facilitates the
assessment of downside risk. The finding of increased conservatism is consistent across three measures
of accounting conservatism, the accrual component of earnings (Bhattacharya et al., 2003), timely loss
recognition measured using Basu (1997) regressions, and conservatism scores as described in Khan and
Watts (2007).
Second, we investigate whether the changes in financial reporting that accompany S&P credit rating
initiation provide benefits to equity market participants as well. LaFond and Watts (2006) show that
conservative accounting practices reduce information asymmetry and should therefore be beneficial to
equity market participants as well as debt market participants. Consistent with their evidence, we find a
positive equity market reaction to the credit rating initiation and a negative market reaction for a smaller
sample for which S&P discontinues providing credit ratings. We argue that the positive equity market
reaction is related to our finding that rated firms provide more conservative financial reporting.
Conservative accounting practices reduce management’s incentives and ability to manipulate reported
earnings and net assets upwards, reducing information asymmetry for equity market participants. We
find that the equity market reaction is related to both conservatism levels before credit rating initiation
and to changes in conservatism when we use working capital accruals as our measure of conservatism.
Finally, as additional evidence that the changes in financial reporting are beneficial to equity market
participants, we show that credit rating initiation results in an increase in analyst following. This
finding corroborates the other two primary findings in the paper. If S&P credit rating initiation prompts
firms to improve their financial reporting by providing more conservative financial reporting, then this
improvement in the information provided by firms will provide greater incentives for analysts to follow
2
the firm. We again find an association between changes in analyst following and conservatism measured
using working capital accruals.
Our study provides new insights into the role of information intermediaries. As Beaver, Shakespeare
and Soliman (2006) note, high profile corporate scandals such as Enron and WorldCom raised questions
about credit rating agencies performance as investment watchdogs since the rating agencies failed to
provide warning of these failures. Beaver et al. (2006) go on to demonstrate significant differences
between the properties of credit ratings provided by certified (e.g., Moody’s, S&P) versus non-certified
(e.g. Egan Jones) rating agencies. Certified agencies such as S&P provide credit ratings that are used in
contractual and regulatory settings, while non-certified agencies provide ratings that serve only an
informational role. Accordingly, Beaver et al. (2006) predict and find that certified agencies provide
ratings that are less timely in response to news generally, are more conservative (ratings that are more
responsive to bad news than to good news), and are slower to move below the investment grade
threshold than ratings provided by non-certified agencies. Their finding that certified agencies act
conservatively provides further support for our conjectured link between S&P (a certified agency) rating
initiation and demands for financial reporting conservatism from rated firms.
Our study also adds to the extensive literature that has investigated the effects of cross-listing equity
shares in foreign markets. The general conclusion of this literature is that cross-listing improves a firm’s
information environment, increasing firm value through positive share price reactions to the listing
announcement (Foerster and Karolyi, 1999; Miller 1999), increasing stock analyst following (Lang,
Lins and Miller, 2003), and resulting in a higher Tobin’s Q (Doidge, Karolyi, and Stultz, 2004). There
are several possible explanations for why cross-listing equity shares could be economically beneficial
for the firm. First, and perhaps most simply, cross-listing raises the visibility of the firm to U.S.
investors, increasing the pool of potential investors, stimulating demand for the company’s stock and
driving up the share price (Merton, 1980; Foerster and Karolyi, 1999). Second, U.S. cross-listing may
3
expose the firm to a variety of regulatory consequences that could affect value1. For example, crosslisted firms may face increased litigation risk and be forced to be more transparent due to SEC
disclosure standards and requirements to provide financial statement data prepared in accordance with
or reconciled to U.S. GAAP.
Related to these regulatory effects, cross-listed firms may increase voluntary disclosures and provide
higher quality information to stimulate interest from investors and financial intermediaries such as
financial analysts. Firms willing to subject themselves to this additional scrutiny may be able to
effectively signal their high quality and separate themselves from lower quality firms in their home
country, thereby enhancing firm value. Consistent with the idea that cross-listing is associated with
increased transparency and monitoring, Doidge et al. (2005) find that firms with higher control rights
for the controlling shareholder and a divergence between the cash flow and control rights of the
controlling shareholder are less likely to cross-list their shares. This finding suggests that the controlling
shareholders of these firms do not want to suffer the decrease in their private benefits of control that
may accompany greater transparency and monitoring. Lang, Raedy and Yetman (2003) provide
evidence suggesting that cross-listed firms engage in less earnings management to meet earnings targets,
report more conservative earnings, and exhibit stronger associations between earnings and share prices,
all consistent with the idea that accounting quality is improved by cross-listing.
In this paper, we do not seek to directly contribute to the cross-listing literature. Rather, we aim to
contribute to the literature more generally by examining an alternative way for foreign firms to be
subject to U.S. financial intermediation. The cross-listing literature suggests that an examination of
firms from emerging markets that choose to subject themselves to the scrutiny of financial
1
Of course, these consequences are a function of how the firm chooses to cross-list. If the firm cross-lists using a
level II or II ADR and is listed on an organized US stock exchange (i.e. NYSE, AMEX or NASDAQ), then the
firm is subject to SEC oversight and US style disclosure requirements. Firms that cross-list in the OTC market
(level I ADR) or under Rule 144a do not subject themselves to the same regulatory scrutiny or the same disclosure
requirements.
4
intermediaries from a more developed market such as the U.S. provides a setting where the effects of
financial intermediation are expected to be large, resulting in ex-ante high power tests.
In summary, our results make several contributions to the literature. First, ours is the first study to
examine the effects of credit ratings on emerging market firms, filling an important void in the
international finance and accounting literatures. Second, our results add to the cross-listing literature by
showing that firms subjecting themselves to scrutiny from S&P experience changes in their information
environment that closely resemble the effects of cross-listing equity shares. Our results add to the
international accounting literature by pointing out that in addition to cross-listing equity shares, credit
ratings by U.S. agencies also cause significant changes in the financial reporting practices of firms
internationally. Our paper also responds to the call of Fan and Wong (2005) for more research on
governance mechanisms affecting emerging market firms. Finally, we contribute to the growing
literature on accounting conservatism by showing a link between credit market information demands
and conservatism in a powerful time-series setting, and by documenting that increased conservatism
benefits equity market participants as well as credit market participants.
The remainder of the paper is organized as follows. Section 2 details the related literature and outlines
the motivation for our empirical tests. Section 3 describes the credit rating process for emerging market
firms. Section 4 presents the results of tests for changes in conditional accounting conservatism after
S&P credit rating initiation. Section 5 presents stock price reactions to the announcement of credit rating
initiation. Section 6 presents the results of tests for changes in equity analyst coverage following credit
rating initiation. Section 7 provides a summary, discussion, and outline for further research.
5
2. Hypothesis Development
That credit rating agencies rely on accounting information and are important users of a firm’s financial
statements is unsurprising (Kaplan and Urwitz, 1979; Ziebart and Rieter 1992). More relevant for this
paper is the recent evidence on the importance that these agencies place on the quality of information
provided by a firm’s financial reports. Ashbaugh-Skaife, Collins, and LaFond (2006) study the extent to
which corporate governance, including financial transparency which reduces the information asymmetry
between borrowers and lenders and facilitates the monitoring of management’s actions, is positively
viewed by rating agencies. They find that two measures of transparency, the quality of working capital
accruals and the timeliness of earnings, are positively related to bond ratings. Consistent with this
argument, they document that bond ratings are higher for firms with greater earnings predictability.2
While it is clear that rating agencies assess the quality of a firm’s financial reporting when establishing
its creditworthiness, it is also likely that a firm will alter this quality in response to the demands of the
credit rating agency. This may particularly be the case for firms in emerging markets where accounting
standards and their enforcement are generally of low quality. Crabtree and Maher (2005) argue that debt
holders are primarily focused on downside risk and should therefore value conservative financial
reporting. S&P’s Corporate Rating Criteria (2006) supports this argument, emphasizing that one
organizational problem they look for is whether “The company is particularly aggressive in the
application of accounting standards…” (p.23). This supports Watts’ (2003a, 2003b) argument that
conservative financial reporting is in part driven by the demands of the credit market. We hypothesize
that firms will respond to scrutiny of their financial statements by S&P by reporting more
conservatively. In this section we formalize this hypothesis and its implications for emerging market
2
In addition to the literature relating accounting quality to bond ratings it should be noted that a significant body
of research has developed to relate properties of financial information to bond yields. Sengupta (1998) documents
that firms perceived to have higher disclosure quality achieve a lower cost of debt while Yu (2005) finds more
transparent firms have lower credit spreads.
6
firms. We also develop related conjectures for how this change in financial reporting will impact other
stakeholders such as equity investors and stock analysts.
