How do insider trading incentives shape nonfinancial disclosures? Evidence from product and business expansion disclosures

Prior literature investigates how managers exploit their discretion over corporate disclosures for opportunistic purposes. For instance, Lang and Lundholm (2000) provide evidence that firms release more good news for a higher stock price before equity offerings. Brockman et al. (2008) show that managers tend to release pessimistic earnings forecasts to deflate stock prices before stock repurchases. Ge and Lennox (2011) find that managers withhold bad earnings news before stock-for- stock mergers. Yermack (1997) and Aboody and Kasznik (2000) document that managers strategically disclose bad news to deflate stock prices before stock option grants to maximize option values. Several studies explore strategic disclosures by managers in the insider trading settings that exclude option grants. Cheng and Lo (2006) find that managers release more bad earnings news prior to purchasing shares of their firms. In parallel, Rogers (2008) find some weak evidence that managers provide lower quality disclosures prior to purchasing shares than they would in the absence of insider trades. Cheng et al. (2013) find that managers tend to release more precise earnings forecasts for good (bad) news than for bad (good) news before selling (buying) shares. Overall, the evidence in this literature indicates that corporate disclosures, which occur shortly before price-relevant events, are subject to the opportunistic incentives of managers. Put differently, it is the opportunistic incentives for equity offerings, stock repurchases, stock-for-stock mergers, option grants, and insider trades, that drive disclosure behaviors, rather than the disclosures causing those events to take place ex post. As such, hypothetically, reverse causality is less concerned in this strategic-disclosure literature; so too is this study which looks at PBE disclosures made before insider trades.

This paper is also related to a growing body of works on opportunistic insider trades. This literature provides evidence that managers tend to trade stocks before disclosures of a wide range of price-relevant corporate news; these include earnings announcements (e.g., Ke et al. 2003; Piotroski and Roulstone 2005), management earnings forecasts (Penman 1982, 1985; Billings and Cedergren 2015), earnings misstatements (e.g., Badertscher et al. 2011; Agrawal and Cooper 2015), equity offerings (e.g., Karpoff and Lee 1991), stock repurchases (e.g., Lee et al. 1992), dividend announcements (John and Lang 1991), bankruptcies (Gosnell et al. 1992; Seyhun and Bradley 1997), disclosures of internal control weaknesses (Skaife et al. 2013), as well as other corporate news releases (Elliott et al. 1984; Givoly and Palmon 1985; Summers and Sweeney 1998; Beneish 1999; Dechow et al. 2016).

While this study makes the first attempt to highlight the impact of managerial incentives on product and business expansion disclosures, it is important to note how such nonfinancial disclosures help managers fulfil self-serving incentives. Firstly, managers have discretion on when and how to make disclosures to the public (e.g., Richardson 2002). Nonfinancial disclosures of PBE plans, especially in respect of disclosure completeness and timeliness, are relatively hard to verify ex post, or at least in a short run, by outside investors who generally do not have access to a firm’s private information. Thus, managers can manipulate the timing and selectivity of PBE disclosures to fulfil personal trading incentives without bearing high disclosure risk. Specifically, managers can selectively release (withhold) good (bad) PBE news to boost stock prices, or selectively disclose (withhold) bad (good) PBE news to reduce stock prices. Given such incomplete disclosures of news, investors are likely to overreact to the good (bad) PBE news, leading to an inflation (deflation) of stock prices and an increase in trading gain by insiders. On the other hand, while withholding PBE news at a specific point in time, managers can defend themselves from litigation and reputational losses by arguing that they were unaware of, or knew with uncertainty, about the news at that point. As such, the potential reputational and legal costs for withholding PBE news would be relatively low for managers, even if outsiders discern the incomplete or untimely PBE disclosures. This line of reasoning is in line with the prevailing literature (e.g., Kothari et al. 2009; Baginski et al. 2018; Bao et al. 2019; He et al. 2021) which shows that managers tend to withhold various material value- relevant news. Secondly, in pricing firm equity, investors, if rational and sophisticated, should rely on nonfinancial information such as product or business expansion to forecast a firm’s long-term streams of future earnings. Consistent with this notion, Nichols (2010) finds evidence of strong market reactions to PBE disclosures; He et al. (2019) find that analysts actively utilize PBE disclosures for their earnings forecasts for firms, and that investors react more strongly to the analysts’ earnings forecasts in response to PBE disclosures. Thirdly, PBE disclosures may occur sporadically and are discretionary in their timing and selectivity, thus making the timing of disclosures relatively easy for managers to fulfil personal trading incentives. In addition, PBE plans involve rich and specific heterogenous information, including varied nature of news, from which managers can select to affect stock prices. In sum, nonfinancial disclosure of PBE plans is a powerful, flexible instrument that managers can use to fulfil their personal incentives. In the next section, I discuss how insider trading incentives impact disclosures of PBE plans.

