Managing earnings surprises in the US versus 12 other countries

Journal of Accounting and Public Policy 20 (2001) 373±398 www.elsevier.com/locate/jaccpubpol

Managing earnings surprises in the US versus 12 other countries Lawrence D. Brown

a,*

, Huong Ngo Higgins

b,1

a b

School of Accountancy, J. Mack Robinson College of Business, Georgia State University, Atlanta, GA 30302-4050, USA Department of Management, Worcester Polytechnic Institute, 100 Institute Road, Worcester, MA 01609, USA

Abstract We compare the distribution of earnings surprises in the US to those of 12 other countries. We expect US managers to be relatively more likely to manage earnings surprises due to di€erences in US corporate governance and legal environments versus those in other countries. An increasing emphasis on stock price performance in the US, as re¯ected by the rapid increase in stock and options compensation to US managers, and increases in litigation upon stock price drops, leads us to expect the tendency of US managers to manage earnings surprises versus those of non-US managers has increased in recent years. Our evidence is consistent with our expectations. We discuss the implications of our ®ndings for public policies to address the earnings surprise game. Ó 2001 Elsevier Science Ltd. All rights reserved.

1. Introduction The distribution of quarterly earnings surprises has changed dramatically in the US in recent years (Matsumoto, 2001, Table 1). The frequency of reported quarterly pro®ts exceeding analyst estimates by a small amount has increased considerably, while the frequency of reported quarterly losses falling short of analyst estimates by a large amount has decreased even more (Brown, 2001,

*

Corresponding author. Tel.: +1-404-651-0545; fax: +1-404-651-1033. E-mail addresses: [email protected] (L.D. Brown), [email protected] (H. Ngo Higgins). 1 Tel.: +1-508-831-5626; fax: +1-508-831-5720 0278-4254/01/$ - see front matter Ó 2001 Elsevier Science Ltd. All rights reserved. PII: S 0 2 7 8 - 4 2 5 4 ( 0 1 ) 0 0 0 3 9 - 4

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Table 1). Consistent with the views of the US Securities and Exchange Commission (Levitt, 1998, p. 14; Turner, 2000), US managers have increased their propensity to manage earnings surprises in recent years: they were more likely to report earnings that beat analyst estimates slightly, and less likely to report earnings that fell short of analyst estimates by a large amount. Managing earnings surprises is harmful because managers who are concerned about meeting short-term earnings targets forego long-term value-creating activities (Eccles et al., 2001, p. 95) such as by cutting R&D levels (Lev, 2001, p. 101). It also distorts corporate decision-making, compromises the integrity of corporate audits, and undermines capital markets (Collingwood, 2001, p. 67). Drawing on three literatures, corporate governance, investor protection and earnings surprise, we formulate two hypotheses. We expect that US managers are relatively more likely than non-US managers to play earnings surprise games during our sample period, by being relatively more likely to: (1) create small positive pro®t surprises relative to small negative pro®t surprises, and (2) avoid large negative loss surprises. 2 Consistent with our two hypotheses, we ®nd that US managers do manage earnings surprises to a relatively greater extent than do non-US managers, especially in recent years. Our ®ndings suggest that US managers focused relatively more on making quarterly earnings numbers, perhaps reducing their comparative advantage versus what it would have been in the absence of playing earnings surprise games. We compare earnings surprise management in the US with 12 other countries. Our approach facilitates an understanding of links between corporate countries' governance characteristics and earnings surprise management. Information asymmetry inherent in the US corporate governance system induces shareholders and sell-side analysts to attach great importance to current-term results (Jacobson and Aaker, 1993, p. 384), causing US managers to focus on earnings surprise games. Our ®ndings have implications for policies to alleviate earnings games by instituting and facilitating disclosures that reduce information asymmetry. 3 Another reason why it is important to compare the US

2 Unlike the extant literature, we focus on annual surprises because we compare the US with other countries, most of which do not report quarterly earnings. We examined pro®ts and losses separately rather than earnings because: (1) Brown (2001, Fig. 1) showed that small positive surprises pertain to pro®ts, not losses, and that large negative surprises pertain to losses, not pro®ts; and (2) a data analysis we undertook indicated that the frequency of pro®ts and losses varied over time and across countries. 3 Our view is consistent with: (1) ``Value Reporting: Beyond the Earnings Game'' (Eccles et al., 2001, Chapter 7), which calls for information on intangibles and non-®nancial value-drivers that are leading indicators of future ®nancial success; and (2) Lev (2001, Chapter 5), who advocates that an improved system of corporate disclosures should include information on intangible investments in ®rms' innovations and cover major phases of the value-chain.

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with other countries is to ascertain if the US results found by the extant literature are an artifact of the US or, instead, are a global phenomenon. Our paper proceeds as follows. We develop hypotheses in Section 2. Sample selection and methodology appear in Section 3. Section 4 presents the principal ®ndings, and Section 5 provides results of sensitivity tests. Section 6 concludes. 2. Hypotheses development We draw on three literatures, corporate governance, investor protection and earnings surprise, to formulate two hypotheses. The corporate governance literature suggests that US managers have strong incentives to enhance shortterm results, especially stock price. The investor protection literature suggests that litigation is a greater threat to US than it is to non-US managers. The earnings surprise literature shows that stock price movements in the US are signi®cantly positively related to the sign and magnitude of earnings surprises, providing strong incentives for US managers to create small positive surprises and avoid large negative ones. 2.1. Corporate governance The Berle and Means' model of corporate governance depicts ®rms as owned by dispersed shareholders (Berle and Means, 1932, Chapter IV), who lack the ability, skill and incentives to monitor manager performance (Roe, 1994, p. 6). As ownership is separate from management, information is distributed asymmetrically (Greenwald and Stiglitz, 1990, p. 160). Managers, who are better informed than shareholders about business prospects, shift earnings from a future time period to the current one (Jacobson and Aaker, 1993, p. 386). Investors are aware of this, but they are unable to determine its extent, pressuring all managers to emphasize current-term results. The information asymmetry makes investors unaware of long-term value-enhancing options, and induces them to attach more importance to current-term results than they otherwise would. US corporations have the most widely dispersed ownership (Shleifer and Vishny, 1997, p. 754), so the US has more information asymmetry between managers and stockholders than do other countries. Majority ownership is uncommon in the US, perhaps due to legal restrictions on ownership and control by banks, mutual funds, insurance companies and other institutions (Roe, 1994, pp. 21±22). 4 In large corporations of most other countries, 4

Ownership is also broadly dispersed in the UK, although it is more institutionally dominated than in the US (Black and Co€ee, 1994, p. 2002; Shleifer and Vishny, 1997, p. 754). Thus, it is not surprising that UK managers are relatively more likely than any other non-US managers to manage pro®t surprises (see Section 4.1 below).