Beginning with Basu (1997), a large number of papers have examined the phenomenon of accounting
conservatism, interpreted by Basu as the tendency to require a higher degree of verification for
recognizing good news than bad news in financial statements. Basu (1997) notes that conservatism leads
to the empirically testable proposition that accounting earnings reflect bad news more quickly than good
news. Following Watts (2003a, 2003b), several recent studies examine the role of debt contracting as a
primary source of the demand for conservative accounting (Bushman and Piotroski, 2006; Ball, Robin
and Sadka, 2005; and Beatty, Weber and Yu, 2007). The argument linking debt market demand for
information and accounting conservatism is quite intuitive. Debt market participants face asymmetric
payoffs with little or no ability to reap the benefits of economic gains experienced by the debtor
company but potential for losses accompanying economic difficulty experienced by the debtor that
affect the likelihood or magnitude of debt repayment. In this setting, a key role of accounting is to
provide information to facilitate a conservative assessment of the debtor’s ability to repay debt. The
results of Beatty, Weber and Yu (2007) show that contractual modifications to accounting numbers only
partially fulfill creditors’ demands for conservative accounting, and that conservatism in financial
reporting and contractual modifications are complementary ways for creditors to manage risk.
These arguments suggest that the impact of credit rating initiation on the information that firms provide
will be observed in more conservative reporting. The effects of demand influences on financial
reporting practices are expected to be strong in our setting because in the absence of demands for
conservative reporting, the emerging market firms have both the incentive and the flexibility to avoid
conservative reporting. For example, our emerging market sample countries tend to have civil law legal
origins (14 countries) and low levels of judicial impartiality (16 countries), which are both
characteristics associated with low levels of conservatism (Bushman and Piotroski, 2006). In addition,
7
our sample countries have on average poorly developed debt markets, which are also expected to result
in low levels of conservatism (Ball, Robin and Sadka, 2005).
We also conduct tests to look for increases in equity values and the number of equity analysts following
the firm after S&P credit rating initiation. While credit market demands may be a primary source of the
demand for conservative financial reporting (Bushman and Piotroski, 2006, Ball, Robin and Sadka,
2005), recent research suggests that conservative financial reporting may also benefit equity market
participants because it reduces information asymmetry. LaFond and Watts (2006) argue that
conservative accounting reduces incentives and flexibility to manipulate accounting numbers.
Conservatism requires there be greater verification and certainty before a gain can be recorded than is
required for recording a loss, restricting the incentives that managers may have to overstate profits and
net assets and therefore reducing information asymmetry. While their primary hypotheses examine how
information asymmetry in equity markets can lead to a demand for accounting conservatism, we test for
a different implication of their framework. In our setting, it is the new demand from S&P that we
expect to increase accounting conservatism, and this increase in conservatism is expected to reduce
information asymmetry. The reduced information asymmetry in turn will result in increase in equity
values and analyst following.
While some international work exists on investors’ reactions to rating initiation (Barron, Clare and
Thomas, 1997) and rating changes for corporate bonds (Matolcsy and Lianto, 1995; Steiner and Heinke,
2001), most has been conducted in the context of well-developed financial markets. In the emerging
markets setting, almost all research focuses on sovereign ratings rather than firm-level credit
assessments.3 In particular, Brooks, Faff, Hiller and Hiller (2004) examine the stock market reaction to
sovereign rating changes and find that announcements from S&P elicit the most significant price
3
One exception is the work of Ferri (2004) which examines the number of credit analysts assigned to emerging
market countries by Moody’s Investors Service.
8
response. Kaminsky and Schumkler (2002) and Gande and Parsley (2005) conduct similar studies but
ask if the impact of rating changes for one country extend to neighbouring nations or those with
institutional links to the country. Both studies confirm that sovereign rating changes in one country have
spillover effects on bond markets in other nations. Kaminsky and Schumkler (2002) establish that these
effects are particularly large in times of crisis while Gande and Parsley (2005) link the size of spillovers
to the trade and capital flows between countries. Our study differs substantially from this limited prior
literature because we focus on the effects of credit-rating initiation at the firm-level for firms from
emerging markets and because we examine primarily the effects on the firm’s information environment.
3. S&P Ratings in Emerging Markets
U.S.-based rating agencies have expanded into foreign markets in a variety of ways. In some cases they
have acquired local rating agencies outright (for example S&P’s acquisition of the Canadian Bond
Rating Service and Australian Ratings) while in others they have opened branch offices or partnered
with foreign firms (Lyons, 1996). Regardless of how they have entered, expansion into foreign markets
has been fast-paced. Figure 1 shows that there were very few corporate ratings maintained by S&P in
emerging markets in 1993. However, as of year-end 2004, over 700 ratings in foreign countries were
outstanding. In this paper, we concentrate on 18 emerging markets that encompass the vast majority of
emerging market ratings.
Typically, a firm approaches the rating agency in order to secure a rating. While unsolicited ratings may
be assigned by an agency, these instances are relatively rare. Kliger and Sarig (2000) report that
approximately 98% of rated U.S. firms pay for their ratings. Paying for a rating allows the issuer to be
actively involved in the rating process and may entitle management to comment on a preliminary rating
before it is publicly announced (Cantor and Packer, 1994). While ratings clearly involve elements of
subjectivity, management can refer to the publicized guidelines from the agencies detailing the factors
9
that will be considered throughout the rating process. These factors include, among others, financial
performance, an evaluation of management and an assessment of the industry. For emerging market
firms, S&P has additional considerations (see pages 37-41 of S&P’s Corporate Ratings Criteria, 2006
for a discussion of country risk assessment). Primary among these is the sovereign rating of the nation
in which the firm resides. In the vast majority of cases, a firm will not be rated above the sovereign.
While the distinction between emerging and developed markets is not always clear, S&P defines a
country with a sovereign rating below investment grade (below BBB) as an emerging market. 4
Therefore, it is not surprising that Figure 2 shows that the majority of firms within our sample received
an initial rating that was below the investment grade threshold. While one firm entered the sample with
a rating of AAA, the median initial rating was only BB and the lowest rating was CC. Other factors
deserving special consideration in emerging markets are extensive but a partial list includes exchangerate risk, government regulation, legal issues, changing tariff barriers, access to capital, and liquidity
restrictions.
Table 1 provides details of our sample firms according to their country of domicile and year of initial
rating. The sample represents all firms rated by S&P in these countries with the necessary data to be
included in some or all of our empirical analysis. The country with the most extensive coverage in the
sample is Korea, followed by Taiwan, Mexico and Brazil . Portuguese firms had the best median initial
rating at A- while Russian and Turkish companies fared the worst with a median rating of B+. Rating
initiations in the sample peaked in 1996 and 1997 with significant growth in these two years. In total,
the sample consists of 209 unique rated firms.5
See “Corporate Utility Credit Analysis in Emerging Markets” S&P publication October, 1999.
While the full sample of rated firms in emerging markets represents 209 firms from 18 countries, the sample of
firms used in the empirical analyses is much smaller ranging from 30 to 180, depending on the availability of data.
The number of rated firms from our sample countries reflected in Figure 1 is much greater than 209 because the
graph depicts all rated firms without regard to data availability.
4
5
10
4. Tests for changes in conditional accounting conservatism
Several papers have examined attributes of accounting quality across countries, companies or time.
There is no clear consensus about which measures provide the most powerful tests for changes in
accounting quality through time or for differences in accounting quality across countries or companies.
Commonly used measures in the recent literature include: 1) earnings smoothing, 2) loss avoidance
(sometimes referred to as earnings management to avoid losses), 3) various measures of accruals quality,
and 4) various measures of accounting conservatism, (e.g., Leuz, Nanda and Wysocki, 2003;
Bhattacharya, Daouk and Welker, 2003, Francis et al., 2005; and Barth et al., 2006).
As discussed earlier, both the prior academic research on accounting conservatism and the credit rating
criteria published by S&P indicate that in the context of credit ratings the most likely influence on
financial reporting is increased conservatism. While it is possible that the demand for improved
information from S&P would affect other aspects of the firm’s information, such as an increase in the
amount of voluntary disclosure, we focus on the effect on accounting conservatism since the extant
evidence suggests a clear link between the information needs of credit market participants and
conservative accounting.
4.1 Accrual-based measures of conservatism
Accounting earnings consist of two primary components, cash flows from operations and accruals
which result from accounting adjustments that recognize in earnings economic events and transactions
either before or after the cash flow related to the event or transaction occurs. Accounting conservatism
is expected to result in earnings reflecting bad news more quickly than good news, i.e. unrealized losses
are recognized in earnings more quickly than unrealized gains. The recording of unrealized losses
11
occurs through the accrual component of earnings – e.g. inventory is written down to lower of cost or
market value by decreasing inventory which does not affect current period cash flows. As a result,
ceteris paribus, relatively more conservative accounting recognition should result in a lower accrual
component of earnings (Bhattacharya et al., 2003). Accordingly, our first and most simple way to assess
accounting conservatism is to examine the accrual component of earnings.