The value of insider trades is associated with stock prices. So, to increase trading gain, managers can exploit their private information and manipulate corporate disclosures to influence stock prices. But insider trading regulations (particularly, the “disclose or abstain” doctrine) require that insiders who possess material private information should disclose it to the public, or abstain from trading ex ante based on the private information. Any insider purchase (sale) preceding price-relevant good (bad) news disclosures is regarded as illegal.⁵ The enactment of the Insider Trading Sanctions Act (ITSA) of 1984 and the Insider Trading and Securities Fraud Enforcement Act (ITSFEA) of 1988 in U.S. substantially increased penalties for illegal insider trades.⁶ Therefore, when managers plan for trading, they can instead opt to selectively disclose good (bad) news to inflate (deflate) stock prices before selling (buying) shares. However, insiders who trade after corporate disclosures may still be suspected of having exploited foreknowledge of price-relevant public disclosures (Huddart et al. 2007). This, if confirmed to be true, would contravene the U.S. securities laws governing the release of forward-looking statements that surround insider trades (Arshadi 1998; Rogers and Stocken 2005). Although insiders still bear some degree of litigation risk for trading after corporate disclosures, this risk is lower than that associated with trading before disclosures (e.g., Garfinkel 1997; Noe 1999; Huddart et al. 2007). Such litigation risk is mainly manifested in the insider selling case for two reasons.

First, when insider sales are followed by a price decline, investors who fail to trade duly would suffer losses and can thereby file a lawsuit against insiders, alleging that the insiders traded on foreknowledge of public disclosures (e.g., Huddart et al. 2007). In contrast, a stock price increase following insider purchases only leads to opportunity losses for investors. Because the opportunity costs of not purchasing shares duly are not considered as damages to investors (Niehaus and Roth 1999), presumably insider purchases after disclosures would not result in litigation to insiders. Second, most private enforcers of insider trading rules focus exclusively on insider selling cases, and the courts often recognize insider sales as an action with scienter (Johnson et al. 2007; Rogers 2008).⁷ In contrast, insider purchases, especially after bad news disclosures, could be construed alternatively as insiders signaling their optimistic beliefs in a firm’s future prospect. Thus, they are not usually recognized as a mechanism to establish evidence of scienter in courts. Insiders could defend themselves against legal and/or reputational risk associated with insider purchases by claiming that the purchases are aimed at signaling good future prospect of firms.

The Stock Exchange’s listing agreement in U.S. seeks to “ensure timely disclosure of information that may affect security values or influence investment decisions (NYSE Listed Company Manual 2019)”. Failure to disclose all PBE news on a timely basis transgresses the listing agreement, and could bring about litigation and associated reputational losses for firm management after the strategic disclosures are discovered. Nevertheless, prior research (e.g., Bradbury et al. 2009; Kim et al. 2011; Baginski et al. 2018; Bao et al. 2019; He et al. 2020) finds evidence that managers withhold bad news for extended periods. The completeness and timeliness of nonfinancial disclosures as to PBE plans are not easily discernable by outside investors. Hence, insiders may bear relatively low disclosure risk for manipulating the timing and selectivity of such nonfinancial disclosures. Such disclosure risk is even lower for selectively releasing bad news and hiding good news, than for selectively disclosing good news and concealing bad news, at a specific point in time. Given low trading risk from insider purchases and low disclosure risk from bad-news PBE disclosures, the perceived costs to insiders of purchasing shares after a bad-news PBE disclosure should be lower compared to the perceived benefits of trading gain. Therefore, managerial incentives to buy shares are expected to induce a higher incidence of a bad-news PBE disclosure in advance of the share purchases. That said, I also allow for the possibility that managers might still scruple about potential reputation losses and litigation that are associated with the opportunistic strategy and thus refrain from adopting such self- serving maneuver. Accordingly, I make my first hypothesis in both the null and alternative forms as follows.

