Financial The Importance of Reporting Incentives:

Institutions Earnings Management in European

Center Private and Public Firms
by
David Burgstahler
Luzi Hail
Christian Leuz

04-07
 The Wharton Financial Institutions Center

The Wharton Financial Institutions Center provides a multi-disciplinary research approach to

the problems and opportunities facing the financial services industry in its search for
competitive excellence. The Center's research focuses on the issues related to managing risk
at the firm level as well as ways to improve productivity and performance.

The Center fosters the development of a community of faculty, visiting scholars and Ph.D.
candidates whose research interests complement and support the mission of the Center. The
Center works closely with industry executives and practitioners to ensure that its research is
informed by the operating realities and competitive demands facing industry participants as
they pursue competitive excellence.

Copies of the working papers summarized here are available from the Center. If you would
like to learn more about the Center or become a member of our research community, please
let us know of your interest.

Franklin Allen Richard J. Herring

Co-Director Co-Director

The Working Paper Series is made possible by a generous

grant from the Alfred P. Sloan Foundation
 The Importance of Reporting Incentives:
Earnings Management in European Private and Public Firms*

David Burgstahler
Gerhard G. Mueller Endowed Professor in Accounting
University of Washington/Seattle

Luzi Hail
Institute for Accounting and Control
University of Zurich

Christian Leuz
The Wharton School
University of Pennsylvania

February 2004

Abstract
This paper examines how capital market pressures and institutional structures shape firms’
incentives to report earnings that properly reflect their economic performance. To isolate the
effects of reporting incentives, we exploit the fact that, within the European Union, privately held
limited companies face the same accounting standards as publicly traded corporations because
accounting regulation is based on legal form. We hypothesize that raising capital in public markets
rather than from private sources and the institutional environment in which a firm operates have a
systematic influence on firms’ accounting quality. We focus on the level of earnings management
as one dimension of accounting quality that is particularly responsive to firms’ reporting
incentives. As hypothesized, our results document that raising capital in public markets and the
quality of the legal system are associated with the level of earnings management across European
countries. We find that earnings management is more pervasive in private firms and that both
public and private firms exhibit more earnings management in countries with weak legal
enforcement. We also document that private and public firms respond differentially to differences
in the tax and accounting rules in the EU.

JEL classification: G14, G15, G30, G32, K22, M41

Key Words: International accounting, Earnings management, Private companies, Legal
system, Accounting harmonization, Earnings properties

*
We thank Bob Bowen, John Core, Wayne Guay, DJ Nanda, Shiva Rajgopal, Terry Shevlin, Laurence van
Lent, Cathy Schrand, Peter Wysocki, and workshop participants at the University of Southern California, the
University of Washington and the Wharton School for helpful comments on earlier drafts. Luzi Hail
gratefully acknowledges the financial support by the research fund of the University of Zurich Association.
1. Introduction

In this paper, we examine the role of incentives stemming from capital market

pressures and legal institutions to report earnings that accurately reflect a firm’s economic

performance. Reporting incentives have been given little attention in the international

accounting debate. Much of the discussion has focused on accounting standards per se,

which are viewed as the primary input for high quality accounting (e.g., Levitt, 1998).

Consistent with this view, several countries have adopted or plan to adopt International

Financial Reporting Standards (IFRS) in an attempt to improve accounting quality.

Similarly, harmonization efforts within the European Union (EU) have largely focused on

eliminating differences in accounting standards across countries (e.g., Van Hulle, 2004).

Accounting standards generally grant substantial flexibility to firms. Measurements

are often based on private information and the application of standards involves judgment.

Corporate insiders can use the resulting discretion in reporting to convey information about

the firm’s economic performance, but they may also abuse discretion when it is in their

interest. For this reason, reporting incentives are likely to play an integral role in

determining the informativeness of reported accounting numbers. While this general

insight is not new (e.g., Watts and Zimmerman, 1986), it is often overlooked in

international standard setting. As Ball (2001) notes, the global debate focuses too much on

the standards and too little on the role of institutional factors and market forces in shaping

firms’ incentives to report informative earnings.

To empirically illustrate the importance of reporting incentives, we examine a setting

in which incentives to report about economic performance differ substantially across sets

of firms and countries, although standards are formally harmonized and largely held

1
constant. We hypothesize that raising capital in public markets rather than from private

sources and the institutional environment in which a firm operates have a systematic

influence on its incentives to report earnings that reflect economic performance. Both

factors shape the way in which information asymmetries between firms and the key

financing parties are resolved, i.e., the role earnings play in the process, which in turn

affects the properties of reported earnings (see also Ball et al., 2000).

International settings are especially powerful along the incentive dimension because

they offer much variation in institutional features and market forces. However, it is

difficult to isolate the effects of reporting incentives on earnings quality when accounting

standards vary across countries. The European setting provides a unique opportunity to

overcome these difficulties. As accounting regulation in the EU is not based on having

publicly traded securities but depends on a firm’s legal form, private limited companies

face the same accounting standards as publicly traded corporations. This feature allows us

to study reporting incentives and demand for information created by public debt and equity

markets, while holding accounting standards constant. At the same time, the European

setting provides variation in institutional factors across countries, which allows us to

examine their role in shaping earnings quality for both private and public firms.

Ideally, our analysis would be based on measures that directly capture the extent to

which firms use discretion to make earnings more informative about economic

performance. However, the use of discretion and the resulting informativeness of earnings

are difficult to measure because true economic performance is unobservable. Moreover,

we do not have stock prices for private firms, which generally serve as a benchmark for

economic performance. We therefore focus on the pervasiveness of earnings management

2
as an inverse proxy for the extent to which reported earnings reflect a firm’s true economic

performance. We believe that this proxy offers several advantages. First, earnings

management is an important dimension of accounting quality and in the extreme unlikely

to be informative.1 Second, earnings management proxies should be particularly

responsive to the use of discretion and firms’ reporting incentives, making our tests more

powerful. Third, there is an extensive literature offering various earnings management

proxies (see Healy and Wahlen, 1999).

As it is difficult to specify ex ante how firms manage earnings, we use four different

proxies based on Leuz et al. (2003) to measure the pervasiveness of earnings management.

These proxies are designed to capture a variety of earnings management practices, such as

earnings smoothing and accrual manipulations, and are constructed taking differences in

firms’ economic processes into account. We compute these proxies separately for public

and private firms within an industry and within a country to further control for industry and

country differences in business processes and economic activities.

We examine the pervasiveness of earnings management across private and public

firms from 13 European countries. Our results show substantial variation in the level of

earnings management across European countries, despite extensive harmonization efforts.

We find that earnings management is more pervasive in private firms than in publicly

traded firms. Thus, the demand for publicly traded capital and associated capital market

pressures appear to curb the level of earnings management and provide incentives to render

earnings more informative. We also find that earnings management is more pervasive in

1
Accounting quality is a broad concept with multiple dimensions. Empirical operationalizations of this
concept have focused on a variety of dimensions including timeliness and conservatism (e.g., Ball and
Shivakumar, 2002) or quality of accruals (e.g., Dechow and Dichev, 2002). In this paper we focus on
firms’ relative tendency to manage earnings as another dimension of accounting quality.
3
countries with German and French legal origins and in countries with weaker legal

systems. This finding highlights the importance of enforcement mechanisms and

documents that institutional differences influence both private and public firms.

We further demonstrate that institutional arrangements can differentially affect private

and public firms. Rules that closely align tax and financial accounting appear to have a

larger impact on the reporting behavior of private firms, consistent with the idea that

communicating firm performance via earnings is less important to private firms, which

allows earnings to assume other roles, such as minimizing tax payments. Conversely,

differences in accounting rules across the European countries that remain despite extensive

harmonization efforts appear to matter more for public firms. The latter finding suggests

that accounting standards designed to make earnings more informative play an incremental

role in reducing earnings management but only if coupled with incentives to report about

economic performance, e.g., capital market pressures. These interaction effects

corroborate our main proposition about the importance of reporting incentives. In either

case, the two main incentive variables remain significant and explain a substantial portion

of the variation in earnings management across European private and public firms.

Aside from presenting novel evidence on private firms, this paper builds on recent

studies highlighting the role of institutional factors and market forces in determining the

properties of earnings and accounting quality. In attempting to hold standards constant and

to isolate the influence of reporting incentives, our paper complements the studies by Ball

et al. (2003) and Ball and Shivakumar (2002). Ball et al. (2003) analyze earnings

timeliness and conservatism for four East Asian countries that have accounting standards

similar to common law countries, but differ in institutional structures. They show that

4
despite the similarity in accounting standards the earnings properties of the East Asian

countries do not resemble those in common law countries, like the U.K. or the U.S. Ball

and Shivakumar (2002) compare the timeliness of loss incorporation for private and public

firms in the U.K. and find evidence consistent with the notion that privately held firms

exhibit lower earnings quality than publicly traded firms.

We extend the Ball and Shivakumar findings to a large set of countries, which allows

us to combine and jointly analyze institutional differences and listing status. Furthermore,

we examine a group of countries for which accounting standards have been formally

harmonized by accounting regulation, which extends the work of Ball et al. (2003). We

also focus on a different dimension of earnings quality than these prior studies, namely

earnings management. Our results are consistent with this earlier work but offer several

new insights about the joint effects of market forces and institutional factors. Together,

these studies provide evidence that firms’ reporting incentives created by market pressures

and institutional structures are important determinants of accounting quality.

This insight has important implications for standard setters, suggesting that effective

accounting harmonization is unlikely to be achieved by accounting standards alone. In this

sense, our study relates to the literature on accounting harmonization (e.g., Gernon and

Wallace, 1995, or Saudagaran and Meek, 1997, for an overview) and accounting

convergence (Land and Lang, 2002; Joos and Wysocki, 2002). Our findings show that

considerable differences in the properties of accounting numbers persist across EU

countries despite decades of harmonization efforts. These findings are consistent with

other studies suggesting that the success of the EU harmonization process has been modest

(Emenyonu and Gray, 1992; Joos and Lang, 1994; Herrmann and Thomas, 1995).