Because not all our sample companies provide cash flow statements that could be used to separately
determine the accrual and cash flow components of earnings, we use the “balance sheet” approach,
inferring the accrual component of income from changes in account balances from the balance sheet. In
particular, we measure the total accrual component of earnings as:
TACC  Current Assets - Cash - Current Liabilitie s 
Current Portion of L  T Debt  Income Taxes Payable  Depreciati on exp ense
(1)
where firm and time subscripts are omitted for simplicity. We begin our analysis by estimating the
following equation:
TACCi ,t   0  1 ( POSTRATE ) i ,t   2 ( ADR) t   3 ( POSTADR ) i
  4 ( ASIA) i   5 ( POSTASIA) i ,t   6 (YR ) t   i ,t
(2)
Since accounting conservatism results in more timely recognition of income decreasing events, total
accruals are expected to be lower in more conservative accounting systems. We also replace TACC with
a measure that includes only the working capital accruals. We refer to this measure as WCACC and it is
equal to TACC + Depreciation expense, so it only includes the accruals related to changes in current
asset and current liability accounts.
12
We create several dummy variables to test our hypotheses and to control for other factors that could
influence conservatism. We include a dummy variable, POSTRATE, equal to one for all observations
that occur after the initial S&P credit rating.6 POSTRATE measures the change in accruals occurring
after the rating date. We include a variable that is set equal to zero in 1991, the initial sample year, and
increases by one for each calendar year in the sample period (YR) to control for sample-wide time
trends in our dependent variables. This ensures that POSTRATE does not just capture general timetrends in the data. Since cross-listing is another decision that may impact the information environment,
and conservatism in particular, we control for cross-listing by including a dummy variable for all firmyear observations for firms that are cross-listed in the United States (hereafter ADR) and another
dummy variable for firm-year observations that follow the decision to issue an ADR (hereafter
POSTADR). The ADR data come from the Bank of New York website. These two variables capture
the differences between cross-listed firms and other firms and capture the informational effects of crosslisting. Finally, our sample consists of firms from emerging markets around the world and includes a
substantial number of Asian firms. In addition, our sample period spans the Asian currency crisis of
1997. We introduce two new dummy variables to capture differences between Asian firms and other
sample firms both before and after the Asian currency crisis. ASIA is a dummy variable equal to one
for all observations from Asian firms, and POSTASIA is a dummy variable equal to one for all
observations from Asian firms from 1998 or later.7
Our primary experimental variable is POSTRATE, and we predict that firms will utilize more
conservative accounting practices that lower the accrual component of earnings, resulting in a negative
coefficient on this experimental variable. In addition, we report the results of the regression with two
additional controls added, firm size (SIZE), measured as market capitalization in U.S. dollars at the
6
We later report the results of sensitivity tests that examine whether firms shift to more conservative accounting
practices before the initial credit rating. This sensitivity test allows for the possibility that firms anticipate both the
request for a credit rating and S&P’s preference for conservative accounting and alter their reporting behaviour
before receiving the initial rating.
7
We code POSTASIA as one for years beginning in 1998 because the currency crisis occurred late in 1997.
13
beginning of the fiscal year, and the price to book ratio (PB), also measured at the beginning of the
fiscal year. To control for the potential effects of outlying observations, we include the percentile rank
of both variables in the regressions. We include these two firm characteristics to control for their
potential effects on accounting conservatism.
We collect the accounting data necessary to calculate the accrual component of earnings and cash flows
from Worldscope. We include all rated firms from our sample countries for which all data necessary to
calculate accruals and cash flows are available during 1991 through 2004, though we lose a substantial
number of our potential sample firms due to lack of the necessary data. We exclude non-rated firms
from our sample countries in our primary tests because we expect that firms self-selecting to be rated by
S&P differ from their country-peers in many ways, and because our tests focus on changes in dependent
variables in the post-rating period for rated firms. As we discuss in more detail later, our results are
qualitatively similar if we focus our analysis on all firms, whether rated or not. Because our sample
firms are from emerging markets with potentially high and variable inflation rates during our sample
period, our main results utilize balance sheet data converted to U.S. dollars. Our results are qualitatively
similar if we use data presented in local currencies. Because we are working with emerging market data
and want to ensure that outliers that could reflect data errors are not responsible for our results, we
truncate the top and bottom 1% of our sample observations on TACC and WCACC.8 This yields a
sample of 994 observations from 105 rated firms.
Table 2 shows descriptive data for this sample of firms and highlights the differences between rated
firms and non-rated firms. Rated firms have lower levels of both TACC and WCACC and higher levels
of cash flow from operations (CFO). The first two results are consistent with greater conservatism for
rated firms, while the last result suggests that rated firms are financially better performers. Rated firms
8
Unless otherwise noted, all of our results are qualitatively similar if we delete the top and bottom 5% of our
sample observations.
14
are much larger than non-rated firms, and have higher P/B ratios consistent with rated firms having
higher valuations and better growth opportunities. Rated firms are much more likely to be cross-listed,
with 60% of rated firms also having equity cross-listed in the United States.
We present the results of estimating equation (2) in Table 3. Columns 1 and 2 present the results
without the SIZE and PB controls for TACC and WCACC, respectively, while columns 3 and 4 show
the results with SIZE and PB added to the equation. All of our regressions include firm-level clusters
because each firm can contribute multiple observations to our analysis. Controlling for firm-level
clustering avoids understatement of standard errors due to the lack of independence in observations at
the firm level. Neither ADR nor POSTADR has a consistently significant relation with TACC, but there
is some evidence of lower WCACC for cross-listed firms. For rated firms, ASIA and POSTASIA are
unrelated to accruals. Consistent with the evidence in Pae, Thornton and Welker (2005), higher price to
book ratios are associated with higher working capital accruals. However, PB is not related to TACC.
There is no trend in accruals over our sample period.
Most importantly, POSTRATE is significant and
negative in all four columns of Table 3, consistent with our prediction that firms employ more
conservative accounting practices in response to S&P’s use of their financial statements in the rating
process.
The results of this analysis continue to strongly support an increase in conservatism in the post-rating
period for rated firms when we repeat the analysis using observations from all firms, whether rated or
not (results not tabulated). In this analysis, we include one additional dummy variable, RATE, which
takes on a value of one for all observations from rated firms and is intended to capture differences
between rated and non-rated firms that exist prior to the rating. In addition to showing continued support
for our hypothesis, these tests do reveal some interesting and intuitively appealing properties of
accounting conservatism that are not visible in our smaller sample of rated only firms. The full sample
results suggest that rated firms had similar levels of accruals to other firms prior to the initial S&P credit
15
rating. In the full sample, there is a significant increase in conservatism after cross-listing, low levels of
conservatism for Asian firms prior to the Asian currency crisis, and an increase in conservatism for
Asian firms after the currency crisis.
We also conducted tests based on the relation between WCACC and CFO as described in Ball and
Shivakumar (2005). In brief, Ball and Shivakumar argue that while in general there is a negative
relation between WCACC and CFO, conservatism would result in an offsetting positive relation
between WCACC and CFO when the latter is negative. We therefore estimate an extended version of
equation 2 that includes CFO, and operating cash flows times a dummy variable that is equal to one
when operating cash flows are negative (DCFO). These terms are then interacted with each of the
variables included in equation 2. Consistent with our prediction of greater conservatism, we find a
significantly positive coefficient on POSTRATE*CFO*DCFO on the rated sample when we either
delete no outliers or trim outliers at the 1% level. However, the rated firms are generally performing
well, and fewer than 10% of the rated firms experience negative CFO years during our sample period,
so this result is sensitive to our outlier decision rule. If we delete 5% of the observations from the lower
tail of the cash flow distribution, the coefficient on POSTRATE*CFO*DCFO is no longer significant.
Because of the lack of negative cash flow observations for the rated sample, we do not tabulate these
results but do note that these results are generally consistent with our expected increase in conservatism
in the period after rating initiation.
4.2 Timely loss recognition measure of conservatism
The next measure of conservatism we use comes from Basu (1997) and is often referred to as timely
loss recognition. Basu (1997) shows that conservatism in accounting recognition results in a predictable
and strong empirical regularity: Earnings are much more strongly associated with the news reflected in
abnormal stock returns when those returns are negative (bad news) than when they are positive (good
16
news).
He demonstrates this empirical regularity by estimating the following regression relating
accounting earnings to equity returns:
EPSi ,t   0  1 ( RET ) i ,t   2 ( DRET ) i ,t   3 ( RET * DRET ) i ,t   i ,t
(3)
where EPS is annual earnings per share divided by stock price at the beginning of the year, RET is the
stock return computed as the holding period return from 12 months before the fiscal-year end to three
months after the fiscal-year end less the corresponding market returns, DRET is a dummy variable equal
to 1 if returns are negative and 0 otherwise. Conservative accounting implies a positive β3 coefficient
relating stock returns to earnings when returns are negative, consistent with earnings being more likely
to incorporate current period bad news than current period good news. We add to this basic model the
same six variables that are included in equation (2) explaining the level of accruals – i.e. POSTRATE,
ADR, POSTADR, ASIA, POSTASIA, and YR. In addition to adding these terms directly to the model,
we also add interaction terms with RET, DRET, and RET*DRET for each variable. In each case,
accounting conservatism is reflected in the interaction terms involving RET*DRET, so we focus our
attention on the coefficient estimates for these terms. Our primary prediction is that conservatism
increases after the initial S&P credit rating, i.e. that the coefficient on (POSTRATE*RET*DRET) is
positive. We again present results with additional terms added to control for SIZE and PB.