As discussed previously, insiders may still bear high trading risk for selling shares after good news disclosures. In this scenario, if the perceived benefits of a trading gain are higher (lower) than the perceived costs associated with the trading and disclosures, insiders would (would not) selectively disclose good news on PBE before selling shares. Thus, my second hypothesis, stated respectively in the null and alternative forms, follows.

The empirical analysis is conducted based on data gathered primarily from six sources: Compustat, Center for Research in Security Prices (CRSP), Institutional Brokers Estimate System (I/B/E/S), Capital IQ, Factset, and Thomson Financial. I obtain insider trading data from the Thomson Financial Insider Research Services Historical Files. Consistent with Huddart and Ke (2007), insider trading transactions used in the empirical tests are limited to open market purchases and open market sales. Non-open-market transactions, such as dividend reinvestments, stock transfers among family members, and pension transactions, are excluded. I further restrict the insider trading transactions to those by officers and directors only, excluding those by non-officer employees who are unlikely to have an influence on corporate disclosure decisions.⁸ To focus on the aggregate influence of firm management team, I sum the purchases and sales by all managers of the same firm in the periods of interest.⁹

I draw the PBE disclosure data from Capital IQ. It maintains a team of over 600 analysts who collect and code key corporate developments from press releases and news outlets for all U.S. listed firms from the year 2002. My sample period starts in that year, and ends in 2012.¹⁰ Since the Regulation Fair Disclosure was enforced in 2001 to prohibit firms from communicating private information to institutional investors, it is unlikely for these investors to be informed enough to front-run insider trades during my sample period. It thus provides a good setting in which to examine the opportunistic insider trades and strategic PBE disclosures.

Capital IQ has data on a variety of corporate development items, including earnings guidance, product announcements, and business expansion announcements. Product and business expansion announcements pertain to stand-alone public disclosures, which, in content, are exclusive of other types of corporate reporting and disclosures; this makes my empirical analysis relatively clean and not subject to systematic confounding effects induced by other concurrent disclosures. The PBE news announcements were all initiated by firms, with each announcement corresponding to a unique news content. I collected 85,535 disclosure observations that have ticker symbols. A firm might make multiple, distinct PBE news announcements at the same date, giving rise to duplicate disclosure observations. I remove these duplicate data, reducing the sample size to 78,106 observations. I then use ticker symbols to merge these disclosure data with the I/B/E/S-CRSP link table, and the sample reduces to 64,869 observations. Further, I require that observations have the necessary data available in the databases to construct the variables of interest for my empirical tests. This procedure yields the final sample that comprises 27,016 disclosure observations for 873 unique firms.¹¹ Panels A, B, and C of Table 1 tabulate the frequency of PBE disclosures by year, quarter, and industry, respectively. Firms in the chemical products, electronic equipment, computer equipment and services industries have the highest frequency of PBE disclosures. Panel A of Table 2 gives descriptive statistics of the variables used in the main tests.

Following Noe (1999), Cheng and Lo (2006), Brockman et al. (2008), Nichols (2010), Ge and Lennox (2011), and He (2018), among others, I use the stock market reaction to identify whether a disclosure conveys good or bad news to the market. Specifically, a PBE disclosure is classified as a good (bad) news disclosure if the cumulative abnormal stock returns over the 3-day window centered on the disclosure date are positive (negative).¹² The cumulative abnormal returns are calculated based on the market model with an estimation period of [− 181, − 2] relative to the PBE disclosure date. In addition, I use an alternative estimation window, [− 181, − 2] plus [2, 52], to construct the measure for disclosure news, and obtain qualitatively identical results; this specification for the news measure, which accounts for a post-disclosure period for the estimation window, also assuages the reverse causality concern (discussed in Sect. 4.3). The dependent variable in my main empirical analyses is Gbnews, which equals 1 if a firm delivers a good news disclosure of PBE information, and equals 0 if a firm makes a bad-news PBE disclosure.¹³ The mean value of Gbnews, reported in Table 2, indicates that 50.7% of the announcements of PBE plans pertain to good news disclosures. Suijs (2007, p.391) contends that “a firm may prefer not to disclose its private information if the firm is uncertain of investor response”. Therefore, consistent with prior research (e.g., Cheng et al. 2013), my sample observations in the main tests are restricted to those that have disclosures of PBE plans.¹⁴