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controlling shareholders exercise nearly complete control over managers (Shleifer and Vishny, 1997, pp. 754±755). 5 Due to the greater ties between shareholders and managers, information asymmetry is smaller in other countries (Jacobson and Aaker, 1993, p. 387). The agency problem in the di€use-ownership corporation is to minimize the sum of costs of aligning managers' and shareholders' incentives (Jensen and Meckling, 1976, pp. 308±309). Corporate governance mechanisms reduce agency problems, but managers focus on short-term performance due to information asymmetries. The three most important corporate governance mechanisms in the US are based on current stock price: independent directors, market for corporate control, and equity-based compensation to managers. German and Japanese ®rms, for example, have independent directors on their boards, but these relationships generally last a long time, placing less emphasis on short-term performance (Kaplan, 1994, p. 142; Kaplan and Minton, 1994, p. 226). Unfriendly mergers and acquisitions rarely occur outside the US and the UK (Shleifer and Vishny, 1997, pp. 756±757), and non-US managers are rarely compensated in stock and options (Share and Share Unalike, 1999, p. 18). Independent directors help to mitigate agency costs (Bacon and Brown, 1975, pp. 4, 7±8; Gilson and Roe, 1993, p. 877). In the US, shareholders bridge their separation from managers by electing non-employee directors to stimulate and oversee competition among top performers (Fama, 1980, p. 293). These directors play an important monitoring role, as seen by higher CEO turnover for poorly performing ®rms with independent board members (Weisbach, 1988, p. 431). 6 Non-US governance systems are less short-term oriented because their managers either are controlling shareholders or relationships between managers and stakeholders are maintained for a long time (Kaplan, 1994, p. 142), mitigating the importance of shortterm performance. In contrast, US managers stress short-term goals such as making the current earnings number (Eccles et al., 2001, pp. 94±95; Collingwood, 2001, p. 67). The market for corporate control serves as a governance mechanism (Manne, 1965, p. 112; Jensen, 1988, p. 28), by removing poorly performing managers (Martin and McConnell, 1991, p. 671). Undervalued ®rms become acquisition targets (Palepu, 1986, p. 18), creating a market-induced incentive 5 For example, German and Japanese corporate governance systems are bank-centered, where a main bank provides a signi®cant share of ®nance and governance to each ®rm (Allen and Gale, 2000, pp. 34±38). In France, cross-ownership and core investors are common (Shleifer and Vishny, 1997, p. 754). In most of the rest of the world, controlling owners are often founders or their o€spring (Shleifer and Vishny, 1997, p. 754). 6 See the American Law Institute (1992, pp. 143±149) for additional discussion on the e€ectiveness of independent directors.

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for managers to boost their ®rms' stock prices to retain their jobs. A popular way to boost stock price is to play the earnings game, whose rules include creating small positive surprises and avoiding large negative ones (Brown, 2001, Fig. 1). Executive compensation contracts, when combined with pressures of product and capital markets, monitor corporate managers (Hart, 1983, p. 381). In recent years, almost every large US ®rm has used equity as a management incentive; 10 years ago, only about half of all US ®rms did (Share and Share Unalike, 1999, p. 18). Japanese managers owned much less equity (Lichtenberg and Pushner, 1994, p. 246), and stock option plans were uncommon (Aoki, 1988, p. 254). In many European countries, the heavy tax on option gains wiped out incentives to issue options so there was considerably less focus on current stock price (Foreign Bodies, 1999, p. 54) and earnings. 7 2.2. Investor protection The legal system for investor protection, rather than corporate governance, helps resolve agency problems (La Porta et al., 2000, p. 4). Countries whose legal systems protect shareholders have relatively larger and broader capital markets (La Porta et al., 1997, p. 1131). Evidence based on 49 countries found that common-law (French civil-law) countries generally had the strongest (weakest) investor legal protections, and German±Scandinavian civil-law countries were in the middle (La Porta et al., 1998, p. 1113). Countries with strong property laws and enforcement mechanisms facilitate informed arbitrage and capitalization of ®rm-speci®c information (Morck et al., 2000, p. 217). The US ranks among the top of all countries in investor protection (La Porta et al., 1998, pp. 1130, 1136, 1142). Managers who knowingly make misleading or false forecasts are subject to liability under both the SEC Act of 1934 via Rule 10b-5 and the principles of common law. A typical 10b-5 lawsuit occurs after a steep stock decline, where plainti€s allege that they bought stock at in¯ated prices because managers either failed to disclose material adverse information or disseminated overly optimistic information (Kellogg, 1984, pp. 186±187; Francis et al., 1994, p. 137). The legal liability climate is more intense in the US than in any other country due to the American rule and its use of class action suits. The American rule requires each party to bear its own costs, inducing litigation (Hughes and Snyder, 1995, pp. 225, 249). In contrast, the English rule, used in most of the Western world, requires losers at trial to pay the winner's legal fees, reducing

7

UK managers had more stock and options than managers in other European countries but the amount was about one-twentieth that of US managers (Foreign Bodies, 1999, p. 54).

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the frequency of low-merit claims (Hughes and Snyder, 1995, p. 249). Legal liability is lower in France, Germany and Japan, whose judicial systems are less favorable to plainti€s than are the US and the UK, and the legal risk is low, but increasing (Frost, 1999, p. 8). Civil litigation is rare in Continental European countries compared to the US and the UK (Ball et al., 2000, pp. 13±14). Non-US managers providing negative earnings surprises faced fewer, less severe legal liabilities. Following the English rule, class action suits generally are not permitted in other countries (Romano, 1993, p. 126, note 21). Once a judge allows a case in the US to receive class action status, defendants often capitulate given the sheer number of plainti€s. The rate of class action suits keeps rising in the US despite laws to curb abusive litigation. 8 Plainti€s agree to settle less often than they used to, and they seek a larger percent recovery. 9 After a stock drops, litigious shareholders seek to be appointed lead plainti€s, as being so named means higher fees. 10 2.3. Earnings surprises Central to this literature is the concept that earnings are related to stock prices (Beaver, 1998, Chapter 5). The notion that share price equals the discounted value of expected future cash ¯ows underlies valuation models linking earnings to share prices (Feltham and Ohlson, 1995, p. 690), and earnings surprises to share price movements (Kormendi and Lipe, 1987, p. 324). In-

8 Congress passed the Private Securities Litigation Reform Act of 1995 to limit abusive securities litigation in US federal courts, but the number of lawsuits increased dramatically in recent years by plainti€s circumventing federal laws by ®ling in state courts (Grundfest and Perino, 1997, p. 2), with the number of cases ®led reaching a record high in 1998 (Levine and Pritchard, 1998, p. 19). Congress passed the Securities Litigation Uniform Standards Act of 1998 to limit redress through state courts (Casey, 1999, p. 142), but it is unlikely to reduce litigation costs (Levine and Pritchard, 1998, p. 24). 9 This may be attributable to uncertainty in the legal landscape. State versus federal court implementation of the pleading requirements may have caused the respective parties' valuation of particular cases to diverge (Feldman, 1998, p. 61). Plainti€s were required to disclose in the notice of settlement their potential claims, which pressured them to obtain a larger percent recovery than they would in the past. 10 Plainti€s used to try to be the ®rst to ®le a complaint against a company whose stock had dropped, because courts often appointed the ®rst-to-®le as the lead plainti€. Because this practice encouraged frivolous claims, the Reform Act of 1995 required parties ®ling class actions to publish notice allowing other class members to come forward so that the court could choose the one with the largest ®nancial interest, such as an institutional investor, to serve as lead plainti€. Institutional investors were reluctant to emerge, so plainti€s competed to issue the ®rst press release, hoping to attract enough members to represent the largest ®nancial interest (Pesner and Rossman, 1999, pp. 195±196). Securities class action lawsuits were not permitted in any of the other countries we examined.