5
 Our study also contributes to the earnings management literature. We demonstrate

that the influence of institutional factors on the level of earnings management extends to

private firms, generalizing the results in Leuz et al. (2003). Prior research has focused

primarily on publicly listed firms, despite the macroeconomic significance of private

firms.2 However, it is not clear that the arguments and prior findings extend to private

firms. Moreover, the evidence on differences in earnings management between public and

private firms is either confined to a particular industry or a particular country (e.g., Beatty

and Harris, 1999; Beatty et al., 2002; Vander Bauwhede et al., 2003). Our study spans a

broad cross-section of industries and countries and casts doubt on the extent to which

earlier findings generalize outside the highly regulated U.S. banking sector.

The paper is organized as follows. Section 2 develops our hypotheses. Section 3

describes the research design. In Section 4, we report the main results linking firms’

listing status and legal environments with the degree of earnings management. In Section

5, we consider additional institutional factors, namely, differences in tax alignment and

residual differences in the accrual accounting rules across Europe. Section 6 concludes.

2. Hypothesis Development

Our analysis is based on the recognition that accounting standards provide

considerable discretion to firms in preparing their financial statements. As corporate

insiders generally have private information about the firm’s economic performance, they

can use the discretion to report earnings that accurately reflect the firm’s underlying

performance. Alternatively, they may decide it is not worth their effort and resources to

make reported earnings more informative for external valuation and contracting purposes.

2
A notable exception is Coppens and Peek (2003) providing evidence on tax influences on private firms’
6
Moreover, corporate insiders can use the flexibility to hide poor economic performance,

achieve certain earnings targets or avoid covenant violations. Given their private

information, it is difficult to constrain such behavior. Thus, reporting incentives are likely

to play a crucial role in determining accounting quality and the informativeness of reported

earnings.3

In particular, we hypothesize that both raising capital in public markets rather than

from private sources and the institutional environment in which a firm operates have a

systematic influence on its incentives to report earnings that reflect economic performance.

Both factors shape the way in which information asymmetries between the firm and its key

financing parties are resolved and the role earnings play in the process, which in turn

affects the properties of reported earnings (see also Ball et al., 2000).

Incentives from Raising Capital in Public Markets

Privately held firms and those with publicly traded debt or equity securities face very

different demands for accounting information. Raising capital in public markets creates

strong incentives to provide information that is useful in evaluating and monitoring the

firm. Investors in public markets generally do not have private access to corporate

information and therefore rely heavily on information that firms provide. Financial

statements and, in particular, earnings play an important role in evaluating and monitoring

traded claims against the firm (e.g., Ball, 2001). Outside investors are aware of potential

conflicts of interests and the fact that corporate insiders prepare the financial statements.

Hence, they will be reluctant to supply capital to firms with low quality financial

tendency to avoid losses. However, they do not explicitly compare private firms to public firms.
3
This logic has also been exploited in the earnings management and accounting choice literature. See
Watts and Zimmerman (1986), Healy and Wahlen (1999) and Dechow and Skinner (2000).
7
statements. That is, in price protecting and demanding a higher cost of capital, investors

provide incentives to report financial information that reflects economic performance.

In contrast, privately held firms have relatively concentrated ownership structures and

hence can efficiently communicate among shareholders information via private channels.

Therefore, financial information and reported earnings are less important in

communicating firm performance, which in turn makes private firms less likely to expend

resources (e.g., hiring a high quality auditor) to produce earnings that are highly

informative about economic performance.4 Moreover, reported earnings can assume a

different role than for public firms. For instance, private firms face less of a tradeoff if

they manage earnings to minimize taxes but make them less informative in the process.

Alternatively, earnings can be used in determining dividends and other payouts to

stakeholders in the firm. Following Ball and Shivakumar (2002), we argue that these other

uses are likely to render earnings less informative for private firms.

While it is reasonable to believe that raising capital in public markets creates strong

incentives to provide earnings that reflect economic performance, we recognize that there

are many tradeoffs and potentially countervailing effects. For instance, Leuz et al. (2003)

argue that private control benefits and expropriation from outside investors create hiding

incentives for corporate insiders. That is, public firms with agency problems between

controlling insiders and outside investors may mask firm performance by managing

reported earnings to prevent outsider intervention. But similar incentives can arise for

private firms, which rely heavily on debt financing from banks and use financial

statements to inform banks about their performance. The relatively heavy use of bank debt

8
together with extensive creditor rights in many European countries make the threat of

creditor intervention even more real for private firms. As before, the fear of interference

and the subsequent loss of private control benefits can create incentives to obfuscate true

performance (see also Fudenberg and Tirole, 1995).5 Thus, it is not clear which agency

problem creates stronger incentives to misrepresent economic performance.

Another argument why public capital markets can create incentives to reduce the

informativeness of earnings in specific situations is based on earnings targets. Beatty et al.

(2002) argue that small investors in stock markets are more likely to rely on simple

heuristics such as earnings targets than fairly sophisticated private investors (e.g., banks),

which makes public firms more likely to engage in earnings management to exceed targets.

Similarly, public firms may manage earnings to meet or beat capital market expectations as

expressed in analyst forecasts. Stock based compensation contracts can further exacerbate

these incentives (e.g., Guttman et al., 2003). Consistent with these arguments, Beatty and

Harris (1999) and Beatty et al. (2002) present evidence from the banking industry

suggesting that public banks engage in more earnings management than their private

counterparts.

In summary, it is ultimately an empirical question whether private or public firms have

stronger incentives to make earnings more informative about firm performance.

4
Private firms may even have incentives to obfuscate firm performance because the EU requires them to
file financial statements to the corporate register. In particular, family-owned private firms may try to
hide true firm performance from employees and the general public.
5
While it is common in some European countries that bank representatives sit on the supervisory board
of public firms, making them effectively corporate insiders, banks rarely assume this role in private
firms, for the most part because supervisory boards are less common. Therefore, reported performance
is likely to be an important trigger for lender intervention.
9
Incentives Stemming from Institutional Structures

Countries differ in the way they channel capital to firms and in the way they reduce

information asymmetries between firms and the key financing parties. These differences

are likely to shape firms’ incentives to report earnings that reflect true economic

performance. We illustrate this idea using a stylized characterization of financial systems.

In an outsider system, like the U.K., firms rely heavily on public debt or equity

markets in raising capital. Corporate ownership is dispersed. Investors are “at arm’s

length” and do not have privileged access to information. Public debt and equity markets

play a major role in monitoring corporate insiders. Consequently, public information

about the firm is crucial as it enables investors to monitor their financial claims.

In contrast, in an insider, relationship-based system, firms establish close relationships

with banks and other financial intermediaries and rely heavily on internal financing,

instead of raising capital in public equity or debt markets. Corporate ownership is

generally concentrated (e.g., La Porta et al., 1999). Corporate governance is mainly in the

hands of insiders with privileged access to information (e.g., board members). In such a

system, information asymmetries are resolved to a large extent via private channels rather

than public disclosure.6 Ball et al. (2000) argue that these features reduce the demand for

high quality earnings, relative to outsider systems.

Prior studies suggest that countries’ legal origins are summary measures of these

institutional differences (e.g., La Porta et al., 1998; Ball et al., 2000). Based on the above

arguments, we expect earnings to be less informative in countries with French and German

6
Moreover, opacity is an important feature of the system because it provides barriers to entry and
protects relationships from the threat of competition (e.g., Rajan and Zingales, 1998). Opacity
effectively grants the financing parties some monopoly power over the firm, which allows insiders to
secure sufficient returns and in turn ensures relationship financing to firms.
10
legal origins compared to English and Scandinavian legal origins as the former are

typically viewed as insider systems (see also Leuz et al., 2003).

To explore specific institutional effects, we examine the influence of the legal

enforcement system. Without proper enforcement, legal institutions remain largely

ineffective. Moreover, in countries with weak legal systems, e.g., where rules are weakly

enforced, courts are inefficient and corruption is widespread, firms are more likely to abuse

accounting discretion. Thus, we hypothesize that ceteris paribus the quality of the legal

system and the informativeness of earnings are negatively associated.

Accounting Harmonization in the European Union

The European setting provides a unique opportunity to analyze the role of both types

of reporting incentives. First, there is a substantial range of institutional differences across

Europe. For example, the U.K. is generally viewed as an outsider economy, while

Germany and Italy are typically referred to as insider economies. The Netherlands and the

Scandinavian countries would generally be viewed as being somewhere in the middle.

Second, accounting regulation within the European Union is not based on listing

status. Privately held companies with limited liability face the same accounting standards

as publicly traded corporations. Thus, within a given country, accounting standards are

held constant across the two sets of firms. Moreover, accounting standards have been

formally harmonized across EU countries for many years (van Hulle, 2004).

The cornerstones of EU accounting harmonization are the Fourth and the Seventh

Directive. The Fourth Directive applies to some five million limited liability companies in

the EU. It requires these firms to prepare audited financial accounts according to the

principles laid out in the directive and to provide publicly accessible financial statements to

11
the corporate registers. The Seventh Directive complements these harmonization efforts

with requirements on when and how firms must prepare consolidated financial accounts.

Both directives effectively prescribe a common set of accounting rules for both

consolidated and unconsolidated (or parent-only) financial statements. The directives had

to be transformed into national laws by the member states during the late 1980s and early

1990s. As a result, accounting standards across EU member states are fairly similar,

though not necessarily equal in every respect. Explicit transformation choices in the

directives as well as so-called “soft transformations” lead to remaining differences (e.g.,

Stolowy and Jeny-Cazavan, 2001).7

Thus, the European setting is unique insofar as it provides substantial within-country

and cross-country variation in capital market and legal incentives while holding the

accounting rules largely constant. Our study exploits this variation and explicitly links it to

reporting incentives stemming from raising capital in public markets and institutional

structures.

3. Research Design and Data

3.1 Proxies for Earnings Management

Our hypotheses call for measures that directly capture the extent to which firms use

reporting discretion to make earnings more informative about the underlying economic

performance. However, both firms’ use of discretion and the resulting informativeness of

earnings are difficult to measure. A firm’s true economic performance is unobservable,

and we do not have stock prices for private firms, which could serve as a benchmark. We

therefore focus on the level of earnings management. Conceptually, earnings management

7
We address this issue in Section 5 by checking to what extent residual differences in the accounting
12
is the extent to which firms’ use reporting discretion to reduce the informativeness of

earnings and, hence, an inverse proxy for our theoretical construct.8 Moreover, earnings

management proxies should be particularly responsive to the use of discretion and firms’

reporting incentives, making our tests more powerful. Finally, we can draw on prior

research in constructing several measures of earnings management (e.g., Healy and

Wahlen, 1999; Dechow and Skinner, 2000).