Insider trading opportunities motivate managers to be strategic to change their nonfinancial disclosure policies. As discussed in Sect. 2.3, if insiders wish for a high trading gain, they should trade shortly after disclosures. In such a scenario, the ex post trading intensity reflects the degree of managers’ ex ante incentives to grab trading gain. Consistent with prior studies (e.g., Sivakumar and Waymire 1994; Noe 1999; Ke et al. 2003; Cheng et al. 2013), I focus on insider trades in the 30-day period after a disclosure, not only because delayed trading after a disclosure would reduce trading profits for insiders, but also because PBE news might not be withheld for long by managers. I obtain qualitatively the same results if I expand the window to be the 90-day period post PBE disclosures. Because the variables for insider trades are highly skewed, I use the logarithm transformation of insider trades for empirical tests.

To separate sale incentives from purchase incentives for a given firm in the periods of interest, I follow the related literature (e.g., Cheng et al. 2013) and define the insider trading variables as follows. Insiderbuy equals the natural logarithm of one plus the net insider purchases (i.e., the number of shares purchased minus the number of shares sold) over a 30-day window after a PBE disclosure, if a firm has a positive amount of net insider purchases over the 30-day window, and equals 0 otherwise. Insidersell equals the natural logarithm of one plus the net insider sales (i.e., the number of shares sold minus the number of shares purchased) over a 30-day period after a PBE disclosure, should a firm have a positive amount of net insider sales over the 30-day window, and equals 0 otherwise. As shown in Table 2, the mean value of Insidersell is significantly higher than the mean Insiderbuy, indicating a higher intensity of insider sales than that of insider purchases after PBE disclosures. Figure 1a plots the frequency of insider trades (TInsiderbuy and TInsidersell) for the whole population of U.S. listed firms during my sample period. From the chart, we can see that, in general, insider sales occur far more frequently than insider purchases. Panel B of Table 2 provides additional summary statistics of insider trades for the whole population. TInsiderbuy (TInsidersell) has the mean value of 0.5245 (2.8644), which is significantly lower than the mean of Insiderbuy (Insidersell) for the disclosure sample. This suggests that the post-PBE-disclosure insider trades are more intense than the insider trades in general.

To reduce legal and reputational risks for informed insider trades, insiders might opt to make routine trades in place of non-routine trades to pretend to be non-opportunistic traders. In such a case, routine insider trades are still opportunistic (He et al. 2021). Therefore, I do not separate routine insider trades from non-routine insider trades in my main analysis. My results still hold for following Cohen et al. (2012) to restrict the measures of insider trades to non-routine trades only.

In addition, following Cheng et al. (2013), I use indicator variables to capture the existence of insider trades in the 30-day period after a PBE disclosure. The indicator variable for insider sales (DInsidersell) equals 1 if the amount of net insider sales is positive (i.e., insider sales are larger than insider purchases) and 0 otherwise. The indicator variable for insider purchases (DInsiderbuy) equals 1 if the amount of net insider purchases is positive and 0 otherwise. The use of this alternative specification of insider trading does not alter any inference drawn in the main empirical tests.¹⁵

Lastly, in the cases when corporate insiders trade their stocks in advance of releases of value-relevant news, stock returns can be used to measure the profitability of the insider trading, on the assumption that investors do not possess private information about the news and thus would not react to it prior to the disclosure of it. Nevertheless, my study concerns the insider trades taking place after strategic managerial disclosures. Specifically, insiders capitalize on investors’ overreaction to the strategic PBE disclosures to profit from their stock trading. As both corporate insiders and outside investors trade on the PBE disclosures, stock returns after the disclosures would not capture the profits of the insider trades. Hence, as with the related literature (e.g., Cheng and Lo 2006; Cheng et al. 2013), I do not examine the insiders’ trading profits in this study either.