Consistent with the research design employed in our earlier analysis, we test for changes in
conservatism using the sample of rated firms. For this analysis, we again use returns, earnings and
prices in U.S. dollars to control for the effects of inflation.9 We again truncate the top and bottom 1% of
our sample for both EPS and RET, yielding a sample of 150 firms and 1,300 firm-year observations.
The sample with the controls for SIZE and PB consists of 150 firms and 1,285 firm-year observations.
EPS ranges between -3.572 and 1.194 and has a mean of 0.040 and a median of 0.066. RET ranges
between -1.022 and 2.949 and has a mean of 0.323 and a median of 0.280.
9
Again, our results of this analysis are qualitatively similar if we use returns, earnings and prices in local
currencies.
17
Table 4 presents the results of estimating the expanded equation 4. Column 1 presents the coefficient
estimates and column 2 the t-statistics from the estimation without the controls for SIZE and PB while
columns 3 and 4 show the results with these controls added. Recall that in this specification,
conservative accounting implies a positive coefficient on RET*DRET. The results in column 1 show
that the coefficient on this term is not significantly different from zero. However, when we control for
SIZE and PB, the coefficient on this term is significantly positive at the 1% level. The coefficient on
POSTRATE*RET*DRET, our primary experimental variable is positive in both columns 1 and 3. The
evidence is consistent with the idea that our sample firms practice more conservative accounting after
the initial S&P credit rating. The only control variables significantly associated with conservatism are
SIZE and PB, both of which have negative relations with conservatism. Our results are qualitatively
similar if we trim outliers at the 5% rather than 1% levels.
4.3 Khan and Watts (2007) Conservatism Scores
We also repeat our tests of accounting conservatism using the firm-year level measure of conservatism
introduced by Khan and Watts (2007). Their measure of conservatism is related to the timely loss
recognition measure introduced by Basu (1997), but utilizes a two-step procedure and instrumental
variables to construct a firm-year measure of conditional conservatism. In the first stage, the RET and
RET*DRET variables from equation 4 are interacted with three instruments, SIZE, PB and leverage as
measured by the ratio of total debt to total assets. This augmented regression is then estimated crosssectionally on an annual basis to determine regression coefficients. For their measure of conservatism,
there are four coefficients of interest, the coefficient on RET*DRET, and the coefficients on the
interactions of this term with each of SIZE, PB and leverage. The firm-level measure of conservatism
(dubbed the conservatism score or “cscore”) is then constructed by multiplying these four regression
coefficients by one and the firm year value of SIZE, PB and leverage, respectively. We adopt this
18
procedure and utilize the full sample for purposes of running the initial regressions to establish
regression coefficients, and then examine the behavior of the cscores for our rated firms. We truncate
the top and bottom 1% of sample observations of EPS, RET, SIZE, PB and leverage. For consistency
with Khan and Watts (2007), we do not use percentile ranks of the explanatory variables in this
regression and we use the end of year rather than the beginning of the year PB ratio.
The results of this analysis are presented in Table 5. Panel A reports the distribution of the regression
coefficients from the first stage regressions across our sample years, panel B provides descriptive
information on the resulting cscores, and panel C shows the results of the regression explaining
variation in cscores for our sample firms. As with our prior tests, the regression results in panel C
include six explanatory variables (POSTRATE, ADR, POSTADR, ASIA, POSTASIA, and YR) and we
present the results for this regression for our sample of rated firms only. We do not run regressions with
controls for firm specific variables such as SIZE and PB as we did in our earlier tests because these firm
specific characteristics are included in the first stage regressions.
The regression coefficients in panel A are all statistically significant. The signs of the coefficients are
also consistent with the signs of the coefficients reported for a sample of U.S. firms by Khan and Watts
(2007), with one exception. For our full sample of emerging market firms, the coefficient on
DRET*RET is negative, indicating that our full sample of emerging market firms is aggressive, not
conservative, in financial reporting. This result is consistent with our assertion that emerging markets
are characterized by little demand from investors or regulators for conservative accounting practices.
The descriptive statistics on the cscores themselves provided in panel B show that the distribution of
cscores for our rated emerging market firms is similar to the distribution of cscores for U.S. firms
reported in Khan and Watts (2007), though our scores are predictably lower reflecting less conservatism.
19
Our mean (median) cscore is .046 (.040) versus the mean (median) score reported in Khan and Watts
(2007) of .093 (.082).
The regression results reported in panel C are again consistent with our primary hypothesis. The cscores
increase with conservative accounting, and the coefficient on POSTRATE is positive and significant.
Cscores for ADR firms suggest less conservative reporting for cross-listed firms, and the positive
coefficient on YR is consistent with a secular increase in conservatism for our sample firms. Other
control variables are not associated with cscores.
In summary, all three tests for changes in accounting conservatism result in a similar conclusion.
Because S&P is primarily concerned with assessing downside risk to unsecured creditors, S&P is
particularly interested in financial reporting that quickly signals declines in the firm’s earnings prospects
and ability to repay debt. This translates into a demand for accounting conservatism, and our tests all
show that firms respond to that demand by shifting to more conservative accounting practices in periods
after they receive their initial credit rating from S&P.
4.4 Sensitivity tests
As indicated earlier in our paper, the inferences from our analysis are similar if we 1) truncate outliers at
the 5th and 95th percentile of the distribution rather than the 1st and 99th percentiles, or 2) include all
firms from our sample countries, whether rated or not, and include a dummy variable, RATE, equal to
one for all observations from rated firms. In addition, we performed three further tests to assess the
sensitivity of our results to sample composition and to our definition of POSTRATE.
20
4.4.1 – Eliminating firms using U.S. GAAP or IFRS
Our analysis includes all rated firms, whether the rated firms used their local accounting standards or
use an alternative set of standards such as U.S. GAAP or International Financial Reporting Standards
(IFRS). Since we believe that the weak investor demand and regulatory oversight will result in minimal
incentives for emerging market firms to provide conservative reporting prior to the initiation of S&P
credit ratings no matter what set of accounting standards the firms use, we think it is appropriate to
include all rated firms in our analysis.
If, however, the use of U.S. GAAP or IFRS constrains
management’s ability to respond to demand forces, or if these standards result in higher levels of
conservatism in the absence of demand effects, then including firms in our sample using U.S. GAAP or
IFRS could weaken the observed relation between S&P credit rating initiation and changes in
conservatism. It is also possible that firms switch accounting standards around the time of the initial
credit rating by S&P and that our main results may be due to changes in accounting standards.10 To
check the effects of including firms using GAAP other than that of their home country, we collected
data on accounting standards from Worldscope. Few of our sample firms use these two sources of
GAAP rather than their local standards. Only nine of our sample firms ever use IFRS during the sample
period and only eight ever use U.S. GAAP during our sample period. Dropping observations that are
not prepared using local GAAP has no impact on the inferences from our study.
10
We note, however, that of the six firms that switch to either US GAAP or IFRS during our sample period (as
opposed to using these alternative standards throughout our sample period), two switch one year after credit rating
initiation, two switch 2 years after credit rating initiation, and two switch prior to the credit rating initiation,
suggesting that our observed changes in conservatism after the initial rating are not likely to be caused by changes
in accounting standards.
21
4.4.2 – Eliminating firms that do not have data both before and after the initial rating
Our main tests include all rated firms with available data during our sample period, and we do not
require that the firms have data available both before and after the initial credit rating. While we believe
that our controls for firm characteristics and time trends make it appropriate to include all available
observations, it is still possible that our results could be affected by minor changes in the composition of
the sample between the pre and post rating periods. Accordingly, we restricted our analysis to firms that
had at least one observation for both the pre and the post rating periods. There is no change in the
inferences from our study from this change in sample composition.
4.4.3 – Recoding POSTRATE as one from period t-1 forward
Our analysis so far assumes that firms begin to utilize more conservative accounting practices after S&P
begins providing credit ratings. However, it is also possible that firms anticipate that they will seek a
credit rating and that S&P prefers conservative reporting. If firms anticipate these two things far enough
in advance then the firm may become more conservative prior to obtaining their initial rating. To test for
this possibility, we repeat our analysis but code POSTRATE as one from the period prior to the initial
rating forward. If firms become more conservative in anticipation of receiving a credit rating, our results
should be stronger with this new coding. With both accruals measures and cscores as our dependent
variables, we do find somewhat stronger results with this new coding. Further, we look at the timeseries pattern of both accruals measures and cscores and find that in both cases there is a change to more
conservative accounting between period -2 and period -1. Both of these results are consistent with firms
adopting more conservative accounting practices the year before they receive their initial credit rating,
presumably in anticipation of the demand for more conservative accounting by S&P. However, the
results of our timely loss recognition regressions are weaker with the recoded POSTRATE variable.