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vestors trading on earnings surprises reap abnormal returns (Latane and Jones, 1977, Table 1; Foster et al., 1984, p. 574; Bernard and Thomas, 1989, p. 1), leading to the development of models to predict earnings surprises (Peters, 1993, p. 47; Brown et al., 1996, p. 17). The literature on earnings surprises is linked to studies on ®nancial analysts. Analysts are sophisticated investors (Debondt and Thaler, 1990, p. 52), whose earnings forecasts are more accurate (Brown et al., 1987a, p. 61) and represent market expectations better than do forecasts of time-series models (Brown et al., 1987b, p. 160). Managing earnings surprises in the US has increased in recent years (Matsumoto, 2001, Table 1). There has been a temporal increase in the propensity of US managers to report small positive quarterly pro®t surprises, and an even greater decrease in their tendency to report large negative quarterly loss surprises (Brown, 2001, Table 1). While there is evidence of US management of earnings surprises, it is unknown if the earnings surprise game is unique to the US or if non-US managers play similar games. Di€erences in US corporate governance and legal environments resulted in a greater focus in the US on current stock price, combined with evidence that US managers did manage earnings surprises, led us to expect that US managers were relatively more likely than non-US managers to manage earnings surprises, by creating small positive surprises and avoiding large negative ones. The increased emphasis on current stock price in the US in recent years, due to rapid increases in stock and option compensation and in securities litigation, led us to expect these phenomena to have been especially evident in recent years. Similar to Brown (2001, Fig. 1), we dichotomize ®rms into those reporting pro®ts and losses. Our two hypotheses in alternative form are: Hypothesis 1. US managers were relatively more likely than non-US managers to report pro®ts that exceeded rather than fell short of analysts' estimates by a small amount, especially in recent years. Hypothesis 2. US managers were relatively more likely than non-US managers to report losses that did not fall short of analysts' estimates by a large amount, especially in recent years.

3. Sample selection and methodology 3.1. Sample selection We accessed the Thomson Financial I/B/E/S Summary File for our three primary data items: analyst earnings forecasts, actual earnings and stock

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prices. We selected the last annual consensus (mean) forecast made prior to the annual earnings announcement. All non-US countries meeting two requirements were included: 1. Market capitalization of at least $200 billion in 1999 (US dollars). 2. Minimum of 100 observations available each year for the 12 years, 1988± 1999. Our ®rst criterion was chosen to ensure that the country's stock market was large enough to be of substantial interest to global investors. Our second criterion was selected to enable adequate sample sizes. Twelve countries satis®ed our criteria: Australia, France, Germany, Hong Kong, Italy, Japan, Netherlands, South Africa, Spain, Sweden, Switzerland, and the United Kingdom. 11 3.2. Methodology We examined the distributions of pro®t and loss surprises in the US versus 12 other countries. The ratio of the frequency of small positive pro®t surprises to that of small negative pro®t surprises (henceforth pro®t surprise ratio) was examined to test our ®rst hypothesis; and the ratio of the frequency of large negative loss surprises to that of total negative loss surprises (henceforth loss surprise ratio) was examined to test our second hypothesis. Earnings surprises were de®ned as actual minus predicted earnings, both divided by the absolute value of actual earnings. Surprises within 5% of reported pro®ts were termed small surprises. 12 Surprises greater than 100% of reported losses were termed large negative loss surprises. We assessed the tendency of managers to manage pro®t (loss) surprises by the pro®t (loss) surprise ratio in that country-period. We compared surprise ratios in the US with those in 12 countries in four consecutive three-year periods: 1988±1990, 1991±1993, 1994±1996, and 1997± 1999, and for the 12 years combined. To enhance validity of ®ndings, we based our comparisons on two tests: a binomial test, a non-parametric test, and a t-test, a parametric test. We conducted ®ve sensitivity analyses. First, to assess sensitivity to different de¯ators and cuto€s, we re-tested our hypotheses with an alternative de¯ator, stock price, and alternative cuto€s, 0.5% of stock price (Hypothesis 1) and 5% of stock price (Hypothesis 2). Second, to see if our results arose from di€erences in accounting systems, we determined if the US evidence was similar to other countries using the same (British±American) GAAP

11

Canada does not meet the second requirement. Many Canadian ®rms are cross-listed in the US, so the US ®rms subsume a large portion of the Canadian equity market. 12 Independent auditors consider errors of more than 5% to be material, and the market typically expects ®rms to beat analyst expectations in the range of 3±5% (Anonymous, 1998, p. 13).

L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398 381

system. Third, to examine if our results were due to the use of a particular source of actual earnings numbers, we repeated tests of Hypothesis 2 using a subset of our data where Thomson Financial I/B/E/S and another data source, Worldscope, agreed on the de®nition of actual earnings. Fourth, we examined if our Hypothesis 2 results were due to earnings in the US being more stable than in other countries by comparing the Thomson Financial I/ B/E/S stability index for the US loss observations with those of other countries. Our ®fth sensitivity analysis involved two multivariate logistic regressions based on ®rm-level data to assess the sensitivity to potentially correlated omitted variables. Our ®rst regression used small pro®t surprises, with small positive (negative) pro®t surprises coded as 1 (0); our second regression used negative loss surprises, with large (other) negative loss surprises coded as 0(1). Our independent variable of interest was US, a binary variable for country, coded 1 if US, 0 if non-US. We hypothesized that earnings surprise management was more pervasive in the US so we expected the US variable to be positive in both regressions. To determine if potentially correlated omitted variables a€ected our primary result that the US was more likely to play earnings surprise games, we added seven other explanatory variables: (1) accounting system; (2) Thomson Financial I/B/ E/S and Worldscope actual earnings agreed or disagreed; (3) earnings stability; (4) ownership dispersion; (5) reporting frequency; (6) litigation risk, and (7) ®scal year. Our primary ®ndings were robust to including these seven other explanatory variables.