Table 3 reports correlation matrix among the variables used in the baseline regression. The correlation between Insiderbuy (Insidersell) and GBnews is − 0.0287 (0.0471) and statistically significant at the 1% level, thus providing initial support for the hypothesis H1_a (H2_a). Panel A (Panel B) of Table 4 reports the results of the two-sample tests of mean for GBnews; the two subsamples are partitioned by whether a given observation’s amount of net insider purchases (net insider sales) is greater than 0. The mean value of GBnews for the subsample observations that have a positive amount of net insider purchases is 0.4512, significantly lower (t-stat. = 4.71) than the mean GBnews for the other subsample observations that have a zero or negative amount of net insider purchases, which is 0.5107. The mean of GBnews is significantly higher (t-stat. = 6.91) for observations that have a positive amount of net insider sales than for observations with a non-positive amount of net insider sales. Panel C compares Insiderbuy and Insidersell between the good-news disclosure sample and the bad-news disclosure sample. The mean value of Insiderbuy (Insidersell) is significantly lower (higher) for the good news sample than for the bad news sample [t-stat. =  − 4.59 (7.51)]. The mean difference in Insiderbuy (Insidersell) is − 0.1281 (0.5194), which is equivalent to 22.57% (15.26%) of its overall sample mean and is economically significant. Thus, these univariate results are all consistent with the hypotheses H1_a and H2_a.

Panel D reports a result for the mapping between PBE news and future earnings. The mean values of earnings for the future three fiscal years post the announcements of good PBE news are all significantly higher than those after the bad news announcements. This result lends support not only to the implications of PBE disclosures for long-term streams of a firm’s future earnings, but also to the return-based classification of good vis-à-vis bad news for PBE disclosures.

The baseline regression model used to test the hypotheses H1 and H2 is as follows:

The dependent variable is GBnews, an indicator variable equaling 1 (0) if a product or business expansion disclosure pertains to a good (bad) news disclosure, as defined previously. The key independent variable is Insiderbuy (Insidersell), which proxies for insiders’ purchase (sale) incentives, as defined earlier. I put Insiderbuy and Insidersell in separate regressions because, by the variable construction, they are opposite in nature.¹⁶ A logit regression is run for Model (1). If the hypothesis H1_a (H2_a) holds, the coefficient on Insiderbuy (Insidersell) should be negative (positive) and statistically significant at a conventional level.

Following Nichols (2010) and He (2018), I control for earnings surprise (Asue), book-to-market ratio (BM), firm size (Size), return on assets (ROA), institutional stock ownership (Insti), capital expenditures (CapitalEx), financial leverage (Debt), and industry-level litigation risk (Litigation). Asue is an indicator variable for whether a firm’s earnings surprise, measured by the actual earnings per share (henceforth, EPS) less the median consensus analyst forecast of EPS, is positive for the fiscal quarter. A positive earnings surprise (Asue) is expected to be associated with a higher incidence of a good news disclosure (Gbnews). Prior research (e.g., Ball and Shivakumar 2005; Francis and Martin 2010; Jayaraman and Shivakumar 2013) documents that conservative corporate reporting and disclosures curb value-destroying investment and financing activities. Therefore, firms with high institutional stock ownership (high financial leverage), subject to higher monitoring from institutional investors (creditors), are likely to be conservative in their disclosures. In a similar vein, larger firms are more mature in operating their business and hence are likely to be more conservative in their corporate disclosures. Hence, Insti, Debt, and Size should be related negatively to GBnews. Firms with good performance are likely to have more good news. Thus, ROA is expected to be positively associated with GBnews. Higher capital expenditures (CapitalEx) imply more promising investment opportunities for a firm and are thus expected to be related to a higher incidence of a good news disclosure (GBnews).

I further control for abnormal trading volume (Abtradvol) and abnormal stock returns (Qtrret) to account for the impact of potential fundamental-related events on corporate disclosures.¹⁷ The controls of Abtradvol and Qtrret also mitigate the potential correlated-omitted-variables bias induced by events that affect firm fundamentals. All the control variables are constructed for the fiscal quarter that precedes the PBE announcement quarter, and are defined in “Appendix 1.” Finally, I include industry-, year-, and quarter-fixed effects in the regression, and cluster the standard errors of the coefficients by firm to control for possible time-series correlations of residuals within firms (Petersen 2009).¹⁸

Table 5 reports the logit regression results for the effect of insider trading incentives on PBE disclosures. The coefficient for Insiderbuy is statistically significant at the 1% level with the expected negative sign. This result indicates that the incidence of a bad news disclosure on PBE information, relative to that of a good-news PBE disclosure, is positively correlated with insider purchases made shortly after the disclosure. Such evidence suggests that, when insiders plan to purchase shares from the stock market, they are more likely to disclose bad news on PBE information beforehand, which is consistent with the hypothesis H1_a.¹⁹