Overall, these tests provide some support for the idea that firms anticipate S&P’s preference for
22
conservative reporting and respond to it prior to obtaining their initial credit rating, but this conclusion
is not supported across all of our conservatism measures.
5. Tests of Market Reaction to Credit Rating Initiation
In this section, we examine the equity market reaction to the event of credit rating initiation by S&P. As
discussed earlier, the increase in conservatism documented in the last section is expected to reduce
information asymmetry for all investors, not just creditors. This reduction in information asymmetry is
expected to lower information risk and increase equity values. We use standard event study
methodology to assess the equity market reaction to rating initiation announcements for emerging
market firms. We obtain the rating initiation announcement dates from S&P and implement the test
procedure by computing ex post abnormal returns as
ARit  Rit  (ˆ i  ˆi Rmt )
(4)
where Rit and Rmt are the daily return of the sample firm i at time t and the daily market index return at
time t, respectively. The coefficients ̂ i and ̂ i are ordinary least squares estimates of the intercept and
slope, respectively, of the market model regression. To compute the abnormal returns, we estimate the
parameters ̂ i and ̂ i with an ordinary least squares regression, using 200 daily returns beginning with
day t = -220 and ending with t = -21 relative to the announcement date.
We construct the cumulative abnormal returns (CARi) between any two dates T1 and T2 as
CARi (T 1, T 2) 
T2
 AR
it
t T 1
23
(5)
and we compute the sample cross-sectional average cumulative abnormal returns as
ACAR(T 1, T 2) 
1
N
N
 CAR (T1,T 2) .
i
(6)
i 1
We use t-statistics to test the hypothesis that the average CARs over any given interval are equal to zero.
As a check on the possibility that the mean return is unduly influenced by outlier returns, we also use a
median test of the null hypothesis. We use a sign-rank test statistic to test the hypothesis that the CARs
over any given interval are distributed symmetrically around zero.
We obtain daily stock prices of individual firms and indices in local currency from Datastream
International. Daily prices are adjusted for dividends and capital distributions. Datastream returns are
unavailable or seriously incomplete until the mid-1990s. For this reason, we end up with 173 sample
firms for which we have available returns data.
Figure 3 presents the cumulative abnormal returns from day -20 to day +50 relative to the day that the
rating was first announced. Stock prices start rising well before the announcement date. Starting from
day -15, stock prices rise and continue to rise up to day +30 and level off thereafter. Panel A of Table 6
reports the cumulative abnormal returns in various windows around the rating event and tests the
significance of announcement reactions. The average CAR over the twenty-day window from day -20 to
day -1 is 2.27 percent and significant at the one percent level. The median CAR is also positive and
significant, although the magnitude is smaller at 1.09 percent. As seen from the figure, this result shows
that there is some leakage of news before the announcement. The average CAR over the two-day
window of 0 to 1 is 0.74 percent and significant at the five percent level. The median CAR is also
significantly positive but the magnitude is smaller at 0.13 percent. The average CAR(-1, 1) and CAR(-2,
24
2) are 0.96 and 1.43 percent, respectively. The median CARs show a similar pattern. These CARs are all
significant at the five percent level. Interestingly, the average and median CAR(1, 20) are also
significantly positive at 2.16 and 1.30 percent, respectively. The average and median CAR(-20, 20) and
CAR(-20, 50) are all significant at the usual significance level. While not reported, we also examine the
abnormal returns estimated from the market-adjusted model as a robustness check. The results are, if
anything, even stronger.
We also examine the market reaction to the news announcements that S&P decides to withdraw its
credit rating, which we call a “de-rating” event. While we could find only a relatively small number of
such events, they allow us to see if the effect of de-rating by S&P is opposite in sign to the effect of
initiating credit rating. One should expect to observe the opposite sign if the positive market reaction to
the rating initiation event is truly driven by the rating itself. Panel B of Table 6 presents the results.
Consistent with the prediction, cumulative abnormal returns over various event windows are all
negative. The average values of CAR(0,1) and CAR(-1,1) are -1.31 percent and -2.60 percent,
respectively, which are both significant. While the mean CARs are all negative, statistical significance
drops for other event windows. None of the medians are significant, perhaps due to the small sample
size.
Overall, our results show that the rating initiation announcement produces a positive response in the
equity market. The positive announcement reaction is consistent with the argument that by getting rated
by a reputable U.S. rating agency and thereby submitting to enhanced scrutiny and monitoring, a
company improves its information environment. We also conduct an additional test to determine if there
is a cross-sectional relationship between changes in conservatism and the equity price reaction we
document. If, as argued by LaFond and Watts (2006), conservatism reduces information asymmetry
and this reduction in information asymmetry benefits equity holders, and if equity holders are able to
rationally predict which rated firms will experience the greatest increase in conservatism, then we
25
should observe a relation between the equity price reaction and the change in conservatism. We explore
two possible relations between accounting conservatism and the equity price reaction. First, if investors
anticipate that the firms most likely to benefit from greater conservatism are the firms with the least
conservative reporting prior to the credit rating initiation, then less conservative accounting prior to the
rating event should lead to a more positive equity price response. Second, if investors correctly
anticipate the extent to which firms will change their financial reporting in response to S&P’s demands
for more conservative accounting, then the equity price response should be positively related to the
change in conservatism.
Because these tests require that we measure the change in conservatism at the firm level, we only
conduct this analysis using the level of working capital accruals and cscores which can be measured at
the firm-year level. A total of 85 firms have sufficient data to calculate both abnormal returns in the
three days surrounding the initiation of the credit rating and the level of working capital accruals both
before and after the rating event. The pre-event and post-event working capital accruals are measured as
the average working capital accruals using all available data from prior to the rating or after the rating,
respectively. 11 The change in accruals is defined as the post-period accruals minus the pre-period
accruals.
We estimate two simple regressions to determine if there is a relation between accounting conservatism
and the equity market reaction. First, we regress the abnormal returns in the three days surrounding the
rating event on the pre-rating working capital accruals. If investors anticipate that the firms with the
greatest benefit from the change in conservatism in the post-rating period are those with the least
conservative reporting prior to the rating, there should be a positive relation between the equity market
reaction and the level of working capital accruals (recall that higher accruals are less conservative).
11
Our results are qualitatively the same if we use only observations within 5 years before or after the credit rating
event.
26
This regression yields a positive coefficient with a robust t-statistic of 3.60, significant at the .001 level.
If investors correctly anticipate which firms will experience the greatest change in accruals, then the
abnormal returns should be negatively related to the change in accruals (the more negative the change in
accruals the more conservative the change in accounting). This regression of abnormal returns on the
change in conservatism yields a negative coefficient and a t-statistic of -2.42, significant at the .018
level. Both regressions are consistent with the notion that the change in conservatism prompted by the
S&P credit rating reduces information asymmetry and enhances equity value. The cscores discussed in
section 4.3 are also available at the firm-year level, so we repeated this analysis to test for a relation
between changes in cscores and abnormal returns. There is no relation between changes in cscores and
abnormal returns. In the next section, we perform several tests to determine whether there are changes in
analyst following which would also be expected if the S&P credit rating increases conservatism and
reduces information asymmetry.
6. Tests of Credit Rating Initiation and Analyst Following
The tests we have conducted so far show that firms respond to S&P demands for more conservative
financial reporting, and that this change in financial reporting also increases equity prices by reducing
information asymmetry. In this section we examine whether firms receiving an initial credit rating by
S&P also experience an increase in analyst following. If, as we have argued above, the initiation of
credit ratings by S&P prompts firms to improve their financial reporting practices toward more
conservative reporting, then equity analysts following should also increase in response to this improved
financial reporting. An alternative theoretically supportable prediction would be that the improved
information provided by the firm after S&P credit rating initiation decreases the demand for private
information and reduces analyst following (i.e. public and private information are substitutes). While
we acknowledge that this prediction can be justified in theory, the existing empirical evidence suggests
27
that analyst following is positively related to the quality of information provided by firms (Lang and
Lundholm, 1993; Lang, Lins and Miller, 2003 and 2004).
We obtain data on analyst following from I/B/E/S. We include all observations from 1991 through
2004 because I/B/E/S data prior to 1991 are sparse in emerging markets and most of the initial S&P
ratings occur after 1991. We include all firms, both rated and non-rated, with available IBES data from
our 18 sample countries, resulting in a sample of 33,480 firm-year observations. For the sample of S&P
rated firms, we have 966 firm-year observations of analyst following as of the 11th month of the firm’s
fiscal year. For the non-rated firms, there are 32,514 firm-year observations of analyst following.