4. Findings 4.1. Reported pro®ts Table 1 provides evidence regarding our ®rst hypothesis. The US pro®t surprise ratio increased monotonically from 1.20 (1988±1990) to 1.34 (1991±1993) to 1.92 (1994±1996) to 2.40 (1997±1999). Aside from Australia and Italy, no other non-US country had a monotonic increase in its pro®t surprise ratio during the sample period. In the aggregate, the mean pro®t surprise ratio for the other 12 countries increased in the ®rst two three-year periods, from 1.11 (1988±1990) to 1.28 (1991±1993), decreased in the third to 1.19 (1994±1996), and increased in the fourth to 1.33 (1997± 1999). To determine if pro®t surprise management was more pervasive in the US than in other countries, we compared the US pro®t surprise ratio with that of each of the 12 non-US countries using a binomial test. Table 1 shows the

382 L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398 Table 1 Ratio of small positive to small negative earnings surprises for pro®table ®rms (uses earnings as de¯ator and 5% cuto€ for small surprises) GAAP

Country

N

1988±90

1991±93

1994±96

1997±99

BA

Australia Hong Kong Netherlands South Africa United Kingdom

397 222 309 309 2473

0.72 1.00 1.22 1.26 1.83

1.15 1.26 1.11 1.23 1.53

1.28 1.11 1.47 1.20 1.55

1.38 0.86 1.49 1.03 1.81

1.16 1.08 1.30 1.17 1.68

CE

France Germany Italy Japan Spain Sweden Switzerland

496 205 144 1390 151 93 225

0.96 0.77 0.98 1.20 0.83 1.25 1.25

1.18 1.43 1.03 1.03 2.36 0.89 1.15

0.87 1.36 1.48 1.36 0.77 0.89 0.94

1.26 1.33 1.75 1.38 1.29 1.00 1.39

1.01 1.17 1.13 1.20 1.12 1.00 1.17

US

US 6458 1.20 Countries with ratios 6 smaller than the US ratio p-value of binomial test n/s (one-tailed test) BA Mean 1.21 CE Mean 1.03 Non-US mean 1.11 St. dev. of non-US mean 0.09 No. of st. dev. US ratio is 1.00 above the non-US mean p-value of t-test n/s (one-tailed test)

1.34 9

1.92 12

2.40 12

1.81 12

0.0002

1988±99

0.0537

0.0002

0.0002

1.26 1.30 1.28 0.11 0.55

1.32 1.10 1.19 0.08 9.13

1.31 1.34 1.33 0.08 13.38

1.28 1.11 1.18 0.05 12.60

n/s

<0.001

<0.001

<0.001

The ratio of small positive to small negative surprises for ®rms reporting pro®ts in the US and 12 other countries is shown in four consecutive three-year periods: 1988±90, 1991±93, 1994±96 and 1997±99, and for the entire 12-year sample period, 1988±99. Firms reporting pro®ts are de®ned as those whose reported annual earnings numbers (source: Thomson Financial I/B/E/S) exceed zero. A small positive surprise is de®ned as one wherein reported pro®ts exceed predicted earnings by no more than 5% of reported pro®ts. A small negative surprise is de®ned as one wherein predicted earnings exceed reported pro®ts by no more than 5% of reported pro®ts. N represents the numerator of the ratio, small positive to small negative surprises for pro®table ®rms for the 12-year sample period. The pro®t surprise ratio in each column is a pooled, temporal cross-sectional number where the data are pooled by country. For example, the 1.16 for Australia for the 12-year sample period represents 397 divided by 342. The numerator is those Australian ®rm-years reporting pro®ts exceeding predicted earnings by no more than 5% of reported pro®ts. The denominator is those Australian ®rm-years reporting pro®ts falling short of predicted earnings by no more than 5% of reported earnings. Predicted earnings are the I/B/E/S mean consensus analyst predictions in either the month of or the month prior to the annual earnings report. The n/s indicates a p-value that is not signi®cant at the 10% level (one-tailed test). BA and CE represent ®rms that adhere to the British±American and Continental European GAAP system, respectively. BA mean, CE mean, and Non-US mean are the means of the ®ve BA countries, seven CE countries, and all 12 non-US countries, respectively.

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number of countries with smaller pro®t surprise ratios than the US. It is evident that no other country had a lower pro®t surprise ratio than the US, either for the last two three-year periods or the entire period. The UK was closest to the US during these time frames. These two countries shared similar corporate governance and investor protection, excepting the American rule and securities class litigation. A binomial test rejected the null hypothesis that the US ratio was smaller than those of other countries for the entire sample and for each of the last two three-year periods (p ˆ 0:0002, one-tailed). 13 In contrast, in the ®rst three-year period, six countries had a smaller pro®t surprise ratio than did the US, an insigni®cant result. In the second three-year period, nine countries had a smaller ratio than did the US, a marginally signi®cant result (p ˆ 0:0537, onetailed). In sum, managing pro®t surprise was more evident in the US than in any other country we examined in the second half, but not the ®rst half of the sample period. To enhance validity, we compared the US pro®t surprise ratio with the mean ratios of 12 other countries using t-tests. We treated each country as an independent observation, calculating the mean and standard deviation of the 12 pro®t surprise ratios. Table 1 contains the mean and standard deviation of the distribution of 12 other countries' pro®t surprise ratios. The t-test results closely mirrored the binomial test results. For the last two threeyear periods and for the entire 12-year period, the US ratio was signi®cantly greater than the non-US mean, suggesting the US was relatively more likely to report small pro®t surprises. It was 9.13, 13.38 and 12.6 standard deviations above the non-US mean for the last two three-year periods and for the 12-year period (p < 0:001, one-tailed). 14 In contrast, it was only 1.00 and 0.55 standard deviations above the non-US mean in the ®rst three-year time periods. Overall, our results strongly support Hypothesis 1. Pro®t surprise management was more evident in the US than in all other countries, especially in recent years. US managers managed pro®t surprises by shifting surprises from the negative to the positive quadrant (Burgstahler and Eames, 1998, pp. 19± 22). They had not always done so; we did obtain signi®cant results in the ®rst two three-year periods. 4.2. Reported losses Table 2 contains evidence pertinent to our second hypothesis. The US loss surprise ratio decreased monotonically from 0.40 (1988±1990) to 0.31 (1991±

13 14

The probability that a fair coin yields 12 consecutive heads equals 0.0002. The last number, 12.6, was calculated as …1:81 1:18†=0:05.

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1993) to 0.28 (1994±1996) to 0.23 (1997±1999). None of the non-US countries exhibited a monotonically decreasing loss surprise ratio. In the aggregate, the average ratio for the other 12 countries decreased from 0.53 (1988±1990) to 0.35 (1991±1993), increased to 0.41 (1994±1996), and then to 0.54 (1997± 1999). To determine if US managers managed loss surprises more than did managers in 12 other countries, we ®rst compared the US loss surprise ratio with that of each country using a binomial test. For the most recent three years, the US had a smaller ratio than did 11 countries (p ˆ 0:0029, one-tailed). Japan, the only country whose managers were strongly encouraged to forecast earnings, was the exception (Japan, 0.21 versus the US, 0.23). In the second-to-last three-year period and in the entire sample period, the US had a smaller ratio than did 10 countries (p ˆ 0:0161, one-tailed): Japan and Switzerland were the exceptions. In the ®rst three-year period, the US had a smaller ratio than did nine countries (p ˆ 0:0537, one-tailed), but in the second three-year period, it had a smaller ratio than only six. Thus, managing loss surprise was more evident in the US than in at least 10 other countries in the second half, but not the ®rst half of the sample period. We also conducted t-tests comparing the US loss surprise ratio with the mean ratios of all other countries. The t-test results closely mirrored the binomial test ®ndings. For the entire sample period and for each of the last two three-year periods, the US loss surprise ratio was signi®cantly smaller (p < 0:05, one-tailed) than the mean loss surprise ratio of the distribution of the 12 non-US countries, suggesting that the US was relatively less likely to report large loss surprises. More speci®cally, it was 2.60, 6.20 and 3.25 standard deviations below the non-US mean for the last two three-year periods and for the 12-year period. In contrast, it was 2.17 and 0.80 standard deviations below the non-US mean in the ®rst two three-year periods (p < 0:10, onetailed, and insigni®cant, respectively). Overall, the results support our second hypothesis. Loss surprise management was more evident in the US than in other countries, especially in recent years. US managers managed earnings surprises by mitigating the frequency of large loss surprises. They had not always done so; our results were insigni®cant for the second three-year time period. Our results are consistent with US managers, in recent years, forewarning capital markets of bad news to stave o€ securities litigation, boost their stock prices, and to keep (or drive) their options in the money. 15