Following Leuz et al. (2003), we compute four different proxies capturing a wide

range of earnings management activities: i.e., the tendency of firms to avoid small losses,

the magnitude of total accruals, the smoothness of earnings relative to cash flows and the

correlation of accounting accruals and operating cash flows. We recognize that these

proxies are not perfect and indicate earnings management only in a relative sense. But in

their defense, extreme realizations of the measures are unlikely to reflect informative

earnings, especially considering that we compute the proxies for a large set of firms over

several years. Moreover, they are constructed relative to outcomes of firms’ economic

processes, such as the magnitude or smoothness of the operating cash flows, which makes

it more likely that they capture firms’ reporting choices to make earnings more or less

informative. Finally, recent studies using these proxies suggest that they yield country

rankings that are consistent with widespread perceptions of earnings informativeness and

that exhibit plausible associations and behavior (e.g., Lang et al., 2003; Wysocki, 2004).

In what follows, we briefly discuss each of the four individual measures and how they

are implemented in our setting. Throughout the paper, the unit of analysis is a set of

rules affect our results.

8
Note that we do not claim that firms always use discretion to reduce the informativeness of reported
performance. In fact, there is evidence for the U.S. that, on average, managers use their discretion in a

13
private (or public) firms in an industry within a country. In computing proxies at the

industry-country level, we attempt to further control for industry and country differences in

firms’ business processes and economic activities.

EM1: Avoidance of Small Losses

Burgstahler and Dichev (1997) and Degeorge et al. (1999) present evidence that U.S.

firms use accounting discretion to avoid reporting small losses. While firms may have

incentives to avoid losses of any magnitude, they have limited reporting discretion and it

becomes increasingly costly to eliminate larger and larger losses. Thus, the incidence of

small profits relative to small losses indicates the extent to which firms use accounting

discretion to avoid reporting losses. A firm-year observation is classified as small profit

(small loss) if positive (negative) after-tax bottom-line net income falls within the range of

one percent of lagged total assets. We calculate the ratio of small profits to small losses at

the industry-country level, for public versus private firms.

EM2: Magnitude of Total Accruals relative to Cash Flow from Operations

More generally, firms can use their reporting discretion to mask or misstate economic

performance. For instance, firms can overstate reported earnings to achieve certain

earnings targets or report extraordinary performance in specific instances, such as an

equity issuance (e.g., Teoh et al., 1998a). Similarly, in years of poor performance, firms

can boost their earnings using reserves and allowances or aggressive revenue recognition

practices. Common to these examples is that earnings are temporarily inflated due to

accrual choices but cash flows are unaffected. Thus, we analyze the magnitude of accruals

way that makes earnings more informative (e.g., Watts and Zimmerman, 1986; Subramanyam, 1996).
We care about the relative informativeness of earnings and its association with reporting incentives.
14
relative to the magnitude of operating cash flow as a proxy for the extent to which firms

exercise discretion in reporting earnings.9 The ratio is computed as the median absolute

value of total accruals for an industry within a country scaled by the corresponding median

absolute value of cash flow from operations, where the scaling controls for differences in

firm size and performance.

Cash flow from operations is calculated using the balance-sheet approach because

U.S. style cash flow statements are generally not available for our sample of private and

public European companies. Following Dechow et al. (1995), we compute the accrual

component of earnings as (∆ total current assets – ∆ cash) – (∆ total current liabilities –

∆ short-term debt) – depreciation expense, where ∆ denotes the change over the last fiscal

year. If a firm does not report information on cash or short-term debt, then the changes in

both variables are assumed to be zero. We scale all accounting items by lagged total assets

to ensure comparability across firms.

EM3: Smoothing of Operating Earnings vis-à-vis Cash Flow

Controlling owners and managers can also conceal changes in their firm’s economic

performance by smoothing reported earnings. Our next measure attempts to capture the

degree of smoothing, i.e., the extent to which corporate insiders reduce the variability of

reported earnings using accruals. To control for differences in the variability of firms’

economic performance, we benchmark the variability of operating earnings to the

variability of the cash flow from operations. Specifically, the measure is computed as the

ratio of the standard deviation of operating income divided by the standard deviation of

9
We recognize that exercising reporting discretion does not imply that earnings are uninformative.
However, prior studies show that extreme accrual observations are indicative of poor earnings quality
(e.g., Sloan, 1996).
15
cash flow from operations. Due to data restrictions, we calculate the standard deviations in

the cross-section for each industry-country unit of analysis. We multiply the resulting ratio

by –1 so that higher values correspond to more earnings smoothing.

EM4: Correlation between Accounting Accruals and Cash Flow from Operations

An alternative approach to measuring whether corporate insiders smooth reported

earnings is to examine accrual choices in response to shocks to the firm’s economic

performance. Firms can use accruals to hide bad current performance as well as to

underreport good current performance and create “hidden reserves” for the future. In

either case, accounting accruals buffer cash flow shocks and result in a negative correlation

between changes in accruals and operating cash flows. While a negative correlation is a

“natural” result of accrual accounting (e.g., Dechow, 1994), larger magnitudes of this

correlation indicate, ceteris paribus, smoothing of reported earnings that does not reflect a

firm’s underlying economic performance (see Skinner and Myers, 1999).10 Consequently,

we define our fourth individual earnings management measure as the contemporaneous

Spearman correlation between the changes in total accruals and the changes in cash flow

from operations calculated for each industry-country unit of analysis. We scale all changes

by lagged total assets and multiply the resulting ratio by –1, so that higher values indicate

higher levels of earnings management.

Aggregate Measures of Earnings Management

Finally, to mitigate potential measurement error, we transform the individual earnings

management scores into percentage ranks (ranging from 0 to 100) and combine the average

16
ranks into indices. We define two sub-categories, “earnings discretion” (EM1 and EM2)

and “earnings smoothing” (EM3 and EM4), and construct an earnings management index

for each of the sub-categories (denoted EMdiscr and EMsmooth) as well as an aggregate index

of earnings management (denoted EMaggr).

3.2 Data, Sample Selection and Descriptive Statistics

The primary source of financial data is the January 2003 version of the Amadeus Top

200,000 database supplied by Bureau van Dijk. Amadeus provides standardized financial

statement data for a vast set of European private and public companies and is compiled

from several well-established national information collectors. Since its coverage is less

detailed in initial years, we focus on the five-year period from 1997 to 2001. Amadeus

provides consolidated financial statements when they are available and parent-only

accounts otherwise. Thus, our analysis is based on a firm’s primary set of financial

statements from an informational perspective.

The main advantage of the relatively new Amadeus database is that it includes

privately held corporations, allowing us to focus on an economically important group of

firms that is fairly under-represented in academic research. But it also has limitations.

First, the accounting information provided for private firms is not as detailed as in standard

databases for public firms. Second, as by definition stock price data are not available for

private firms, we cannot use market values as an independent benchmark in evaluating

accounting information. Finally, the distinction between publicly traded and privately held

corporations is based on the listing status indicator. Each release of Amadeus only

10
As accounting systems likely under-react to economic shocks, insiders signaling firm performance use
accruals in a way that on average results in a less negative (and in specific cases even positive)
correlation with cash flows.
17
contains the listing status of a company’s current fiscal year. No time-series data on listing

status are provided and all past information is classified under the latest firm type

available. Thus, for firms that have changed listing status over the five-year sample

period, some firm-years may be misclassified.11

Sample Construction

The initial sample consists of all firm-year observations from private and public

companies that have their domicile in one of the 15 member states of the European Union

(EU), where current year’s net income and previous year’s total assets are available on

Amadeus. By adopting size restrictions similar to those laid out in the Fourth EU

Directive, we explicitly exclude small privately held firms to which the EU directives may

not fully apply.12 Specifically, we require privately held firms to meet at least two of the

following three criteria in every year: (1) total assets greater than EUR 2.5 millions, (2)

sales greater than EUR 5 millions, and (3) number of employees greater than 50. We also

exclude banks, insurance companies and other financial holdings (SIC codes between 6000

and 6799), public administrative institutions (SIC codes above 9000), as well as privately

held subsidiaries of quoted companies as indicated in Amadeus. Investment, financing and

operating decisions in the latter firm category are likely to be influenced by parent

companies, which may bias our analyses. These sampling criteria result in 298,290 firm-

11
We would prefer to eliminate firms that are going public over the sample period as they have been
documented to exhibit higher levels of earnings management (e.g., Teoh et al., 1998b). But since data
restrictions do not allow us to identify these firms, we ignore changes in listing status in our analyses.
Their proportion in the overall sample, however, is likely to be very small.
12
The Fourth EU Directive distinguishes between small, medium-sized and large companies depending
on the three criteria balance sheet total, net turnover and average number of employees (Article 11 and
27). Small and medium-sized companies are subject to certain exemptions from reporting
requirements, e.g., they are allowed to draw up abridged balance sheets and income statements.
18
year observations from non-financial private and public companies located in 15 EU

countries.

We further eliminate observations from two sample countries with missing accounting

and legal institutional data. Amadeus does not provide data on operating income and

depreciation expenses for companies from Ireland, and several institutional proxies used in

the analysis are missing for Luxembourg.13

To mitigate the influence of outliers and potential data errors we truncate accounting

items needed in the calculation of our proxies at the first and 99th percentile and delete

firm-year observations where accounting items are exactly equal to zero, most likely

indicating missing data. For robustness, we check that our results do not hinge on either of

these two design choices. The final sample consists of 287,354 firm-year observations

from private and publicly traded, non-financial companies over the fiscal years 1997 to

2001 across 13 European countries.

Several of our earnings management proxies have to be computed for a group of firms.