We match our emerging market sample firms from I/B/E/S with the Worldscope database to extract
information regarding firm characteristics such as firm size and growth opportunities. We lose a large
number of observations in our tests that use these control variables because the overlap in I/B/E/S and
Worldscope coverage is not great, ending up with only 518 firm-year observations for the rated sample
and 10,608 for the non-rated sample. Tests that introduce additional controls suffer an even greater
decline in sample size.
Table 7 provides descriptive statistics for selected variables of the S&P rated firms versus the non-rated
firms from our 18 emerging market countries where all firms also have available I/B/E/S data. Rated
firms have greater analyst coverage, averaging over 10 analysts per firm-year observation compared to
only 5 analysts per firm-year observation for non-rated firms. The median number of analyst following
exhibits a similar pattern with 9 and 3 analysts for rated and non-rated firms, respectively. The average
total assets in U.S. dollars for rated and non-rated samples are $9,461 and $1,364 million, respectively.
The median total assets for the two groups of firms are $3,064 for rated firms and $296 million for nonrated firms. The average price-to-book ratios are 1.88 and 2.35 for the rated and non-rated sample,
respectively. This is surprising since one would expect rated firms to be growth firms. It turns out that
28
the median price to book ratio of the rated sample is larger at 1.47 than that of non-rated sample at 1.36,
suggesting that there are a few unusually large price to book ratios in the non-rated sample. As we saw
in the larger samples used in previous sections, rated firms are far more likely to be cross-listed, 35% of
rated firm-years have an ADR compared to around 6% of non-rated firm-years.
While not reported, we compute the mean number of analysts following the firm by year and also by
country. For both rated and non-rated samples, the number of observations and the average number of
analysts peaks in the late 1990s. The largest number of observations for S&P rated firms comes from
Mexico and Korea, providing 136 and 101 firm-year observations for rated sample firms with analyst
following, respectively. Thailand provides the fewest number of observations for rated firms, 13. For
both rated and non-rated firms, analyst following is lowest in Chile and highest in Mexico. In all
countries, analyst following is greater for rated firms than for non-rated firms.
We predict that credit rating initiation affects the information provided by rated firms so that firms are
followed by more analysts subsequent to the rating initiation. We examine the number of analysts
following the firm as one measure of the firm’s information environment. As in the cross-listing setting,
we anticipate that the improved scrutiny and monitoring provided by S&P will enhance the quality of
the information available about rated firms, providing analysts with greater incentives to follow these
companies. Following the past literature, we measure analyst following as the number of analysts
(NANAL) forecasting annual earnings in the eleventh month of the fiscal year, since analyst activity
levels off after this month (O’Brien and Bhushan, 1990; Lang, Lins and Miller, 2003).
Any increase in analyst following could be a result of changing firm characteristics rather than the initial
credit rating itself. Following our research design from our earlier tests, we control for cross-listing,
Asian firms, an annual time trend, and in additional tests include controls for firm characteristics shown
to be associated with analyst following in the past literature.
29
Our initial equation explaining the number of analyst following the firm is:
NANALi ,t   0  1 ( POSTRATE ) i ,t   2 ( ADR) i   3 ( POSTADR ) i ,t
  4 ( ASIA) i   5 ( POSTASIA) i ,t   6 (YR ) t   i ,t
(7)
where all variables are as previously defined. Following Rock et al. (2000), we use negative binomial
estimation for the analyst following model.
The results of estimating this initial equation are presented in column 1 of Table 8. The coefficient on
POSTRATE is positive and is statistically significant at the 1% level in this specification.
The
coefficient estimate on this variable is 0.260, which translates into the marginal effect of 2.58. This
increase means that when our sample firms get rated, there is on average an increase of 2.58 analysts
following the firm. Since there are 9 analysts following the sample firms on average before the rating,
the credit rating initiation by S&P leads to an increase of almost 30% in analyst following. The
coefficient estimates on ADR and POSTADR are positive but are not significant. The coefficient
estimate on POSTASIA is significantly positive, suggesting that there is more analyst following after
the Asian currency crisis. Finally, the coefficient on year, the annual time trend, is negative and
significant, indicating a downward trend in the number of analyst following emerging market firms
during our sample period.
This regression model in column 1 of Table 8 does not control for other firm characteristics that could
be changing around the credit rating initiation, which could impact analysts’ willingness to follow the
firms. In columns 2 and 3 we add several control variables, at the cost of significant reductions in
sample size since not all control variables are available for our rated firms. In column 2 we report the
results of the regression with two additional controls added, firm size (SIZE), measured as market
30
capitalization in U.S. dollars at the beginning of the fiscal year, and the price to book ratio (PB), also
measured at the beginning of the fiscal year. Consistent with our accruals based tests reported earlier,
we include the percentile rank of both variables in the regressions. We include size because the past
literature suggests that analysts follow larger firms, and price to book since analysts may be more
inclined to follow growth oriented firms. The coefficients on both variables are positive but only the
coefficient on SIZE is significant. The inclusion of these controls has no impact on other variables.
Column 3 reports the results after including three additional variables shown in the past literature to
affect analyst following: return standard deviation, the correlation between earnings and returns, and
earnings surprise. Definitions of these variables are provided in Table 8. While the sample size
significantly drops to 368, the coefficient on POSTRATE remains significant and positive. These results
are consistent with our previous findings suggesting that firms provide more conservative information,
reducing information asymmetry and enhancing analyst following.
We also once again conduct tests that examine the relation between the change in conservatism and the
change in analyst following to further corroborate our interpretation that S&P credit rating initiation
leads firms to provide more transparent and conservative financial reporting, and that this change in
reporting reduces information asymmetry and improves equity market performance and provides greater
incentives for analysts to follow the firm. Our tests follow the same general design that we used in
testing for a relation between changes in conservatism and the equity market reactions. We measure the
change in analyst following as the mean of the post-rating period number of analysts per year minus the
mean of the pre-rating period number of analysts per year. We again look at two measures of accruals to
measure reporting conservatism – the average level of accruals in the pre-rating period, and the
difference between the mean level of accruals in the post-rating period and the pre-rating period. Our
predictions mirror those made in the analysis of abnormal returns – if increased conservatism is
reducing information asymmetry and providing analysts with better incentives to follow the company,
31
then the increase in analyst following should be positively related to the level of accruals in the prerating period, and negatively related to the change in accruals that occurs after the rating initiation.
There are only 33 firms with available data to measure both the change in conservatism and analyst
following, so the results of this analysis must be interpreted with caution. However, the regression
results are generally consistent with our earlier findings. The regression of the change in analyst
following on the pre-period level of accruals yields a positive coefficient with a t-statistic of 2.45,
consistent with the idea that firms with a high level of pre-rating accruals (i.e. with the least
conservative accounting) have the most to gain by providing more conservative data and therefore
attract the most new analysts. Similarly, the coefficient on the change in accruals has a negative
coefficient and t-statistic of -3.91, indicating a strong relation between negative changes in accruals
(more conservative reporting) and increases in analyst following.12
7. Conclusions and Suggestions for Future Research
We investigate two primary research questions. First, we examine whether firms change their financial
reporting practices due to the informational demands associated with S&P credit rating initiation. We
find a significant increase in firms’ conditional conservatism in the period after the initial rating by
S&P, consistent with rated firms responding to demands by S&P for conservative reporting that
facilitates the assessment of downside risk. Conservative reporting helps creditors assess downside risk
because it speeds the recognition of potential losses and facilitates a conservative estimate of firm value.
The finding of increased conservatism is consistent across three measures of accounting conservatism.
12
Due to the small sample size for this analysis, the results using only 5 years of pre and post-rating data to
measure the change in accruals or pre-period accruals are not significantly different from zero. We again repeated
this analysis using the change in cscores and failed to find a relation between changes in cscores and changes in
analyst following.
32
Second, we investigate whether the changes in financial reporting that accompany S&P credit rating
initiation provide benefits to equity market participants as well. We find a positive equity market
reaction to the credit rating initiation and a negative market reaction for a smaller sample when S&P
discontinues providing credit ratings. We find that the equity market reaction is related to both
conservatism levels before credit rating initiation and to changes in conservatism when we use working
capital accruals as our measure of conservatism.
Finally, as additional evidence that the changes in financial reporting are beneficial to equity market
participants, we also find that credit rating initiation results in an increase in analyst following. This
finding corroborates the other two primary findings in the paper. If S&P credit rating initiation prompts
firms to improve their financial reporting by providing more conservative financial reporting, then the
past literature suggests that this improvement in the information provided by firms will provide greater
incentives for analysts to follow the firm. We again find an association between changes in analyst
following and conservatism measured using working capital accruals. These results provide
confirmation that our earlier results are attributable to the change in the information provided by firms
after their initial credit rating by S&P.