15 So€er et al. (2000, Table 1) showed that stock price was higher at the time of earnings announcements if managers preannounce bad news than if they waited to announce the bad news at the time of the earnings report.

L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398 385 Table 2 Ratio of large negative to total negative earnings surprises for loss ®rms (uses absolute value of earnings as de¯ator and 100% cuto€ for large negative surprises) GAAP

Country

N

1988±90

BA

Australia Hong Kong Netherlands South Africa United Kingdom

58 71 37 34 326

0.75 0.67 0.29 1.00 0.63

0.21 0.75 0.34 0.50 0.44

0.48 0.87 0.46 0.31 0.50

0.42 0.63 0.45 0.70 0.44

0.44 0.72 0.39 0.61 0.47

CE

France Germany Italy Japan Spain Sweden Switzerland

118 103 49 273 33 63 28

0.53 0.29 0.44 0.52 0.47 0.50 0.23

0.33 0.39 0.28 0.29 0.23 0.18 0.31

0.36 0.42 0.30 0.21 0.38 0.43 0.24

0.58 0.55 0.75 0.21 0.83 0.47 0.40

0.42 0.43 0.32 0.24 0.34 0.39 0.29

US

US 1110 Countries with ratios larger than the US ratio p-value of binomial test (one-tailed test) BA Mean CE Mean Non-US mean St. dev. of non-US mean No. of st. dev. US ratio is below the non-U.S. mean p-value of t-test (one-tailed test)

0.40 9

0.31 6

0.28 10

0.23 11

0.29 10

0.0537

n/s

0.0161

0.0029

0.0161

0.67 0.38 0.53 0.06 2.17

0.45 0.32 0.35 0.05 0.80

0.52 0.30 0.41 0.05 2.60

0.53 0.49 0.54 0.05 6.20

0.53 0.36 0.42 0.04 3.25

<0.05

1991±93

n/s

1994±96 1997±99 1988±99

<0.05

<0.001

<0.01

The ratio of large negative to total negative earnings surprises for ®rms reporting losses in the US and 12 other countries is shown in four consecutive three-year periods: 1988±90, 1991±93, 1994±96 and 1997±99, and for the entire 12-year sample period, 1988±99. Firms reporting losses are de®ned as those whose reported annual earnings numbers (source: Thomson Financial I/B/E/S) are less than zero. A small positive surprise is de®ned as one wherein reported pro®ts exceed predicted earnings by no more than 5% of reported pro®ts. Large negative loss surprises are de®ned as cases wherein reported losses exceed predicted earnings by at least 100% of the absolute value of the reported loss. N represents the numerator of the ratio, large negative loss surprises to total negative loss surprises for the 12-year sample period. The ratio in each column is a pooled, temporal crosssectional number where the data are pooled by country. For example, the 0.44 for Australia for the 12-year sample period represents 58 divided by 132. The numerator is those Australian ®rm-years that do report losses for which analysts make an error of at least 100%. The denominator is those Australian ®rm-years reporting losses that give rise to negative surprises. Predicted earnings are the I/B/E/S mean consensus analyst predictions in either the month of or the month prior to the annual earnings report. The n/s indicates a p-value that is not signi®cant at the 10% level (one-tailed test). BA and CE represent ®rms that adhere to the British±American and Continental European GAAP system, respectively. BA Mean, CE Mean, and Non-US mean are the means of the ®ve BA countries, seven CE countries, and all 12 non-US countries, respectively.

386 L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398

5. Sensitivity analyses We conducted ®ve sensitivity analyses. First, we examined if our results were robust to an alternative de¯ator and cuto€ for earnings surprises. Second, we investigated if our ®ndings were an artifact of the accounting system, British± American (BA) or Continental European (CE). Third, we determined if our results were due to di€erences between Thomson Financial I/B/E/S and another data source, Worldscope, in de®ning earnings. Fourth, we examined if earnings in the US were more stable than in other countries. Fifth, we used logistic regressions to test our two hypotheses in cross-sectional, multivariate settings. 5.1. An alternative de¯ator and cuto€s for earnings surprises Our results to this point used reported pro®t (Table 1) and the absolute magnitude of reported loss (Table 2) as de¯ators. There are two problems with this de¯ator: (1) the ratio explodes when earnings approach zero; and (2) the ratio is unde®ned [and hence the observation is omitted] when earnings equal zero. To examine the robustness of our results to an alternative de¯ator and cuto€s, Table 3 contains results for the entire sample period using stock price as the de¯ator. Small pro®t surprise was de®ned as a pro®t surprise within 0.5% of stock price (panel A), and large loss surprise was de®ned as a loss surprise at least 5% of stock price (panel B). We obtained similar results with many other analyses using other cuto€s based on actual earnings and price as de¯ators (available upon request). Panels A and B provide results for pro®t and loss surprises, respectively, for the entire sample period. The panel A results are similar to the Table 1 ®ndings. The US pro®t surprise ratio was larger than that of 11 non-US countries (p ˆ 0:0029, one-tailed), nine standard deviations above the mean ratio of the non-US countries (p < 0:001, one-tailed). The panel B results were similar to the Table 2 ®ndings. The US had a smaller ratio than those of 11 non-US countries (p ˆ 0:0029, one-tailed), four standard deviations below the mean ratio of the non-US countries (p < 0:01, one-tailed). 5.2. Alternative GAAP systems Our results may be due to inherent di€erences in GAAP models, i.e., the processes for standard setting and enforcement. GAAP system, characterized as common-law, a market process, or codi®ed-law, a govermental process, cause international di€erences in accounting income numbers (Ball et al., 2000, p. 13). Accounting in the 12 non-US countries can be classi®ed as adhering either to the British±American (BA) accounting model, consisting of Australia,

L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398 387 Table 3 Ratio of small positive to small negative surprises for pro®table ®rms. Ratio of large negative surprises to total negative surprises for loss ®rms (uses stock price as de¯ator, 0.5% cuto€ for small surprises, and 5% cuto€ for large negative surprises) GAAP