In light of this requirement, we have to define a unit of analysis that is fine enough so that

group members share similar characteristics, but at the same time results in enough

observations per group to reliably capture the degree of earnings management. With only

9,693 firm-level observations the number of public companies is clearly the limiting factor

(compared to 277,661 private-firm observations). We conduct all our analyses on the

industry-level using the industry classification in Campbell (1996). That is, we calculate

the individual and aggregate earnings management scores by country and industry for both

public and private firms, resulting in possibly 312 observations (= 13 countries × 12

13
If we include Luxembourg in the analysis adopting the legal institutional data from Belgium, all the
results and inferences remain the same.
19
industry classes × 2 firm types).14 We require a minimum of ten firm-year observations

per unit of analysis. This requirement reduces the sample to a total number of 274

industry-level observations, of which 152 are from privately held firms and 122 are from

publicly listed firms. If we increase the required number of firm-year observations per

group to N ≥ 25 (N ≥ 50), we lose 35 (72) public firm observations, and 5 (9) private firm

observations, respectively.15

Descriptive Statistics for Dependent Variables and Firm-level Controls

Panel A of Table 1 presents descriptive statistics for the four individual earnings

management measures (EM1 through EM4) as well as the overall earnings management

index (EMaggr). On the country-level, mean values of industry observations from privately

held companies generally exceed their public counterparts. For EMaggr, only Greece

exhibits less earnings management among private companies, which may partly reflect the

fact that Greek publicly traded firms exhibit more earnings management than public

companies in any other sample country.16 On the other end of the spectrum, public firms

from the U.K. and Finland exhibit low levels of earnings management. For the sample as a

whole, mean and median values calculated from listed companies are significantly lower

than private company means and medians.

Panel B of Table 1 reports Spearman correlation coefficients between earnings

management scores. All four individual measures are highly correlated and well

represented by the aggregate index. Since the Amadeus database has not been used much

14
Observations with missing industry data in Amadeus are grouped together in a separate industry class.
If we delete this ad hoc group from our analyses, the results and the inferences remain unchanged.
15
The substantial reduction in public firm observations weakens the statistical power, resulting in lower
significance levels for the public firm indicator in some of the analyses but without changing the tenor
of the results.

20
in prior studies, we also benchmark our earnings management measures with those in Leuz

et al. (2003) based on public firms from the frequently used Worldscope database. In

(untabulated) analyses we find that the correlations between their measures and the public

firm observations from our sample are above .65 for all individual EM scores, except EM1,

and above .90 for the aggregate index. As reliably measuring loss avoidance is likely to

require a substantial number of firm-years, it could well be that the relatively low

correlation of EM1 is driven by the smaller number of public firms in the Amadeus

database. As a robustness check, we repeat all our analyses dropping EM1 from the

construction of the aggregate index. The results are very similar.

Table 2 presents descriptive statistics for firm characteristics used as control variables

in the multivariate tests. We choose proxies for which prior work suggests an association

with the level of earnings management (or accruals) and which are also likely to differ

across private and public firms. SIZE is the book value of total assets at the end of the

fiscal year (in EUR thousands) and is used to control for differences between public and

private companies in size. Since access to capital and other corporate financing decisions

depend on the extent of agency costs and asymmetric information (e.g., Titman and

Wessels, 1988; Rajan and Zingales, 1995) and these attributes likely differ between public

and private companies, we include financial leverage as a control variable. We calculate

financial leverage, LEV, as the ratio of total debt to the sum of total debt plus book value

of equity.17 Two additional sources of variation between privately held and public

companies are firm growth and profitability. GROWTH is defined as the annual

16
This finding is consistent with Leuz et al. (2003) where Greece (together with Austria) ranked highest
in terms of earnings management for an international sample of quoted companies.

21
percentage change in revenue, and ROA is the yearly return on assets as measured by

bottom-line net income divided by lagged total assets. All firm attributes are computed as

industry-level medians.

Table 2 shows that, as expected, public firms are on average larger, more profitable

and experience higher growth rates than private firms. These relations hold for every

single country in the sample except Portugal. As the differences in size between the two

firm categories are striking, we also perform sensitivity analyses using only the largest

quartile of private firms in order to make the two groups more comparable sizewise (results

not tabulated). This drastic reduction in available firm-years leaves the results virtually

unchanged and the inferences remain the same. With respect to financial leverage, no clear

pattern arises, which may be due to our inability to distinguish between various forms of

debt. Overall, mean and median values of LEV are not statistically different across private

and public firms. The remainder of Table 2 provides information by country on the legal

and institutional variables discussed in the next two sections.

4. Main Results on the Role of Reporting Incentives

Our main tests focus on reporting incentives that stem from raising capital in public

markets and the institutional environment in which a firm operates. Both factors shape the

way in which information asymmetries between firms and the key financing parties are

resolved and the role of earnings in communicating economic performance.

A binary variable indicates observations stemming from firms with publicly traded

debt or equity securities (PUBL). Institutional incentives are captured by a legal variable

17
This definition abstracts from the fact that for European countries liabilities oftentimes include
provisions arising from national labor market contracts or country-specific regulations with no relation
to financing decisions (Giannetti, 2003).
22
drawn from La Porta et al. (1998). LEGAL represents the average score across three

proxies measuring the quality of the legal enforcement system: (1) an index of the judicial

system’s efficiency, (2) an index of the rule of law, and (3) the level of corruption.

LEGAL ranges from zero to ten with higher values corresponding to a higher degree of

legal quality and enforcement.

In selected analyses, we also examine the level of earnings management by a

country’s legal origin (i.e., English, Scandinavian, French, or German), which is probably

the most exogenous classification of international legal institutions. However, legal

origins also capture ownership patterns and differences in financial development (e.g., La

Porta et al., 1999 and 2000). Hence, they are more appropriately viewed as institutional

summary measures. The main analysis therefore focuses on the LEGAL variable in an

attempt to more specifically identify institutional effects stemming from the quality of

legal enforcement.

Univariate Comparisons across the Capital Market and Legal Incentives Dimensions

Table 3 provides univariate tests across the two incentive dimensions under study.

Panel A summarizes the results of within-country univariate regressions with an indicator

variable to compare earnings management for public versus private firms. Summary

results are shown for the sub-category indices as well as for the aggregate index. For the

earnings discretion sub-category, the earnings management index is significantly greater

for private firms in 10 of the 13 countries and there is no significant difference between

public and private firms in the remaining 3 countries. For the earnings smoothing sub-

category, the earnings management index is significantly greater for private firms in 7

countries, and there is no significant difference between public and private firms in 5

23
countries. In the remaining country, Greece, the sub-index is significantly greater for

public firms. For the aggregate earnings management index, the index is significantly

greater for private firms in 8 of the 13 countries, there is no significant difference between

public and private firms in 4 countries, and in the remaining country, Greece (see

discussion in Section 3), the index is significantly greater for public firms.

Thus, for most, but not all, individual countries, the evidence suggests that earnings

management is more pervasive among private firms than among public firms. This result

also holds if we contrast each of the four individual earnings management scores (EM1

through EM4) by country for private versus public firms. The bottom row in Panel A of

Table 1 provides an assessment of statistical significance for each EM measure using

Monte Carlo simulation.

Panel B of Table 3 shows correlations among the earnings management indices, listing

status, and the LEGAL variable. The top three lines of panel B describe the correlations

among the earnings management indices. As expected, the indices are highly correlated

and significantly different from zero at the 1% level. The correlations of the earnings

management indices with the PUBL variable are negative, reflecting the general result

showing lower earnings management scores for public firms. The correlation is

substantially more negative for EMdiscr (–.49) than for EMsmooth (–.17). In contrast, the

strength of the negative correlation between the earnings management indices and the legal

enforcement variable is stronger for the smoothing sub-index than for the discretionary

sub-index, and strongest for the aggregate earnings index, presumably reflecting the

reduction of measurement error. Again, all correlations are significantly different from

zero at the 1% level.

24
 Taken as a whole, the evidence presented here and in Table 1 shows no obvious

advantage to using the individual components or either of the two sub-indices. Given this

evidence, and in light of the potential measurement advantage from combining all earnings

management scores into a single index, the remainder of the analysis in the paper focuses

on the aggregate earnings management index, EMaggr.18

Panel C of Table 3 compares the aggregate earnings management index across groups

of countries defined by their legal origins, which broadly capture investor protection and

ownership regimes across countries (La Porta et al., 1999). The ordering of the mean and

median values of EMaggr follows the expected ordering of legal origins, with the English

origin countries showing the lowest earnings management index, followed by slightly

higher values for the Scandinavian origin countries, followed by significantly higher

earnings management among French origin countries, and still higher values for the

German origin legal system. These findings are consistent with Ball et al. (2000) and Leuz

et al. (2003). Pairwise comparisons of all the differences described above are significant at

the .05 level or better, except for the difference between the English and Scandinavian

countries. Thus, even within the set of EU countries and despite accounting

harmonization, we observe significant variation in the level of earnings management

reflecting institutional differences.

Panel D of Table 3 reports average EMaggr for subgroups defined by the two main

variables, legal quality and listing status, to see how they separately and jointly affect the

informativeness of earnings. We create a binary variable indicating high and low legal

quality countries splitting LEGAL by the median value. As hypothesized, publicly traded

18
Unreported factor analysis shows that there is only one factor with an eigenvalue above one and that all
of the four individual scores exhibit substantial loadings onto this factor, further supporting the use of a

25
firms in countries with high legal quality show the lowest level of earnings management.

In contrast, private firms located in low quality legal environments exhibit the highest

levels of earnings management. As the off-diagonal comparisons show both variables

appear to play a role for the pervasiveness of earnings management. In fact, they do so in

an almost equal manner, as the private firms/high legal quality and the public firms/low

legal quality cells are statistically indistinguishable. These findings suggest that both

incentive dimensions separately contribute to the reporting behavior of European private

and public firms.

Multivariable Tests of Capital Market and Legal Incentives Hypotheses

Table 4 presents results of regressions that examine the role of capital market forces

and legal factors, separately and jointly, thereby including additional controls for

differences in firm characteristics. Throughout the discussion, assessments of statistical

significance are based on heteroskedasticity-corrected t-statistics.19

The first four columns of Table 4 examine the effect of listing status, introducing

controls for differences in firm characteristics that may be systematically related to

earnings management and differ across public and private firms. The first column shows

our benchmark case without any control variables, and confirms earlier results of

significantly higher earnings management scores for private firms. The second column

introduces size and leverage controls, which are potential explanators of earnings

management, although neither is empirically significant in this setting. The third column

introduces two more control variables, growth and return on assets (ROA), which

single aggregate index.