Our study adds to our understanding of the forces that shape financial reporting decisions internationally
by showing that credit rating initiation by S&P, similar to cross-listing equity shares, affects firms’
financial reporting decisions. This finding also furthers our understanding of accounting conservatism
by demonstrating an association between credit market informational demands and changes in
accounting conservatism in a powerful, time-series setting. We also add to our understanding of
governance mechanisms internationally and governance for emerging market firms specifically. Our
study also points out the need for more study of firm-level accounting conservatism in international
settings. Our study has examined one international transaction, obtaining credit ratings from S&P, for
33
emerging market firms. The accounting literature would also benefit from additional studies that
examine firm-level conservatism for international firms that may be exposed to more than one liability
regime, interact with international debt markets without obtaining S&P credit ratings, or otherwise
engage in transactions internationally that may affect accounting conservatism at the firm-level.
34
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37
Table 1 - Sample rating initiations in emerging markets by country and year
The sample consists of firms from 18 emerging market countries that have been rated by S&P. The table provides the frequency of initial rating for
each of our sample countries in each year. For each country, the median initial rating level is provided in the last column.
Country
Argentina
Brazil
Chile
Greece
India
Indonesia
Israel
Korea
Malaysia
Mexico
Philippines
Poland
Portugal
Russia
South
Africa
Taiwan
Thailand
Turkey
Total
Pre1994
1994
1
3
1
1
1
3
1
2
1995
4
1
1997
1998
1999
1
3
2
2
3
2
3
2
1
1
2
7
1
2
2
2
1
1996
1
1
3
2
5
2
4
3
2
7
3
1
12
1
13
3
10
28
1
4
2
1
2
1
2
3
2001
3
2
1
2
2
1
1
1
5
1
2000
2
4
2
36
1
1
1
1
4
13
38
2003
2004
2
1
1
2
1
2
2
2
1
5
1
2
1
2
1
1
1
1
4
1
2
1
2
1
1
19
2002
2
13
1
1
4
8
1
1
17
20
14
5
1
14
Total
10
17
14
6
7
14
9
23
9
19
13
6
5
9
9
21
11
7
209
Median
Initial
Rating
BBBBBBB+
BBBBB
BB
BBB
BBBBBB
BB
BB
BBBAB+
BBB
BBBBBB
B+
Table 2 - Descriptive statistics – Accruals sample
The table presents summary statistics of the variables used in the analyses. The sample consists firm-year
observations from 18 emerging market countries over the sample period of 1991-2004. TACC is total
accruals, WCACC is working capital accruals, CFO is cash flow from operations, P/B is the ratio of
stock price to book value of equity per share and Size is the market capitalization in millions of U.S. $.
P/B and Size are measured as of the beginning of the year. ADR is a dummy variable set equal to one for
any company that has been cross-listed in the U.S. stock exchanges during the sample period and zero
otherwise.
Variable
TACC
WCACC
CFO
Market capitalization
Price to book ratio
ADR
Rated firms
Obs.
Mean
994
994
994
964
964
994
-0.045
0.009
0.141
3,320.48
2.05
0.60
39
Median
-0.043
0.005
0.124
1,653.47
1.59
1
Non-rated firms
Obs.
Mean
Median
34,113
34,113
34,113
31,479
31,478
34,113
-0.027
0.017
0.097
476.20
1.94
0.10
-0.033
0.008
0.093
68.50
1.06
0.00
Table 3 - Credit rating initiation and accounting conservatism – level of accruals
The table presents the results from the regression model where the dependant variable is either total
accruals (Tacc) scaled by lagged total assets or working capital accruals scaled by lagged total assets
(Wcacc). POSTRATE is set equal to one for all observations occurring after the date on which the rated
firm receives its initial credit rating from S&P and zero otherwise. Year is set equal to zero in 1991, the
initial sample year, and increases by one for each calendar year represented in the sample. ADR is set
equal to one for any company with an ADR during the sample period and zero otherwise. POSTADR is
set equal to one for all observations occurring after the effective date of the ADR for firms with ADRs
and zero otherwise. Firm size is the percentile rank of total assets in U.S. dollars at the beginning of the
fiscal year. Price to book ratio is the percentile rank of the ratio of stock price to book value of equity per
share at the beginning of the fiscal year. The sample consists of up to 994 firm-year observations from 18
emerging market countries covered by Worldscope during the sample period of 1991-2004. Numbers in
parenthesis are t-statistics.
Variables
POSTRATE
tacc
(1)
-0.017
(-2.60)
Dependent Variable
wcacc
tacc
(2)
(3)
-0.011
-0.017
(-2.46)
(-2.54)
wcacc
(4)
-0.011
(-2.56)
ADR
-0.018
(-2.11)
0.000
(0.04)
-0.013
(-1.49)
0.002
(0.42)
POSTADR
-0.003
(-0.44)
0.000
(0.03)
-0.007
(-1.00)
-0.003
(-0.61)
ASIA
-0.005
(-0.58)
0.006
(0.94)
-0.007
(-0.73)
-0.002
(-0.42)
POSTASIA
-0.002
(-0.25)
-0.003
(-0.39)
-0.001
(-0.12)
0.002
(0.24)
YR
-0.001
(-0.60)
0.000
(0.47)
-0.001
(-0.63)
0.001
(1.13)
SIZE
-0.000
(-0.59)
-0.000
(-0.83)
PB
-0.000
(-0.25)
0.000
(3.76)
Constant
-0.016
(-2.30)
0.010
(1.98)
-0.011
(-1.11)
-0.002
(-0.21)
994
994
964
964
No. of observations
40
Table 4 - Credit rating initiation and accounting conservatism – timely loss recognition
The table presents the results from the regression model where the dependant variable is earnings per
share scaled by price (EPS) and the explanatory variables are abnormal returns net of local market returns
cumulated over the fifteen months ending three months after fiscal year end (RET) and a dummy variable
set equal to one (DRET) when RET is negative. These explanatory variables are also interacted with
several other variables included in our earlier tests. POSTRATE is set equal to one for all observations
occurring after the date on which the rated firm receives its initial credit rating from S&P and zero
otherwise. Year is set equal to zero in 1991, the initial sample year, and increases by one for each
calendar year represented in the sample. ADR is set equal to one for any company with an ADR during
the sample period and zero otherwise. POSTADR is set equal to one for all observations occurring after
the effective date of the ADR for firms with ADRs and zero otherwise. Firm size is the percentile rank of
total assets in U.S. dollars at the beginning of the fiscal year. Price to book ratio is the percentile rank of
the ratio of stock price to book value of equity per share at the beginning of the fiscal year. The sample
consists of firm-year observations from 18 emerging market countries covered by Datastream and
Worldscope during the sample period of 1991-2004.
Variables
(1)
EPS
(2)
t-value
(3)
EPS
RET
-0.034
-0.35
-0.247
-1.01
DRET
-0.093
-0.96
-0.237
-0.65
RET*DRET
-0.530
-0.80
1.472
2.59
POSTRATE
0.017
0.44
0.024
0.61
POSTRATE*RET
-0.097
-1.29
-0.101
-1.33
POSTRATE*DRET
-0.008
-0.17
-0.013
-0.29
POSTRATE*RET*DRET
0.222
1.87
0.307
2.42
ADR
0.011
0.33
0.008
0.23
ADR*RET
0.048
0.63
0.018
0.22
ADR*DRET
0.032
0.66
0.040
0.78
ADR*RET*DRET
-0.014
-0.08
0.449
1.41
POSTADR
-0.081
-2.04
-0.069
-1.90
POSTADR*RET
0.112
1.31
0.074
0.91
POSTADR*DRET
0.080
1.08
0.063
1.08
-0.089
-0.28
-0.115
-0.42
ASIA
0.016
0.49
0.001
0.02
ASIA*RET
0.011
0.20
-0.012
-0.19
ASIA*DRET
-0.002
-0.04
0.076
1.21
ASIA*RET*DRET
-0.104
-0.52
0.407
1.26
POSTASIA
-0.077
-1.35
-0.044
-0.79
POSTASIA*RET
0.028
0.27
0.015
0.15
POSTASIA*DRET
0.033
0.46
-0.031
-0.34
POSTADR*RET*DRET
41
(4)
t-value
-0.321
-0.95
-0.453
-1.36
YR
0.008
0.98
0.006
0.83
YR*RET
0.002
0.11
0.008
0.60
YR*DRET
0.006
0.39
0.008
0.59
YR*RET*DRET
0.109
1.17
0.069
1.06
-0.001
-0.37
SIZE*RET
0.003
1.07
SIZE*DRET
0.003
0.75
-0.020
-2.13
PB
0.001
1.67
PB*RET
0.000
-0.21
PB*DRET
-0.002
-2.08
PB*RET*DRET
-0.009
-2.19
-0.002
-0.01
POSTASIA*RET*DRET
SIZE
SIZE*RET*DRET
Constant
0.021
No. of observations
1,300
42
0.39
1,285
Table 5 - Credit rating initiation and accounting conservatism – Khan and Watts cscores
The table presents the results from the regression model where the dependant variable is the conservatism
score (cscore) as described in Khan and Watts (2007). POSTRATE is set equal to one for all
observations occurring after the date on which the rated firm receives its initial credit rating from S&P
and zero otherwise. Year is set equal to zero in 1991, the initial sample year, and increases by one for
each calendar year represented in the sample. ADR is set equal to one for any company with an ADR
during the sample period and zero otherwise. POSTADR is set equal to one for all observations occurring
after the effective date of the ADR for firms with ADRs and zero otherwise. Firm size is total assets in
U.S. dollars at the beginning of the fiscal year. Price to book ratio is the ratio of stock price to book value
of equity per share at the end of the fiscal year. LEV is the ratio of total debt to total assets. The sample
consists of firm-year observations from 18 emerging market countries covered by Worldscope during the
sample period of 1991-2004.