Country

N

Pro®t Surprise Ratio

N

Loss Surprise Ratio

BA

Australia Hong Kong Netherlands South Africa United Kingdom

493 250 370 353 2985

1.13 1.11 1.35 1.18 1.75

78 89 65 45 458

0.59 0.90 0.69 0.80 0.66

CE

France Germany Italy Japan Spain Sweden Switzerland

690 392 197 3161 202 161 297

1.05 1.09 0.86 1.10 1.11 1.11 1.10

177 152 115 406 70 84 60

0.64 0.63 0.74 0.35 0.71 0.52 0.62

US

US 8400 Countries with pro®t (loss) surprise ratios smaller (larger) than the US ratio p-value of binomial test (one-tailed test) BA mean CE mean Non-US mean St. dev. of non-US mean No. of st. dev. US pro®t (loss) ratio is above (below) the non-US mean p-value of t-test (one-tailed test)

1.70 11

1871

0.49 11

0.0029

0.0029

1.30 1.06 1.16 0.06 9.00

0.73 0.60 0.65 0.04 4.00

<0.001

<0.005

The ratio of small positive to small negative surprises for ®rms reporting pro®ts and the ratio of large negative to total negative surprises for ®rms reporting losses in the US and 12 other countries are shown for the entire 12-year sample period, 1988±99. Firms reporting pro®ts (losses) are de®ned as those whose reported annual earnings numbers (source: Thomson Financial I/B/E/S) exceed (are less than) zero. A small positive (negative) surprise is de®ned as one wherein reported pro®ts exceed (fall short of) predicted earnings by no more than 0.5% of stock price. Large negative loss surprises are de®ned as cases wherein reported losses exceed predicted earnings by at least 5% of stock price. Total negative loss surprises are de®ned as those wherein predicted earnings exceed reported earnings. The N represent the numerator of the ratios, small positive to small negative surprises for pro®table ®rms and large negative to total negative surprises for loss ®rms. The ratio in each column is a pooled, temporal cross-sectional number where the data are pooled by country. For example, the 1.13 for Australia for pro®t ®rms represents 493 divided by 436, and the 0.59 for Australia for loss ®rms represents 78 divided by 132. Predicted earnings are the I/B/E/S mean consensus analyst predictions in either the month of or the month prior to the annual earnings report. BA and CE represent ®rms that adhere to the British±American and Continental European GAAP system, respectively. BA Mean, CE Mean, and Non-US mean are the means of the ®ve BA countries, seven CE countries, and all 12 non-US countries, respectively.

388 L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398

Hong Kong, the Netherlands, South Africa, and the United Kingdom; or to the Continental European (CE) accounting model, consisting of France, Germany, Italy, Japan, Spain, Sweden, and Switzerland (Ball et al., 2000, p. 20; Mueller et al., 1997, exhibits 1±4). The US adheres to the BA system so comparing its pro®t and loss surprise ratios to those of other BA countries can determine if the above US results are an artifact of its using a particular GAAP system. The Table 1 data showed that the BA pro®t surprise ratio was higher than the CE pro®t surprise ratio for the 12 years combined (1.28 versus 1.11), and for two of the four three-year periods. The US uses the BA accounting system so this result is consistent with the notion that its relatively large pro®t surprise ratio is an artifact of its use of the BA system, but two facts are inconsistent with this notion: (1) the US ratio is much higher than the BA surprise ratio (1.81 versus 1.28) for the 12 years combined, and (2) in contrast to the US ratio, which increased monotonically over the sample period (see Section 4.1), the BA ratio did not: it increased from 1.21 (1988±1990) to 1.26 (1991±1993) to 1.32 (1994±1996), then decreased to 1.31 (1997±1999). The Table 2 data revealed that the US ratio of 0.29 for the 12 years combined was lower than that of either the BA or the CE group (0.53 and 0.36, respectively), and the larger ratio for the BA group was inconsistent with the notion that the US ®nding was due to its use of BA accounting. Additional evidence inconsistent with the view that the US results simply mirrored the BA results was that the US ratio decreased monotonically over the 12-year period (see Section 4.2), while the BA ratio did not. 5.3. Thomson Financial I/B/E/S actuals may di€er from those of other sources Thomson Financial I/B/E/S strives to put actual and forecasted earnings on a common footing, but it was likely to be relatively more successful at doing so when ®rms did not include non-recurring or unusual events in their earnings reports. 16 Non-recurring or unusual events were generally negative, often occurring when earnings were unusually poor (Elliott and Hanna, 1996, Table 2). If Thomson Financial I/B/E/S was more 16 According to I/B/E/S (1999, pp. 7±8): ``With very few exceptions analysts make their earnings forecasts on a continuing operations basis. This means that I/B/E/S receives an analyst's forecast after discontinued operations, extraordinary charges, and other non-operating items have been backed out. While this is far and away the best method for valuing a company, it often causes a discrepancy when a company reports earnings. I/B/E/S adjusts reported earnings to match analysts' forecasts on both an annual and quarterly basis. This is why I/B/E/S actuals may not agree with other published actuals, i.e., Compustat.''

L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398 389 Table 4 Ratio of large negative to total negative earnings surprises for loss ®rms (actual reported annual losses obtained from Worldscope) GAAP

Country

N

1988±90 1991±93 1994±96 1997±99 1988±99

BA

Australia Hong Kong Netherlands South Africa United Kingdom

22 14 33 11 83

0.80 1.00 0.42 1.00 0.65

0.25 N/A 0.29 0.57 0.51

0.25 1.00 0.42 0.67 0.38

0.65 0.71 0.43 0.60 0.40

0.49 0.88 0.37 0.65 0.47

CE

France Germany Italy Japan Spain Sweden Switzerland

4 92 10 54 3 26 16

0.33 0.40 0.67 0.67 0.50 0.40 0.64

0.33 0.61 0.25 0.48 0.50 0.28 0.33

0.25 0.50 1.00 0.36 N/A 0.42 N/A

0.33 0.69 0.50 0.36 N/A 0.58 1.00

0.31 0.58 0.42 0.41 0.50 0.43 0.48

US

US Countries with ratios larger than the U.S. ratio p-value of binomial test (one-tailed test) BA mean CE mean Non-US mean St. dev. of non-US mean No. of st. dev. US ratio is below the non-US mean p-value of t-test (one-tailed test)

248

0.46 8

0.34 5

0.28 8

0.18 11

0.26 12

n/s

n/s

n/s

0.77 0.48 0.62 0.07 2.29

0.40 0.33 0.40 0.04 1.50

0.54 0.25 0.52 0.09 2.67

<0.05

<0.10

<0.05

0.0029

0.0002

0.56 0.67 0.57 0.06 6.50

0.57 0.40 0.50 0.04 6.00

<0.001

<0.001

The ratio of large negative surprises to negative surprises for ®rms reporting losses (loss surprise ratio) in the US and 12 other countries is shown in four consecutive three-year periods: 1988±90, 1991±93, 1994±96 and 1997±99, and for the entire 12-year sample period, 1988±99. Firms reporting losses are de®ned as those whose reported annual earnings numbers (source: Worldscope ®eld 5201 and Thomson Financial I/B/E/S) are less than zero, but whose extraordinary charge resulting from non-recurring or unusual events equals zero (source: Worldscope ®eld 1254). Extreme negative loss surprises are de®ned as cases wherein reported losses exceed predicted earnings by at least 100% of the absolute value of the reported loss. N represents the numerator of the ratio for the 12-year sample period. The ratio in each column is a pooled, temporal cross-sectional number where the data are pooled by country. For example, the 0.49 for Australia for the 12-year sample period represents 22 divided by 45. Predicted earnings are the I/B/E/S mean consensus analyst predictions in either the month of or the month prior to the annual earnings report. BA and CE represent ®rms that adhere to the British±American and Continental European GAAP system, respectively. BA Mean, CE Mean, and non-US mean are the means of the ®ve BA countries, seven CE countries, and all 12 non-US countries, respectively.