19
As the legal variable varies only at the country level, we repeat our analyses using robust standard
errors clustered by country, resulting in a more conservative assessment of statistical significance.

26
systematically affect the level of accruals. Finally, column four adds industry controls

based on the industry classification in Campbell (1996). Moving across the four

specifications, the magnitude of the coefficient on PUBL remains relatively constant

except in Model 3. Correspondingly, the coefficient on PUBL is significant at the .01 level

in all but Model 3, where it is significant at the .05 level. Thus, introduction of variables

reflecting controls for firm and industry characteristics do not alter the basic conclusion

that raising capital in public markets results in greater demand for high-quality earnings as

evidenced by the lower level of earnings management for public firms.

The next four columns examine the legal incentive variable separate from the public

versus private hypothesis. Column five shows that the earnings management index is

significantly negatively related to the enforcement variable (benchmark case). Columns

six to eight show that this significantly negative relation continues to hold if we introduce

size, leverage, growth, ROA, and industry controls.

The final column presents a combined model simultaneously including the capital

market and the legal variable together with the controls. The coefficients on the variables

of interest are highly significant. In fact, the t-statistic for the PUBL coefficient in the

complete model is larger than in any other model, except for the simple univariate

comparison in column one.20

Design choices to this point maintain sample size but allow differences in industry

composition between the public versus private observations and between countries. To the

However, the inferences are the same and the legal variable remains significant at the 1% level in all
models presented in Table 4.
20
In order to explicitly account for the varying numbers of input data, we also estimated the regressions in
Table 4 using weighted least squares. The square root of the number of firm-level observations serves
as weighing variable. With the exception of the coefficient on PUBL in Model 3, which has a p-value
of .18, all previously reported results remain significant at the .01 level.
27
extent that economic differences between industries are systematically related to the

informativeness of reported earnings, it is possible that the differing incidence of industry

observations among public versus private firms and among countries accounts for some of

the reported results. To address this issue, we construct a balanced sample that includes

only observations where we have corresponding observations from the same industry for

both public and private firms and for all countries included. With this design, the sample

is less than half the number of observations in the preceding analysis, and includes 5

industries from 11 countries, resulting in a total of 110 public-firm and private-firm

observations. The (untabulated) results are essentially unchanged from those reported in

Table 4. Overall, the results are consistent with the hypothesis that listing status and legal

enforcement have a significant influence on firms’ incentives to report earnings that reflect

economic performance, which is the paper’s main proposition.

5. Additional Analyses on the Role of Tax and Accounting Rules

In this section we explore differences in institutional factors in tax and accounting

regimes across countries and how they relate to our main results on capital market

incentives. Although accounting standards are formally harmonized within the EU, there

are still differences in the way tax and financial accounting are aligned as well as residual

differences in the accounting rules themselves. These differences are likely to affect

private and public firms differentially and hence are potentially confounding factors in our

analyses.

Influence of Tax Alignment

We first introduce a variable to control for the effects of different tax alignment

regimes throughout the EU. Ball et al. (2000) hypothesize that the link between financial

28
and tax accounting can play an important role in firms’ reporting behavior. Moreover,

recent work suggests that taxes could play a role in observed earnings distributions as well

as firms’ tendency to avoid losses (Coppens and Peek, 2003). We rely on a tax status

classification provided by Alford et al. (1993) and Hung (2001). It indicates countries with

a high alignment of financial and tax accounting, for instance, because financial statements

serve as basis for taxation purposes, or tax laws explicitly require that certain items are

treated equally in both sets of accounts (see Table 2). The TAX variable takes on a value

of one when financial and tax accounts are highly aligned and zero otherwise. For the

three countries with missing tax information (Austria, Greece and Portugal), we assume a

tax status of one.21

Table 5 Panel A examines the relation between the TAX variable and the earnings

management index. The results for Model 1 show that TAX is significant in a univariate

model (benchmark case). When the capital market and legal incentive variables are added

in Model 2, the TAX variable provides incremental explanatory power significant at the

.05 level, while the coefficients on PUBL and LEGAL remain significant at the .01 level

and are similar in magnitude to those in the combined model in Table 4.

It is quite plausible that the degree of tax alignment of financial accounting has a

differential effect on private and public firms. For instance, as explained in Section 2,

private firms face less of a tradeoff than public firms if they manage earnings to minimize

taxes but make them less informative in the process. Moreover, tax considerations matter

more to private firms, many of which prepare (unconsolidated) parent-company accounts,

21
The rationale for this assumption is that Austria is close to the German system, which is classified as a
one, and Portugal and Greece are both French legal origin countries, which generally exhibit a high
alignment. If we exclude the three countries without explicit classification from the tax analysis, the
results do not materially change.
29
for which the tax link is more explicit. To examine these arguments, we first estimate the

effect of TAX in separate models for public and private firms. Model 3 shows a positive

and significant coefficient on TAX for private firms and Model 4 a positive but

insignificant coefficient for public firms. These findings are consistent with our

expectations. However, an ad hoc test based on the assumption that the two coefficient

estimates are independent normal variates with variances equal to their estimated variances

indicates that the effects of TAX on earnings management are not significantly different

for public versus private firms.

In order to provide a more formal test of the difference between the TAX coefficients

for public versus private firms, we introduce a model to estimate the coefficients jointly.

The natural way to conduct joint estimation is to include a term that interacts the PUBL

indicator variable and the TAX variable. However, empirically this interaction term, the

TAX proxy and the LEGAL variable are highly correlated, which makes the estimate of

the interaction term difficult to interpret.

To mitigate this problem, we empirically decompose the TAX variable into a

component related to the general institutional and legal environment and a residual

component that reflects the separate effect of the TAX variable. Specifically, we estimate

a first-stage regression of the TAX variable on the legal origins and the natural log of GDP

per capita. This first stage regression is intended to separate the systematic component of

TAX related to the general nature of the institutional environment from the component of

TAX that is uniquely attributable to tax alignment. The legal origin captures this general

nature together with a country’s wealth, which likely plays an important role for the quality

of the legal institutions. The predicted value of TAX is included in the second-stage

30
regression (reported in the table as Model 5) to capture institutional incentives (similar to

the LEGAL variable before) while the residual component is included as main effect and

interacted with the PUBL variable. Thus, the coefficients on the residual TAX component

and the interaction of the residual TAX component and PUBL provide an assessment of

whether the effect of tax alignment differs across public and private firms.

The estimates for Model 5 show a positive coefficient on TAX, corresponding to the

unique effects of tax alignment for private firms, and a negative coefficient on the

interaction of TAX and PUBL, which corresponds to a lower tax effect for public firms.

However, both coefficients are not statistically significant at a two-tailed level. The

coefficient on PREDICT is significant at the .01 level, which is consistent with our earlier

findings that incentives from the institutional framework have substantial explanatory

power. Finally, the coefficient on PUBL remains negative and highly significant in the

presence of these additional explanatory variables.

To further corroborate and even strengthen the results in Table 5 Panel A, we rerun

separate analyses using only observations which come from consolidated financial

statement data or only observations which come from unconsolidated parent-company

accounts.22 For countries where there is a link between financial reporting and taxes, we

expect the effect of the tax variable to be much more important for the unconsolidated

observations than for the consolidated observations, because the link with taxes is

commonly with the unconsolidated parent-only accounts. Untabulated results

corresponding to those in Table 5 differ between the two samples as one would expect.

Consistent with our expectations, the tax variable for the unconsolidated sample is always

31
significant (including the coefficients of the main effect and the interaction term in Model

5) whereas the tax variable loses its significance in the consolidated sample.

Taken as a whole, accounting for differences in tax alignment does not alter the

primary conclusion that capital market pressure increases firms’ incentives to report

earnings that reflect economic performance. At the same time, the finding that the tax

alignment variable is more important for private firms than for public firms is consistent

with the conjecture that reported earnings can serve different purposes in private

companies.

Influence of Remaining Differences in Accounting Rules

A second potential confounding factor on reporting incentives are differences in the

accounting rules remaining after formal harmonization. Comprix et al. (2003), for

instance, argue and provide evidence that the remaining differences in EU accounting rules

give rise to stock market reactions to regulatory action towards further harmonization and

stricter enforcement. We use the “accrual rules index”, ACCRUAL, constructed by Hung

(2001) and updated for EU countries by Comprix et al. (2003), to capture the remaining

differences in the accounting rules. This index measures the use of accrual accounting

rules to accelerate recognition of economic transactions in the accounts (e.g., R&D

activities, pension plans). It proxies for the extent to which a country’s stated accounting

rules are designed to produce timely and informative reported earnings. Higher index

values correspond to more extensive accrual rules that allow for accelerated recognition

(see Table 2).

22
For completeness, we also reexamined the results in Table 4 for these two subsamples and, despite the
reduction in sample size for the subsamples, the significance levels for the major variables are
qualitatively unchanged from those reported in Table 4.
32
 Table 5 Panel B examines the relation between the accrual accounting rules and the

earnings management index. The results for Model 1, our benchmark case, show that

ACCRUAL is significant at the .01 level. When PUBL and LEGAL are added in Model 2,

the ACCRUAL variable continues to be significant, but so are the two incentives variables.

Thus, our previous results are not adversely affected by an attempt to explicitly control for

residual differences in EU accounting rules.

As with tax alignment, it is quite plausible that the effects of accrual accounting rules

differ across private and public firms. Rules that are designed to produce timely and

informative earnings are likely to be more important for public firms, which heavily rely

on financial statements in their communications with outside investors. We again estimate

separate models for public and private firms. Model 3 shows a negative and significant

coefficient on ACCRUAL for private firms and Model 4 shows a more negative and

significant coefficient for public firms. Using the ad hoc test described above, the

difference between the coefficients from the two separate regressions is significant at the

.10 level. These results suggest that the effects of accrual accounting rules on earnings

management are significantly different for public versus private firms and, as expected,

more important for the former group.