The coefficient on Insidersell is positive and statistically significant, consistent with the hypothesis H2_a. Thus, insiders tend to disclose good news on PBE information before selling shares. This result is in contrast with the prior research (e.g., Cheng and Lo 2006) which finds no evidence that insiders provide good news earnings forecasts before selling shares. Indeed, insider sales after a good news disclosure are subject to high legal jeopardy, because the litigation risk associated with insider sales is amplified in good-news earnings forecasts that, in themselves, entail high disclosure risk. However, the trading risk for insider sales might not be manifested in nonfinancial disclosures which entail relatively low disclosure risk. Thus, the costs associated with a good-news PBE disclosure being made before insider selling are likely to be perceived by insiders as being lower than the expected trading gain. This explains why I find a high incidence of good-news PBE disclosures made before insider sales.

Baghai et al. (2014) argue that it is more straightforward to interpret the economic significance of results based on OLS models.²⁰ Thus, I also apply OLS regression to Model (1) using a continuous variable, car, as the dependent variable. car is the cumulative abnormal stock returns surrounding a PBE disclosure and is as defined in Sect. 3.2. The regression results (not tabulated) are consistent with the hypothesis H1_a (H2_a): The coefficient on Insiderbuy (Insidersell) is negative (positive) and statistically significant (t-stat. =  − 4.08 (4.72)). A one-standard-deviation increase in Insiderbuy (Insidersell) is associated with an increase in car by 0.00071 (0.00262), which accounts for around 23.28% (85.90%) of the sample mean of car and is thus economically significant.

Overall, my baseline regression results are consistent with the notion that managers manipulate the timing and selectivity of PBE disclosures to fulfil self-serving trading incentives. Nonetheless, the baseline results are inconsistent with, and thereby rule out, the possibility that managers trade in response to plausible market under-reactions to PBE disclosures, because under this under-reaction possibility, managers should buy (sell) shares, rather than sell (buy) shares, after the good (bad) news disclosures.

The idea behind the hypothesis tests is to investigate how insider trading incentives shape nonfinancial disclosure strategies. The causality flow runs from trading motives to disclosures, where the former is proxied empirically by insider trades after PBE disclosures. But there might be two main sources of endogeneity in my research context. The first is that both PBE disclosures and insider trades are simultaneously driven by some unobserved firm characteristic(s). The second type of potential endogeneity is reverse causality. In particular, more insider purchases (sales) occurring after a bad (good) news PBE disclosure can indicate either one or both of the following: (1) Managers’ incentives to buy (sell) shares motivate a bad (good) news disclosure, as hypothesized in H1_a (H2_a); (2) When stock price decreases (increases) after a bad (good) news disclosure, managers buy (sell) shares in response to the decreased (increased) stock price. In the latter case, the insider trading can be regarded as a passive response to disclosure choices. As such, reverse causality arises in a way that disclosures induce insider trades.

Because strategic PBE disclosures and insider trading are made in conjunction within a short window, identification of exogenous shocks (especially enforcements of insider trading regulations that apply simultaneously and equally to insider purchases and insider sales) to conduct a natural experiment is unlikely to address the endogeneity issues in my setting, and thus I seek other approaches. To address the first type of endogeneity, I use a firm-fixed-effects model and a reduced- form difference-in-differences specification. To cope with both the first and second types of endogeneity, I follow Cheng and Lo (2006) and Cheng et al. (2013) to employ a two-stage instrumental-variables regression technique, and importantly, also conduct a falsification test. The remainder of this section discusses each of the foregoing approaches and the associated results.

Prior research (e.g., Holthausen and Verrecchia 1988; Teoh and Wong 1993; Veenman 2002; Khan and Lu 2012; Lu et al. 2014) shows that value-relevant corporate news triggers greater revisions of investors’ prior expectations about firm fundamentals when prior information is more uncertain and prior beliefs are more divergent. While PBE disclosures may reduce outsiders’ uncertainty about corporate PBE activities (He et al. 2019), the extent to which the perceived uncertainty is resolved will be larger when investors’ prior expectations are more uncertain and more diversified. In such a case, the expected price impact of PBE disclosures will be greater, and investors are likely to overreact to the strategic PBE disclosures to a larger extent. Accordingly, I predict that the strategic PBE disclosures made for opportunistic trades will be more pronounced. To test this, I use analyst forecast dispersion as the proxy for the expected price influence of PBE disclosures. The higher the analyst forecast dispersion, the greater the investors’ uncertainty and/or divergence about firm fundamentals (Barron 1995; Bamber et al. 1997; Hong and Stein 2007), and thus the larger the expected price impact of PBE disclosures, as per prior studies (e.g., Veenman 2002; Khan and Lu 2012). I partition my sample into two subsamples based on the sample median of analyst forecast dispersion (namely, Disper, which is defined in “Appendix 1”), and estimate Model (1) separately for these two subsamples.