390 L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398

successful at placing actual and forecast earnings on a common footing for US ®rms, we may have committed a type 1 error: erroneously concluding that US managers were more likely to manage loss surprises when they were not. Non-recurring or unusual events are least likely to occur when another data source agrees with the Thomson Financial I/B/E/S de®nition of earnings. We used Worldscope as an alternative source of actual earnings (item 5201). Because Worldscope treats extraordinary items di€erently for di€erent countries, we omit ®rms reporting extraordinary charges for this test (item 1254). Table 4 replicates our Table 2 analyses based on this subset of our data. Our Table 4 results are very similar to those in Table 2. The US loss surprise ratio decreased every period, from 0.46 (1988±1990) to 0.34 (1991±1993) to 0.28 (1994±1996) to 0.18 (1997±1999). In contrast, the non-US mean decreased from 0.62 (1988±1990) to 0.40 (1991±1993); then increased to 0.52 (1994±1996) and again to 0.57 (1997±1999). For the most recent three-year period, the US had a smaller ratio than 11 countries (p ˆ 0:0029, one-tailed) did, 6.50 standard deviations below the non-US mean (p < 0:001, one-tailed). Moreover, the US had the smallest ratio of all countries over the 12-year period combined (p ˆ 0:0002, one-tailed), 6.00 standard deviations below the non-US mean (p < 0:001, one-tailed). 5.4. Stability indices US earnings may have been relatively more stable than in other countries, making it easier for US managers to avoid extreme negative surprises when reporting losses (Hypothesis 2). If so, our Table 2 ®ndings may have been due to US earnings being relatively more stable over our study period, rather than to US managers consciously choosing to mitigate large negative loss surprises. To assess the validity of this argument, we examined the median Thomson Financial I/B/E/S stability index, for each country when ®rms report losses. 17 We pooled our data temporally and cross-sectionally. In contrast to the notion that US earnings are relatively more stable, US ®rms reporting losses had relatively less stable earnings than did all other countries (p ˆ 0:0002, onetailed). The median stability indices for each country were: Australia (27.99), France (18.11), Germany (19.50), Hong Kong (23.84), Italy (19.31), Japan (31.71), Netherlands (28.21), South Africa (32.04), Spain (33.81), Sweden

17

The Thomson Financial I/B/E/S stability measure equals the mean absolute percentage di€erence between the actual reported earnings per share and a ®ve-year historical EPS growth trend line, expressed as a percentage of trend line earnings per share. The smaller the stability index, the greater is the stability.

L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398 391

(29.27), Switzerland (25.70), United Kingdom (25.55) and United States (49.79). 5.5. Multivariate logit analyses Having shown that our results were robust to four potential validity threats examined in a univariate fashion (Sections 5.1±5.4), we next conducted multivariate analyses. Table 5 presents the results of multivariate logistic regressions for our two hypotheses. The independent variables included three represented in Sections 5.1±5.4 plus four others. Each regression had the form: SM ˆ a1 GAAP ‡ a2 CODE ‡ a3 STAB ‡ a4 ODISP ‡ a5 FREQ ‡ a6 LITR ‡ a7 YEAR ‡ a8 US where · SM is the surprise management, coded 1 in the ®rst logistic regression if pro®t surprise is small and positive, and 1 in the second logistic regression if loss surprise is negative and not large; 0 otherwise. · GAAP is the type of accounting system, coded 1 if BA, and 0 if CE (Source: Ball et al., 2000). [Expected sign: positive for both regressions because BA countries, being market-oriented, are more focused on current-term results.] · CODE is the coding of earnings before special items, coded 1 if Worldscope and Thomson Financial I/B/E/S actuals agree, and 0 otherwise. (Sources: Worldscope item # 5201 and item ``FY-0 actual EPS'', Thomson Financial I/B/E/S Background data ®le). [Expected sign: positive for second regression because coding di€erences are likely to create large surprises, unknown sign for ®rst regression.] · STAB is the earnings stability, measured by the Thomson Financial I/B/ E/S stability index. (Source: item ``5-Year EPS stability'' from Thomson Financial I/B/E/S Background data ®le). [Expected sign: negative for second regression because ®rms with more stable earnings are less likely to exhibit large negative loss surprises, unknown sign for ®rst regression.] · ODISP is the ownership dispersion, the ratio of public market capitalization and total market capitalization. (Source: Worldscope). [Expected sign: positive for both regressions because ®rms with disperse ownership focus more on short-term performance and earnings surprise management.] · FREQ is the earnings reporting frequency, coded 1 if annual, 2 if semiannual, and 4 if quarterly. (Source: item ``Periodicity'' from Thomson Financial I/B/E/S actual ®le). [Expected sign: positive for both regressions because ®rms reporting frequently are more likely to manage earnings surprises.]

392 L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398 Table 5 Multivariate logistic regression results Panel A: Test of Hypothesis 1 Number of observations Chi-Square for model (eight degrees of freedom) p-value Independent variable Predicted Estimated relation coecients

10528 191.84 <0.0001 Standard errors

Intercept GAAP CODE STAB ODISP FREQ LITR YEAR US

0.754 0.056 0.045 0.000 0.094 0.027 0.086 0.008 0.092

+ ? ? + + + + +

)4.538 0.223 )0.053 )0.000002 0.140 0.051 0.364 0.048 0.165

Panel B: Test of Hypothesis 2 Number of observations Chi-square for model (eight degrees of freedom) p-value Independent variable Predicted Estimated relation coecients

2581 144.1251 <0.0001 Standard errors

Intercept GAAP CODE STAB ODISP FREQ LITR YEAR US

1.552 0.121 0.093 0.000 0.204 0.091 0.135 0.016 0.294

+ + ) + + + + +

)4.202 )1.002 0.151 )0.000001 0.812 )0.009 0.012 0.049 1.111

Chi-square p-value 36.246 16.057 1.359 6.950 2.211 3.650 17.799 39.018 3.235

<0.0001 <0.0001 0.2437 0.0084 0.1371 0.0561 <0.0001 <0.0001 0.0721

Chi-square p-value 7.327 68.923 2.646 0.623 15.827 0.010 0.007 9.307 14.262

0.0068 <0.0001 0.1038 0.4301 <0.0001 0.9202 0.9315 0.0023 0.0002

SM ˆ a1 GAAP ‡ a2 CODE ‡ a3 STAB ‡ a4 ODISP ‡ a5 FREQ ‡ a6 LITR ‡ a7 YEAR ‡ a8 US. SM ˆ 1 in Panel A regression if the small pro®t surprise is positive and 1 in Panel B regression if the negative loss surprise is not large; otherwise 0 [sign equals 1 when surprise management occurs]. GAAP ˆ 1 if the country uses the BA system of accounting; 0 if it uses the CE system of accounting. CODE ˆ 1 if I/B/E/S and Worldscope actuals agree; 0 if they disagree. STAB ˆ I/B/E/S stability index. ODISP ˆ public market capitalization/total market capitalization. FREQ ˆ earnings reporting frequency, coded 1 for annual, 2 for semi-annual, and 4 for quarterly. LITR ˆ technology ®rm coded 1; 0 otherwise. YEAR ˆ 1988±1999 for ®scal year end 1988±1999. US ˆ 1 if US; 0 otherwise.