To further explore this issue, Model 5 provides an assessment of whether the

ACCRUAL variable provides incremental explanatory power relative to PUBL when its

effect is allowed to vary between public and private firms. The accrual rules variable is

included in Model 5 in two parts, derived from a first-stage regression of ACCRUAL on

the legal origins and the natural log of GDP per capita. The estimates for Model 5 show a

negative but insignificant coefficient on ACCRUAL. The coefficient on the interaction of

33
ACCRUAL and PUBL is negative and significant at the .05 level, which corresponds to a

negative effect of accrual rules for public firms. The coefficient on PREDICT is

significant at the .01 level. As in Panel A, the coefficient on PUBL remains negative and

highly significant in the presence of these additional explanatory variables.

The results in Panel B suggest that the design of the accounting rules has relatively

little effect on earnings management by private firms but a significant effect for public

firms.23 This finding suggests that accounting standards that are intended to produce

timely and informative earnings can play an incremental role but only if coupled with

incentives to report about economic performance. This interaction effect corroborates our

main proposition about the importance of reporting incentives.

6. Conclusion

This study extends and generalizes previous evidence that firms’ reporting incentives

created by market forces and institutional factors are important determinants of accounting

quality. In an international setting where accounting standards are largely harmonized

while institutional factors vary substantially, we examine whether raising capital in public

markets rather than from private sources and strong legal enforcement exert a systematic

influence on firms’ incentives to report earnings that accurately reflect economic

performance.

For a large sample of private and public firms from 13 European countries, we find

that one dimension of earnings quality, earnings management, is more pervasive in private

firms than in publicly traded firms, consistent with the hypothesis that capital market

incentives result in higher earnings quality. These results are interesting in light of recent

23
However, this result has to be interpreted cautiously, as Leuz et al. (2003) show that the influence of the
accounting rules is potentially endogenous across countries.
34
allegations that capital markets exacerbate incentives to mask economic performance. The

results also contribute to the earnings management literature, suggesting that prior

evidence in Beatty and Harris (1999) and Beatty et al. (2002) showing greater evidence of

earnings management among public firms than private firms for U.S. banks is not

generalizable to a broader set of industries and countries. The results also extend those in

Leuz et al. (2003) by demonstrating that the influence of institutional factors on the level

of earnings management extends to private firms. The results show that for both private

and public firms, earnings are less transparent in countries with weak legal systems.

Finally, we document a residual role of tax alignment and accounting rules, which

seem to differentially affect private and public firms. Rules that align tax and financial

accounting have a larger impact on the reporting behavior of private firms, consistent with

the idea that private firms emphasize tax minimization relative to communicating firm

performance via earnings. On the other hand, residual differences in the accounting rules

across the European countries are more important for public firms, suggesting that

accounting standards designed to make earnings more informative play a more important

role when they are coupled with capital market pressures to report earnings that reflect

economic performance.

Taken as a whole, the evidence suggests that, despite extensive efforts to harmonize

accounting standards, there remains substantial variation across European countries in the

informativeness of earnings and that reporting incentives created by capital markets and

institutional structures explain much of this variation.

35
References

ALFORD, A.; J. J. JONES; R. W. LEFTWICH; AND M. E. ZMIJEWSKI. “The Relative

Informativeness of Accounting Disclosures in Different Countries.” Journal of
Accounting Research 31 (1993): 183-223.

BALL, R. “Infrastructure Requirements for an Economically Efficient System of Public

Financial Reporting and Disclosure.” In Brookings-Wharton Papers on Financial
Services, edited by R. E. Litan and R. Herring, pp. 127-69. Washington DC:
Brookings Institution Press, 2001.
BALL, R., AND L. SHIVAKUMAR. “Earnings Quality in U.K. Private Firms.” Working paper,
University of Chicago and London Business School, May 2002.

BALL, R.; S. P. KOTHARI; AND A. ROBIN. “The Effect of International Institutional Factors
on Properties of Accounting Earnings.” Journal of Accounting & Economics 29
(2000): 1-51.

BALL, R.; A. ROBIN; AND J. S. WU. “Incentives Versus Standards: Properties of

Accounting Income in Four East Asian Countries.” Journal of Accounting &
Economics (2003): forthcoming.
BEATTY, A. L., AND D. G. HARRIS. “The Effects of Taxes, Agency Costs and Information
Asymmetry on Earnings Management: A Comparison of Public and Private Firms.”
Review of Accounting Studies 4 (1999): 299-326.
BEATTY, A. L.; B. KE; AND K. R. PETRONI. “Earnings Management to Avoid Earnings
Declines across Publicly and Privately Held Banks.” The Accounting Review 77
(2002): 547-70.
BURGSTAHLER, D., AND I. D. DICHEV. “Earnings Management to Avoid Earnings
Decreases and Losses.” Journal of Accounting & Economics 24 (1997): 99-126.
CAMPBELL, J. Y. “Understanding Risk and Return.” Journal of Political Economy 104
(1996): 298-345.
COMPRIX, J. J.; K. A. MULLER, III; AND M. STANFORD. “Economic Consequences from
Mandatory Adoption of IASB Standards in the European Union.” Working paper,
Arizona State University, Pennsylvania State University, and Texas Christian
University, July 2003.
COPPENS, L., AND E. PEEK. “An Analysis of Earnings Management by European Private
Firms.” Working paper, University of Maastricht, June 2003.
DECHOW, P. M. “Accounting Earnings and Cash Flows as Measures of Firm Performance:
The Role of Accounting Accruals.” Journal of Accounting & Economics 18 (1994):
3-42.

36
DECHOW, P. M., AND D. J. SKINNER. “Earnings Management: Reconciling the Views of
Accounting Academics, Practitioners, and Regulators.” Accounting Horizons 14
(2000): 235-50.

DECHOW, P. M., AND I. D. DICHEV. “The Quality of Accruals and Earnings: The Role of
Accrual Estimation Errors.” The Accounting Review 77 (2002): 35-59.
DECHOW, P. M.; R. G. SLOAN; and A. P. SWEENEY. “Detecting Earnings Management.”
The Accounting Review 70 (1995): 193-225.
DEGEORGE, F.; J. PATEL; AND R. ZECKHAUSER. “Earnings Management to Exceed
Thresholds.” Journal of Business 72 (1999): 1-33.
EMENYONU, E. N., AND S. J. GRAY. “EC Accounting Harmonisation: An Empirical Study
of Measurement Practices in France, Germany and the UK.” Accounting and
Business Research 23 (1992): 49-58.
FUDENBERG, D., AND J. TIROLE. “A Theory of Income and Dividend Smoothing Based on
Incumbency Rents.” Journal of Political Economy 103 (1995): 75-93.
GERNON, H., AND R. S. O. WALLACE. “International Accounting Research: A Review of
its Ecology, Contending Theories and Methodologies.” Journal of Accounting
Literature 14 (1995): 54-106.
GIANNETTI, M. “Do Better Institutions Mitigate Agency Problems? Evidence from
Corporate Finance Choices.” Journal of Financial and Quantitative Analysis 38
(2003): 185-212.
GUTTMAN, I.; O. KADAN; AND E. KANDEL. “Adding the Noise: A Theory of
Compensation-Driven Earnings Management.” Working paper, Hebrew University,
November 2003.
HEALY, P. M., AND J. M. WAHLEN. “A Review of the Earnings Management Literature and
its Implications for Standard Setting.” Accounting Horizons 13 (1999): 365-83.
HERRMANN, D., AND W. THOMAS. “Harmonisation of Accounting Measurement Practices
in the European Community.” Accounting and Business Research 25 (1995): 253-
65.
HUNG, M. “Accounting Standards and Value Relevance of Financial Statements: An
International Analysis.” Journal of Accounting & Economics 30 (2001): 401-20.
JOOS, P., AND M. H. LANG. “The Effects of Accounting Diversity: Evidence from the
European Union.” Journal of Accounting Research 32 (1994): 141-68.
JOOS, P., AND P. D. WYSOCKI. “(Non)Convergence in International Accrual Accounting:
The Role of Institutional and Real Operating Effects.” Working paper, Sloan
School of Management MIT, April 2002.

37
LA PORTA, R.; F. LOPEZ-DE-SILANES; AND A. SHLEIFER. “Corporate Ownership around the
World.” Journal of Finance 54 (1999): 471-517.
LA PORTA, R.; F. LOPEZ-DE-SILANES; A. SHLEIFER; AND R. W. VISHNY. “Law and
Finance.” Journal of Political Economy 106 (1998): 1113-55.
____. “Investor Protection and Corporate Governance.” Journal of Financial Economics
58 (2000): 3-27.
LAND, J., AND M. H. LANG. “Empirical Evidence on the Evolution of International
Earnings.” The Accounting Review 77 (2002): 115-33.

LANG, M. H.; J. SMITH RAEDY; AND M. HIGGINS YETMAN. “How Representative Are Firms
that Are Cross-Listed in the United States? An Analysis of Accounting Quality.”
Journal of Accounting Research 41 (2003): 363-86.
LEUZ, C.; D. NANDA; AND P. D. WYSOCKI. “Earnings Management and Investor
Protection: An International Comparison.” Journal of Financial Economics 69
(2003): 505-27.
LEVITT, A., JR. “The Importance of High Quality Accounting Standards.” Accounting
Horizons 12 (1998): 79-82.
RAJAN, R. G., AND L. ZINGALES. “What do we know about Capital Structure? Some
Evidence from International Data.” Journal of Finance 50 (1995): 1421-60.

____. “Which Capitalism? Lessons from the East Asian Crisis.” Journal of Applied
Corporate Finance 11 (1998): 40-48.

SAUDAGARAN, S. M., AND G. K. MEEK. “A Review of Research on the Relationship

between International Capital Markets and Financial Reporting by Multinational
Firms.” Journal of Accounting Literature 16 (1997): 127-59.

SKINNER, D. J., AND L. A. MYERS. “Earnings Momentum and Earnings Management.”

Working paper, University of Michigan, April 1999.

SLOAN, R. G. “Do Stock Prices Fully Reflect Information in Accruals and Cash Flows
about Future Earnings?” The Accounting Review 71 (1996): 289-315.
STOLOWY, H., AND A. JENY-CAZAVAN. “International Accounting Disharmony: The Case
of Intangibles.” Accounting, Auditing & Accountability 14 (2001): 477-96.
SUBRAMANYAM, K. R. “The Pricing of Discretionary Accruals.” Journal of Accounting &
Economics 22 (1996): 249-81.
TEOH, S. H.; I. WELCH; AND T. J. WONG. “Earnings Management and the
Underperformance of Seasoned Equity Offerings.” Journal of Financial Economics
50 (1998a): 63-99.