Table 10 reports the results. The coefficient on Insiderbuy is negative and statistically significant at the 1% level for the high-Disper subsample but is not statistically significant for the low-Disper subsample. The coefficient on Insidersell is incrementally more positive in the high-Disper subsample relative to the low-Disper subsample; the difference in the coefficients for Insidersell amounts to 0.0164 and is statistically significant (p = 0.0272). These results suggest that managers have stronger incentives to make strategic trades and PBE disclosures if ex ante, they feel that their disclosures would trigger stronger stock price reactions. Such evidence and inference reinforce my earlier view that managers are able to foresee the market reactions to PBE disclosures.

To explore insiders’ ability to pursue the strategic disclosure and trading behavior, I also examine whether my main results vary with CEO power. As with Ke et al. (2019), I define CEO power as strong, if (1) the CEO is the chairman of the board and (2) the proportion of independent directors on the board is below the sample median. When a CEO is more powerful, s/he will be more able to make, and convince other managers of, the PBE disclosure decisions beneficial to trading incentives. Thus, such strategic behavior would likely be more pronounced. To test this conjecture, I divide my sample into two subsamples based on whether CEO power is strong, and then run Model (1) on both subsamples. The data used for constructing the CEO power variable are obtained from the Institutional Shareholders Services (ISS) database which covers S&P 1500 firms for the period starting from the year 2007.

Table 11 reports the results. The coefficient on Insiderbuy is negative and highly significant in the high-CEO-power subsample but is only marginally significant in the low-CEO-power subsample. Insidersell takes on a significantly positive coefficient in the high-CEO-power subsample, but a statistically nonsignificant coefficient in the low-CEO-power subsample. Together, these results suggest that managers’ strategic PBE disclosures made for purpose of insider trades are more evident for firms with strong CEO power.

Due to the availability of PBE disclosure data, my sample period starts from the year 2002, i.e., after the enforcement of Regulation Fair Disclosure in 2001. Under this regulation, firms are prohibited from conveying private information to institutional investors. Therefore, it is interesting to see whether, in the absence of access to management’s private information, institutional investors are still able to effectively monitor insiders in a way that curbs opportunistic insider trades and avoids losses from such trades. To offer insights into this issue, I test whether my main results vary across firms with high vis-à-vis low institutional stock holdings. To this end, I split my sample into two subsamples based on the sample median of institutional ownership (Insti), and then estimate Model (1) separately for the two subsamples.

Table 12 reports the results. While the coefficients for Insiderbuy (Insidersell) are negative and statistically significant at the 1% level for both the high-Insti subsample and the low-Insti subsample, the difference in the coefficients on Insiderbuy (Insidersell), which amounts to 0.0151 (0.0011), is statistically nonsignificant [p = 0.1826 (0.8323)]. This suggests that the strategic PBE disclosures made for opportunistic insider trades are equally likely for firms with high institutional stock ownership vis-à-vis those with low institutional ownership. This result is inconsistent with the proposition that institutional investors play an effective monitoring role in curbing opportunistic insider trades and strategic disclosures. However, it is consistent with the notion that insiders’ exploitation of disclosure opportunities for personal trading gain is at the expense of investors that are inclusive of institutional investors. Such an inference reconciles with some prior research (Glosten and Milgrom 1985; Kyle 1985; Sias and Whidbee 2010; Indjejikian et al. 2014) which argues that opportunistic insider trades are to the detriment of investors including sophisticated, yet uninformed, investors.

The hypothesis tests are conditioned on management’s decisions to voluntarily disclose PBE information. This might give rise to sample selection bias because observations that have insider trades but do not have a PBE disclosure are omitted from the regression analyses. The intensity of insider trades might differ between the PBE-disclosure sample and non-PBE-disclosure sample. To account for this possibility, I further control for industry-level insider trades (Industry_Insiderbuy (Industry_Insidersell)), which are measured by the average of net insider purchases (sales) made by the same- industry corporate insiders during the 30-day period after a PBE disclosure. Such a control serves to mitigate the impact of the potential sample selection bias on the coefficient estimates for Insiderbuy and Insidersell. Results are reported in Panel A of Table 13 and remain qualitatively the same in support of the hypotheses H1_a and H2_a.