L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398 393

· LITR is the litigation risk, coded 1 if ®rm is in the technology sector, i.e., with high litigation risk; and 0 otherwise (Source: Worldscope's Dow Jones Market Sector). [Expected sign: positive for both regressions because technology ®rms have more litigation risk, increasing their propensity to manage earnings surprises.] 18 · YEAR is the ®scal year, 1988±1999 (Source: Thomson Financial I/B/E/S). [Expected sign: positive for both regressions because earnings surprise management is expected to have increased in recent years.] · US is coded 1 if US, 0 otherwise (Source: Thomson Financial I/B/E/S). [Expected sign: positive for both regressions because earnings surprise management is expected to be more pervasive in the US.] Panel A shows that, as expected, pro®t surprise management was signi®cantly positively related to GAAP, FREQ, LITR, YEAR, and to US. ODISP had its expected positive sign, but it was insigni®cant. CODE and STAB were not predicted a priori, but they had an insigni®cant and a signi®cant negative coecient, respectively. The model was signi®cant (p < 0:0001, one-tailed), based on 10,528 observations. Our primary result holds: the US variable remains positive and signi®cant after including seven potentially confounding factors. Panel B shows that, as expected, loss surprise management was signi®cantly positively related to ODISP, YEAR, and to US. As expected, loss surprise management was positively related to CODE at a nearly signi®cant level (p ˆ 0:1038, one-tailed). STAB and LITR had their expected signs, but both were insigni®cant (p ˆ 0:4301 and 0.9315, one-tailed, respectively). FREQ had an unexpected sign but it was insigni®cant (p ˆ 0:9202, one-tailed). Contrary to our expectations, but consistent with Table 2, loss surprise management was signi®cantly negatively related to GAAP, indicating that ®rms using the BA system of GAAP were less likely to manage loss surprises. The model, based on 2581 observations, was signi®cant (p < 0:0001, one-tailed). Our primary result holds: the US variable remains positive and signi®cant after including seven potentially confounding factors.

6. Conclusions We obtained two primary ®ndings. First, US managers who report pro®ts, especially in recent years, were relatively more likely than managers in all 12 non-US countries we examined to instigate small positive surprises and suppress small negative surprises. Second, US managers who reported losses, 18

According to Grundfest and Perino (1997, p. 3), technology companies are the most frequent targets of litigation.

394 L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398

especially in recent years, were relatively more likely than managers in most other countries to quell large negative surprises. Our analyses of loss surprises identi®ed one country whose loss surprise ratio was smaller than the US; Japan, the only country whose managers were strongly encouraged to forecast current annual earnings. As those managers who forecasted earnings were unlikely to subsequently report losses that resulted in large negative surprises, our basic premise holds. We considered ®ve potential validity threats, and rejected them all. First, we considered an alternative de¯ator and cuto€s, and obtained qualitatively similar results. Second, we explored if our results were due to a particular GAAP model. They were not. Third, we examined if our results concerning large loss surprises were attributable to the Thomson Financial I/B/E/S de®nition of actual earnings numbers. Finding our results to be robust to a subset of our data wherein Thomson Financial I/B/E/S and Worldscope agreed on the de®nition of actual earnings, we ruled this out as a confounding factor. Fourth, we determined if US earnings were relatively more stable, providing a potential measurement error explanation for our ®nding concerning large loss surprises. We found the opposite, namely US earnings for the loss sample were less stable than in all 12 other countries. As a ®fth sensitivity analysis, we conducted two multivariate logistic regressions, one for each hypothesis. We focused on the binary variable, US, equal to 1 for the US and 0 for other countries, and included seven control factors to ensure that our primary results were not due to potentially correlated omitted variables. Three control variables were included in the sensitivity analyses described above: GAAP (BA coded 1; CE coded 0); CODE (1 if Thomson Financial I/B/E/S and Worldscope actuals agreed; 0 if they disagreed); and STAB (the larger the number, the more unstable the earnings). We included four other control variables: ODISP, FREQ, LITR and YEAR, respectively, measuring ownership dispersion, reporting frequency, litigation risk, and the ®scal year end, 1988±1999. We found that the US variable had its expected positive sign and was signi®cant in both regressions, suggesting that our results were not due to the seven potentially correlated omitted variables we considered. We did not use direct proxies of the corporate governance and legal environment, so we cannot unambiguously attribute earnings surprise management to corporate governance and legal environments. Managing earnings surprises is harmful because managers concerned about meeting short-term earnings targets forego long-term value-creating activities (Eccles et al., 2001, pp. 95±96) such as R&D (Lev, 2001, p. 101). It is also harmful because it distorts corporate decision-making, compromising the integrity of corporate audits and undermining capital markets (Collingwood, 2001, p. 67).

L.D. Brown, H. Ngo Higgins / Journal of Accounting and Public Policy 20 (2001) 373±398 395

Policy-makers are concerned that earnings games have adverse consequences for America's ®nancial reporting system and diminish the market's strength and success (Levitt, 1998, p. 14). Our ®ndings suggest that policymakers aiming at mitigating earnings surprise management in the US should reduce information asymmetry by instituting and facilitating disclosures of long-term value-creating activities. The role of intellectual assets as drivers of today's wealth and growth emphasize the urgency of improving disclosures about intangibles (Lev, 2001, p. 132). Investors, who see how companies really create value, are likely to place less emphasis on current-term results, putting less presssure on managers, analysts and auditors to play earnings games (Eccles et al., 2001, pp. 95±96).

Acknowledgements We are grateful to the research assistance of Indrarini Laksmana, Emad Mohammad and the comments of Sudipta Basu, Carol Frost, Trevor Harris, Amy Lau, Siva Nathan, Grace Pownall, Gordon Richardson, participants of the 1999 Southeast Summer Accounting Research Colloquium, a research workshop at American University, and two anonymous reviewers. The authors gratefully acknowledge the contribution of Thomson Financial for providing earnings per share forecast data, available through the Institutional Brokers Estimate System. These data have been provided as part of a broad academic program to encourage earnings expectation research.

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