38
____. “Earnings Management and the Long-Run Market Performance of Initial Public
Offerings.” Journal of Finance 53 (1998b): 1935-74.
TITMAN, S., AND R. WESSELS. “The Determinants of Capital Structure Choice.” Journal of
Finance 43 (1988): 1-19.
VAN HULLE, K. “From Accounting Directives to International Accounting Standards.” In
The Economics and Politics of Accounting, edited by C. Leuz, D. Pfaff, and A. G.
Hopwood, pp. 351-76. Oxford: Oxford University Press, 2004.
VANDER BAUWHEDE, H.; M. WILLEKENS; AND A. GAEREMYNCK. “Audit Firm Size, Public
Ownership, and Firms’ Discretionary Accruals Management.” International
Journal of Accounting 38 (2003): 1-22.
WATTS, R. L., AND J. L. ZIMMERMAN. Positive Accounting Theory. Englewood Cliffs,
N.J.: Prentice-Hall, 1986.
WYSOCKI, P. “Earnings Management, Tax Compliance, and Institutional Factors: A
Discussion of Haw et al. [2004].” Journal of Accounting Research 42 (2004):
forthcoming.

39
 TABLE 1
Descriptive Statistics for the Individual and Aggregate Earnings Management Scores
Panel A: Mean Earnings Management Scores of Private and Public Firms by Country
Listing Firm- Industry Earnings Discretion Scores Earnings Smoothing Scores Total Score
Country Status Years Obs. EM1 EM2 EM3 EM4 EMaggr
Austria private 1,536 11 3.143 0.743 -0.585 0.916 66.6
public 88 3 0.667 0.656 -0.511 0.920 53.0
Belgium private 20,495 12 3.326 0.746 -0.543 0.907 69.2
public 206 9 1.056 0.642 -0.646 0.840 42.1
Denmark private 7,879 12 2.400 0.616 -0.658 0.834 44.4
public 259 9 0.611 0.562 -0.591 0.850 39.6
Finland private 6,734 11 2.991 0.526 -0.680 0.822 36.8
public 350 10 0.700 0.385 -0.654 0.741 22.6
France private 47,482 12 3.554 0.677 -0.622 0.885 57.9
public 1,282 12 2.667 0.519 -0.757 0.806 33.6
Germany private 4,338 12 4.296 0.784 -0.495 0.908 75.7
public 932 10 2.998 0.592 -0.600 0.893 49.8
Greece private 5,499 12 1.383 0.702 -0.625 0.885 47.7
public 1,122 12 1.257 0.654 -0.528 0.938 60.5
Italy private 47,383 12 4.091 0.729 -0.512 0.935 78.3
public 479 11 3.564 0.670 -0.626 0.877 55.9
Netherlands private 13,762 12 2.067 0.635 -0.607 0.885 46.9
public 628 11 2.015 0.506 -0.723 0.808 36.2
Portugal private 5,888 11 7.242 0.736 -0.506 0.938 80.6
public 141 7 2.571 0.686 -0.610 0.839 55.5
Spain private 42,934 12 4.566 0.664 -0.613 0.869 60.4
public 526 12 1.604 0.445 -0.611 0.873 40.5
Sweden private 15,038 11 2.581 0.582 -0.654 0.820 39.5
public 301 4 3.500 0.622 -0.852 0.642 33.3
United private 58,693 12 1.889 0.655 -0.593 0.844 47.8
Kingdom public 3,379 12 2.313 0.446 -0.743 0.750 24.1
Mean private 277,661 152 3.332 0.677 -0.592 0.881 57.9
(Total) public 9,693 122 1.979 ** 0.554 ** -0.651 ** 0.834 ** 41.5 **

Median private 152 2.864 0.684 -0.593 0.891 56.1

public 122 1.183 ** 0.541 ** -0.649 ** 0.867 # 42.9 **

Standard private 152 2.193 0.111 0.104 0.067 18.7

Deviation public 122 2.030 0.155 0.179 0.131 19.9
Within-Country (# higher/# not distinguishable/# lower) (4/9/0) (8/4/1) (9/2/2) (7/5/1)
(continued)
 TABLE 1 (continued)
Panel B: Spearman Correlation Coefficients among Earnings Management Scores
Earnings Discretion Scores Earnings Smoothing Scores Total Score
Variables (N = 274) EM1 EM2 EM3 EM4 EMaggr
Earnings Discretion Scores EM1 1.000
EM2 0.372 ** 1.000
Earnings Smoothing Scores EM3 0.244 ** 0.458 ** 1.000
EM4 0.199 ** 0.364 ** 0.604 ** 1.000
Total Earnings Management Score EMaggr 0.593 ** 0.744 ** 0.794 ** 0.747 ** 1.000
The initial sample consists of 287,354 firm-year observations from private and public, non-financial companies over the fiscal years 1997 to 2001 across 13 countries of the European
Union. We exclude Ireland and Luxembourg because of insufficient financial and institutional data. Financial data are obtained from the January 2003 version of the Amadeus Top 200,000
database. The computations are at the industry-level using the classification in Campbell (1996), i.e., we calculate each score by country, industry and listing status. We require a minimum
of ten firm-year observations per subgroup. We truncate firm-level realizations at the first and 99th percentile before computing the score per subgroup. The table presents mean values by
country and listing status and Spearman correlations of four individual earnings management scores and the aggregate earnings management index (see Leuz et al., 2003). EM1 is the
number of “small profits” divided by the number of “small losses”. A firm-year observation is classified as small profit (small loss) if positive (negative) net income falls within the range
of one percent of lagged total assets. EM2 is the median ratio of the absolute value of total accruals to the absolute value of cash flow from operations. Total accruals are calculated as
follows: (∆ total current assets – ∆ cash) – (∆ total current liabilities – ∆ short-term debt) – depreciation expense. Cash flow from operations is equal to operating income minus total
accruals. EM3 is the ratio of the cross-sectional standard deviations of operating income and cash flow from operations (multiplied by –1). EM4 is the Spearman correlation between the
change in total accruals and the change in cash flow from operations (multiplied by –1). All accounting items are scaled by lagged total assets. The aggregate earnings management index,
EMaggr, is the average percentage rank across all four individual scores, EM1 to EM4. EM scores are constructed such that higher values imply higher levels of earnings management. We
evaluate sample means (medians) using t-tests (Wilcoxon rank sum tests). In the within-country analysis (Panel A), we assess the statistical significance of the four individual EM scores
across the two subsamples of public and private firms using Monte Carlo simulation. The table reports the number of countries where the EM scores for private firms are (1) significantly
higher, (2) not distinguishable, and (3) significantly lower than the EM scores for public firms at the 5% level. **, *, and # indicate statistical significance at the 1%, 5%, and 10% levels
(two-tailed), respectively.
 TABLE 2
Descriptive Statistics for the Control and Institutional Variables by Country
Country Listing Status Industry Obs. SIZE LEV GROWTH ROA Legal Origin LEGAL TAX ACCRUAL
Austria private 11 62,595 0.0% 5.6% 3.1%
German 9.36 1 0.55
public 3 233,350 0.0% 13.5% 4.5%
Belgium private 12 18,833 40.7% 6.8% 2.1%
French 9.44 1 0.64
public 9 341,348 37.2% 12.1% 3.8%
Denmark private 12 21,887 17.4% 8.2% 4.4%
Scandinavian 10.00 0 0.55
public 9 183,292 23.9% 11.7% 4.9%
Finland private 11 12,866 26.3% 9.0% 5.3%
Scandinavian 10.00 1 0.77
public 10 427,521 28.8% 14.4% 7.8%
France private 12 16,420 15.9% 7.1% 2.9%
French 8.68 1 0.64
public 12 292,407 31.1% 11.6% 4.5%
Germany private 12 50,996 2.4% 3.9% 2.4%
German 9.05 1 0.41
public 10 307,400 12.2% 7.6% 3.5%
Greece private 12 10,771 41.4% 13.1% 3.0%
French 6.82 1 0.41
public 12 82,866 23.6% 16.7% 7.3%
Italy private 12 19,978 35.2% 9.1% 1.1%
French 7.07 1 0.59
public 11 291,237 19.2% 10.2% 3.2%
Netherlands private 12 49,622 31.2% 7.4% 4.0%
French 10.00 0 0.77
public 11 580,869 35.0% 14.8% 6.6%
Portugal private 11 23,906 44.7% 10.9% 2.4%
French 7.19 1 0.55
public 7 72,063 51.0% 6.5% 2.1%
Spain private 12 14,353 33.9% 12.3% 4.4%
French 7.14 1 0.77
public 12 1,450,759 35.3% 12.3% 5.6%
Sweden private 11 12,229 15.0% 8.6% 3.5%
Scandinavian 10.00 1 0.64
public 4 75,398 16.4% 16.1% 3.8%
United private 12 23,786 43.2% 11.8% 4.1%
English 9.22 0 0.86
Kingdom public 12 209,665 28.4% 14.7% 5.8%
Mean private 152 25,969 26.9% 8.8% 3.3%
public 122 390,152 ** 27.9% 12.5% ** 5.1% **