Panel A: Mean Coefficients from Initial Estimation Regressions
Variables
Coefficient estimates
Constant
t-statistics
0.046
14.01
-0.011
-2.89
RET
0.094
6.65
RET*SIZE
0.013
4.77
RET*P/B
-0.007
-3.15
RET*LEV
-0.085
-2.87
DRET*RET
-0.161
-4.89
DRET*RET*SIZE
-0.065
-4.14
DRET*RET*P/B
0.006
1.51
DRET*RET*LEV
0.396
5.23
DRET
Panel B: Descriptive Statistics of C_Scores
Mean
Standard
Deviation
C_Score
0.046
0.151
43
Q1
Median
Q3
-0.038
0.040
0.143
Table 5 - Credit rating initiation and accounting conservatism – Khan and Watts cscores
(continued)
Panel C: Cross-Sectional Regression of C_Scores
Numbers in parentheses are t-statistics
Variables
Cscore
POSTRATE
0.057
(2.38)
ADR
-0.074
(-2.29)
POSTADR
-0.035
(-1.15)
ASIA
-0.022
(-1.13)
POSTASIA
0.003
(0.13)
YR
-0.003
(-0.79)
Constant
5.345
(0.79)
No. of observations
1,057
44
Table 6 - Mean and median cumulative abnormal returns (CAR) for rating initiation
announcement dates and de-Rating announcement dates
The rating initiation sample comprises 173 firms where we have stock return data available during 19912004. The de-rating sample comprises 56 firms. The initial public announcement dates of the rating
initiation events and de-rating events are obtained from S&P. We compute abnormal returns by
estimating the market model using 200 trading days of return data ending 20 days before the rating
initiation announcement. AD denotes the initial announcement date. Numbers in parenthesis are p-values
for the test that the mean/median is equal to zero. Numbers in bracket are p-values for one-sided t-test that
the mean is equal to zero.
Event windows
Mean
Panel A: Credit rating initiation
(AD-20, AD-1)
Median
2.270
(0.01)
1.086
(0.10)
(AD, AD+1)
0.738
(0.04)
0.322
(0.03)
(AD-1, AD+1)
0.955
(0.03)
0.379
(0.02)
(AD-2, AD+2)
1.431
(0.01)
0.267
(0.05)
(AD+1, AD+20)
2.158
(0.02)
1.304
(0.03)
(AD-20, AD+20)
4.712
(0.00)
1.164
(0.00)
(AD-20, AD+50)
4.728
(0.00)
1.647
(0.07)
-1.503
(0.53)
-0.095
(0.78)
(AD, AD+1)
-1.311
(0.10)
-0.207
(0.37)
(AD-1, AD+1)
-2.603
(0.05)
-0.427
(0.16)
(AD-2, AD+2)
-2.856
(0.16)
0.018
(0.83)
(AD+1, AD+20)
-3.249
(0.14)
-0.140
(0.47)
Panel B: Credit de-rating
(AD-20, AD-1)
45
(AD-20, AD+20)
-5.974
(0.19)
-0.562
(0.33)
(AD-20, AD+50)
-8.189
(0.20)
1.116
(0.45)
46
Table 7 - Descriptive statistics – number of analysts
The table presents summary statistics of the variables used in the analyses of analyst following. The
sample consists of up to 33,480 firm-year observations from 18 emerging market countries covered by the
International I/B/E/S database during the sample period of 1991-2004. Number of analysts includes those
analysts who provide annual earnings forecast during the 11th month of the company’s fiscal year. P/B is
the ratio of stock price to book value of equity per share. Total assets and price to book ratio are obtained
from Worldscope. ADR is a dummy variable set equal to one for any company that has been cross-listed
in the U.S. stock exchanges during the sample period and zero otherwise.
Variable
Number of analysts
Total assets
Market capitalization
Price to book ratio
ADR
Rated firms
Obs.
Mean
966
564
564
518
966
10.32
9,461.10
2,954.52
1.88
0.35
47
Median
9.00
3,063.52
1,241.51
1.47
0.00
Nonrated firms
Obs.
Mean
32,514
12,405
12,412
10,608
32,514
4.91
1,364.64
525.70
2.35
0.06
Median
3.00
295.62
158.52
1.36
0.00
Table 8 - Credit rating initiation and number of analyst following
The table presents the results from the negative binomial regression model where the dependent variable
is the number of analysts providing an annual earnings forecast during the 11 th month of the company’s
fiscal year. RATE is set equal to one for all observations from firms receiving a credit rating from S&P
and zero otherwise. POSTRATE is set equal to one for all observations occurring after the date on which
the rated firm receives its initial credit rating from S&P and zero otherwise. Year is set equal to zero in
1991, the initial sample year, and increases by one for each calendar year represented in the sample.
ADR is set equal to one for any company with an ADR during the sample period and zero otherwise.
POSTADR is set equal to one for all observations occurring after the effective date of the ADR for firms
with ADRs and zero otherwise. Firm size is the logarithm of total assets in U.S. dollars at the beginning
of the fiscal year. Price to book ratio is the ratio of stock price to book value of equity per share by the
beginning of the fiscal year. Return STD is the standard deviation of monthly returns over the fiscal three
years before the year in which the dependent variable is measured. Return-earnings correlation is
spearman correlation between annual stock return and annual earnings over the three years before the year
in which the number of analyst is measured. Earnings surprise is the absolute value of the difference
between current earnings per share and earnings per share form the prior year, divided by the firm’s stock
price. The sample consists of firm-year observations from 18 emerging market countries covered by the
International I/B/E/S database during the sample period of 1991-2004. Numbers in parenthesis are tstatistics.
Variables
POSTRATE
(1)
0.260
(3.21)
(2)
0.206
(2.20)
(3)
0.363
(2.87)
-0.041
(-2.40)
-0.080
(-4.44)
-0.096
(-3.90)
ADR
0.159
(1.03)
0.085
(0.49)
0.041
(0.17)
POSTADR
0.179
(1.31)
0.204
(1.44)
0.286
(1.38)
ASIA
0.158
(1.08)
0.318
(2.00)
0.414
(2.17)
POSTASIA
0.181
(1.71)
0.237
(1.83)
0.201
(1.38)
SIZE
0.014
(4.22)
0.014
(3.81)
PB
0.001
(0.71)
0.001
(0.65)
YR
Return STD
-0.244
(-0.41)
Return-earnings correlation
0.005
(0.12)
48
Earnings surprise
Constant
No. of observations
-0.006
(-0.77)
2.254
(16.52)
1.320
(3.73)
1.356
(3.61)
966
518
368
49
S&P Ratings in Emerging Markets
800
All Emerging Markets
700
Sample Countries
Number of Rated Firms
600
500
400
300
200
100
0
1993
1994
1995
1996
1997
1998
1999
2000
2001
2002
2003
2004
Year
Figure 1. S&P Ratings in Emerging Markets. The figure shows the dramatic growth in ratings coverage by S&P for firms in emerging markets.
The top line indicates the number of corporate ratings provided by Standard & Poor’s in 53 emerging market countries while the bottom line
shows the number of ratings for the 18 sample countries chosen for this study. In both cases, there has been a dramatic increase in the number of
ratings provided since the mid-1990s.
50
Initial Rating Assignments
35
30
Number of Firms
25
20
15
10
5
AA
A
AA
+
AA
AA
-
A+
A
A-
BB
B+
BB
B
BB
B-
BB
+
BB
BB
-
B+
B
B-
+
C
C
C
C
C
C
C
C
-
+
C
C
C
C
C
0
Figure 2. Initial Rating Assignments. The figure shows the number of firms that receive initial rating assignments at each possible rating level
from the highest rating of AAA to the lowest initial rating in our sample, CC. For the majority of firms within our 18 sample countries the initial
rating is below the investment grade threshold of BBB.
51
Cumulative Abnormal Return Around Rating Initiation Announcement
6
5
In percent
4
3
2
1
0
-20
-15
-10
-5
0
5
10
15
20
25
30
35
40
45
50
Day relative the announcement
Figure 3. Cumulative Abnormal Return Around Rating Initiation Announcements. The figure shows the cumulative abnormal returns of rating
initiation announcements for the event window of day-20 to day+50.
52