Median private 152 14,571 28.0% 8.1% 3.2%

public 122 149,609 ** 27.2% 11.1% ** 4.9% **

Standard private 152 42,675 17.9% 4.0% 1.6%

Deviation public 122 1,072,473 13.9% 7.4% 2.5%
The initial sample consists of 287,354 firm-year observations from private and public, non-financial companies over the fiscal years 1997 to 2001 across 13 countries of the European Union. We
exclude Ireland and Luxembourg because of insufficient financial and institutional data. Financial data are obtained from the January 2003 version of the Amadeus Top 200,000 database. The analysis is
based on country-industry medians by listing status, i.e., private and public firms in a particular industry and country form separate subgroups. We require a minimum of ten firm-year observations per
subgroup. We truncate firm-level realizations at the first and 99th percentile before computing the subgroup median. The industry classification is based on Campbell (1996). The table presents means
for the control variables by country and listing status. SIZE is the book value of total assets at the end of the fiscal year (in EUR thousands). Financial leverage (LEV) is measured as the ratio of total
debt to the sum of total debt plus book value of equity. GROWTH is the annual percentage change in revenue. ROA stands for yearly return on assets and equals bottom-line net income divided by
lagged total assets. The legal variables consist of two measures from La Porta et al. (1998): (1) the classification of the legal origin, and (2) the general quality of the legal system and enforcement
(LEGAL) measured by the mean of three institutional variables (i.e., efficiency of the judicial system, rule of law, and corruption index). TAX is an indicator variable taking on the value of one if
financial accounts for external reporting and tax purposes are highly aligned and zero otherwise (see Alford et al., 1993, and Hung, 2001). We assume a tax status of one for the three countries with
missing tax information (Austria, Greece and Portugal). ACCRUAL is the accrual index from Hung (2001) (updated for European countries by Comprix et al., 2003), and captures differences in accrual
accounting rules across countries. We evaluate sample means (medians) using t-tests (Wilcoxon rank sum tests). ** indicates statistical significance at the 1% level (two-tailed).
 TABLE 3
Univariate Analysis of Earnings Management Indices,
Listing Status and Legal Variables
Panel A: Within-Country Analysis of Public versus Private Firms
Variables EMdiscr EMsmooth EMaggr
Within-Country (# negative/
# insignificant/# positive) (10/3/0) (7/5/1) (8/4/1)

Panel B: Spearman Correlation Coefficients

Variables (N = 274) EMdiscr EMsmooth EMaggr
EMdiscr 1.000
EMsmooth 0.424 ** 1.000
EMaggr 0.812 ** 0.862 ** 1.000

PUBL -0.498 ** -0.175 ** -0.362 **

LEGAL -0.256 ** -0.351 ** -0.375 **

Panel C: Pairwise Comparison across Legal Origins

Legal Origins English Scandinavian French German
English Mean 35.9
(N = 24) Median 36.1 (Variable = EMaggr)
Scandinavian Mean (0.19) 36.6
(N = 57) Median (0.38) 39.2
French Mean (4.31) ** (6.18) ** 54.8
(N = 157) Median (4.23) ** (5.99) ** 56.4
German Mean (6.31) ** (8.28) ** (2.44) * 63.8
(N = 36) Median (4.94) ** (6.18) ** (2.23) * 62.1
Panel D: Pairwise Comparison across Legal Quality/Listing Status Subgroups
Legal Quality Listing Status Test of
(Variable = EMaggr) Public Private Difference
High Quality Mean 33.5 50.3 (5.54) **
Median 29.0 44.7 (4.67) **
(N = 58) (N = 81)
Low Quality Mean 48.7 66.6 (5.93) **

Median 52.2 68.1 (5.20) **

(N = 64) (N = 71)
Test of (4.56) ** (5.95) **

Difference (4.26) ** (5.57) **

The sample comprises 274 industry-level observations from 13 European countries. The aggregate earnings
management index, EMaggr, is the average percentage rank across all four individual earnings management scores,
EM1 to EM4, as described in Table 1. EM discr and EMsmooth are calculated similarly for the two sub-categories
“earnings discretion” (EM1 and EM2) and “earnings smoothing” (EM3 and EM4). EM scores are constructed such
that higher values imply higher levels of earnings management. In Panel A, we assess differences in the three
aggregate EM indices across public and private firms using regression analysis by country. The panel reports the
number of countries where the coefficient on the public/private indicator is (1) significantly negative, (2)
insignificant, and (3) significantly positive at the 5% level, indicating either higher, indistinguishable, or lower
levels of earnings management among private firms. In Panel B, we report Spearman correlation coefficients for
the aggregate EM indices and the incentive variables of interest. PUBL is an indicator variable taking on the value
of one if the observation stems from publicly traded firms and zero otherwise. LEGAL stands for the mean of
three variables in La Porta et al. (1998), which measure the quality of the legal system and enforcement (i.e.,
efficiency of the judicial system, rule of law, and corruption index). In Panels C and D, we report means and
medians of EMaggr by subgroup (i.e., legal origin and legal quality/listing status). We assign observations to the
high (low) legal quality subgroup according to the median value of LEGAL. In parentheses, we report t-stats and Z-
stats for pairwise differences across groups. ** and * indicate statistical significance at the 1% and 5% levels (two-
tailed), respectively.
 TABLE 4
Earnings Management and Reporting Incentives: The Role of Listing Status and Institutional Differences
EMaggr,i = α0 + α1PUBLi + α2LEGALi + α3SIZEi + α4LEVi
+ α5GROWTHi + α6ROAi + ∑αjIndustry Controlsi + εi
Variables Capital Market Incentives Legal Incentives Combined
(N = 274) Model 1 Model 2 Model 3 Model 4 Model 1 Model 2 Model 3 Model 4 Model
PUBL -16.424 ** -18.218 ** -8.390 * -14.945 ** -20.142 **
(-6.99) (-4.35) (-2.20) (-3.56) (-5.58)
LEGAL -6.104 ** -6.244 ** -5.316 ** -5.476 ** -6.300 **
(-6.69) (-6.31) (-5.67) (-6.07) (-6.97)

Intercept 57.880 ** 48.956 ** 74.009 ** 49.150 ** 103.700 ** 147.828 ** 142.493 ** 140.394 ** 91.701 **
(38.20) (3.66) (6.09) (3.47) (12.86) (12.09) (12.38) (11.91) (5.88)
SIZE 0.784 -0.237 2.430 # -3.751 ** -2.204 ** -1.631 * 3.995 **
(0.57) (-0.20) (1.72) (-5.31) (-3.33) (-2.46) (3.16)
LEV 4.709 -1.987 -0.739 -9.332 -12.426 # -10.993 -11.123
(0.65) (-0.31) (-0.12) (-1.06) (-1.69) (-1.45) (-1.60)
GROWTH 14.596 16.596 -17.841 -34.059 # -7.761
(0.63) (0.62) (-1.03) (-1.74) (-0.38)
ROA -4.434 ** -4.481 ** -4.081 ** -4.052 ** -3.462 **
(-7.51) (-7.31) (-7.65) (-7.49) (-6.65)
Industry
Controls included included included

R2 15.4% 15.6% 32.9% 42.8% 12.8% 20.2% 39.9% 48.4% 54.2%

The sample comprises 274 industry-level observations from 13 European countries. The dependent variable, EM aggr, is the average percentage rank across all four individual earnings
management scores, EM1 to EM4, as described in Table 1. EM scores are constructed such that higher values imply higher levels of earnings management. PUBL is an indicator
variable taking on the value of one if the observation stems from publicly traded firms and zero otherwise. LEGAL stands for the mean of three variables in La Porta et al. (1998),
which measure the quality of the legal system and enforcement (i.e., efficiency of the judicial system, rule of law, and corruption index). SIZE is the book value of total assets at the
end of the fiscal year (in EUR thousands). We use the natural log of the size variable in the analysis. Financial leverage (LEV) is measured as the ratio of total debt to the sum of
total debt plus book value of equity. GROWTH is the annual percentage change in revenue. ROA stands for the yearly percentage return on assets. Industry controls based on the
classification in Campbell (1996) are included in some regressions where indicated. The table reports OLS coefficient estimates and t-statistics based on robust standard errors (in
parentheses). **, *, and # indicate statistical significance at the 1%, 5%, and 10% levels (two-tailed), respectively.
 TABLE 5
Earnings Management and Reporting Incentives: Controlling for the
Degree of Tax Alignment and Remaining Differences in Accrual Rules
EMaggr,i = α0 + α1Conditional Variablei + α2Conditional Variablei*PUBLi
+ α3PUBLi + α4LEGALi + α5PREDICTi + εi
Incremental Effect Differential Effect
Variables Model 3 Model 4 Model 5
(N = 274) Model 1 Model 2 (Private Firms) (Public Firms) (Residual Values)
Panel A: Influence of Tax Alignment (N = 152) (N = 122)
TAX 14.201 ** 6.371 * 9.035 ** 2.933 6.289
(5.71) (2.43) (2.71) (0.69) (1.44)
TAX*PUBL (t-stat = -1.13) -10.687
(-1.57)
PUBL -16.894 ** -15.903 **
(-7.88) (-7.49)
LEGAL -5.351 ** -4.713 ** -6.244 **
(-5.21) (-3.44) (-4.00)
PREDICT 28.137 **
(9.15)
R2 8.7% 30.9% 19.3% 18.1% 31.6%
Panel B: Influence of Accrual Accounting Rules
ACCRUAL -57.621 ** -41.388 ** -29.677 ** -54.504 ** -12.917
(-7.58) (-5.68) (-2.72) (-5.66) (-0.83)
ACCRUAL* (t-stat = -1.71) -43.039 *
PUBL (-1.96)
PUBL -16.595 ** -15.607 **
(-8.19) (-7.38)
LEGAL -4.909 ** -5.140 ** -4.582 **
(-5.37) (-4.17) (-3.38)
PREDICT -75.756 **
(-8.93)
R2 14.6% 36.3% 20.1% 31.4% 32.0%
The sample comprises 274 industry-level observations from 13 European countries. The dependent variable,
EMaggr, is the average percentage rank across all four individual earnings management scores, EM1 to EM4, as
described in Table 1. PUBL is an indicator variable taking on the value of one if the observation stems from
publicly traded firms and zero otherwise. LEGAL stands for the mean of three variables in La Porta et al. (1998),
which measure the quality of the legal system and enforcement (i.e., efficiency of the judicial system, rule of law,
and corruption index). TAX is an indicator variable taking on the value of one if financial accounts for external
reporting and tax purposes are highly aligned and zero otherwise (see Alford et al., 1993, and Hung, 2001).
ACCRUAL is the accrual index from Hung (2001) (updated for European countries by Comprix et al., 2003), and
captures differences in accrual accounting rules across countries. The table reports OLS coefficient estimates and
t-statistics based on robust standard errors (in parentheses). All regressions include an intercept term. We evaluate
the difference between the coefficients on TAX or ACCRUAL across private and public firms with a t-test.
Model 5 uses the residuals from a regression of TAX (ACCRUAL) on the legal origins and the natural log of
GDP per capita as main effect and to construct interaction terms. To control for the general nature of the
institutional environment, the model also includes the predicted values (PREDICT) from this first stage
regressions. **, *, and # indicate statistical significance at the 1%, 5%, and 10% levels (two-tailed), respectively.