I also employ a two-stage Heckman probit regression model to tackle the potential sample selectivity problem. The first-stage regression is modeled by a probit regression of management’s decision to voluntarily make a PBE disclosure, where the dependent variable is the incidence of a PBE disclosure (namely, Inci, which is defined in “Appendix 1”). The Inverse Mills ratio estimated from the first-stage regression is then included in the second-stage regression, which is modeled by Model (1), to control for the selectivity bias. An effective Heckman (1979) model requires that exclusionary variables be validly included (excluded) in (from) the first (second) stage regression (Little 1985; Puhani 2000; Lennox et al. 2012). In my case, the valid exclusionary variables should be related to the incidence of a PBE disclosure (Inci), but unrelated to the disclosure news (GBnews) except indirectly through Inci.

I identify four exclusionary variables, namely, earnings volatility (EarningsVol) and three distinct proxies for proprietary costs of disclosures (EntryCo, Mktsize, and Substi per Karuna (2007)), all of which are defined in “Appendix 1”. High business risk for a firm, featured by high earnings volatility, would trigger outsiders’ greater demand for disclosure transparency. Therefore, EarningsVol is expected to be positively associated with the incidence of a PBE disclosure (Inci). However, EarningsVol should have little direct association with the likelihood of a good-news PBE disclosure relative to that of a bad-news PBE disclosure (GBnews), and hence could be validly excluded from the second-stage regression.

Disclosures of PBE information increase a firm’s risk of leaking its relevant proprietary information to product market competitors. A firm that has lower product substitutability (lower Substi), lower entry costs (lower EntryCo), or larger market size of competing products (higher Mktsize) faces more intense industry-level product market competition and thus is subject to higher proprietary costs of disclosures (Karuna 2007). Accordingly, the incidence of a PBE disclosure (Inci) should be negatively associated with Mktsize and positively associated with Substi and Entryco. However, given the decision to disclose PBE information, the proprietary costs of disclosures should not have further direct impact on the disclosure news (GBnews).²² Hence, EntryCo, Mktsize, and Substi should also be the valid exclusionary variables for the Heckman model.

In the first-stage probit regression, Inci is regressed on the four exclusionary variables and the same set of independent variables as included in Model (1). Panel B of Table 13 reports the regression results. EntryCo and Mktsize are statistically significant in the expected sign. The coefficients on Insidersell and Insiderbuy are qualitatively identical to those reported in Table 5. It is possible that there is no systematic difference in firm characteristics across the disclosure sample and nondisclosure sample. In this case, selectivity will not bias the coefficient estimates. Should the exclusionary variables be valid for the Heckman estimation procedure, rho equaling 0 would indicate that there exists no sample selection bias. As reported in the table, rho is not statistically different from 0. This thus serves as another piece of evidence that my main results are immune from the potential sample selection bias. In addition, I conduct a multinomial logit regression for Model (1) using the full sample, whereby the potential selection bias would be corrected (Bourguignon et al. 2007). The inferences for the hypotheses H1 and H2 remain unchanged for applying the multinomial logit specification.

If some PBE disclosures are bundled contemporaneously with management earnings forecasts, the return-based measures of PBE disclosure news may introduce bias into my results. I deal with this issue as per, Ball et al. (2012) and He (2018), for example. Specifically, I regress GBnews on earnings forecast news, which is defined as management forecast of EPS minus the median consensus analyst forecast of EPS that is issued within 90 days prior to the management forecast date, for those “bundled” PBE disclosure observations. I then treat the residual as the market reaction to the PBE disclosure news only. Should the residual be positive (negative), the PBE disclosure is classified as a good (bad) news disclosure. For the PBE disclosures that coincide with the announcements of actual earnings, I apply a similar procedure to separate the PBE-related news from the earnings- announcement news. In particular, I regress GBnews on earnings surprise, which is defined as the actual EPS minus the median consensus analyst forecast of EPS that is issued within 90 days prior to the EPS announcement date, to obtain the residual for the disclosure news classification. My inferences for the hypotheses H1 and H2 remain unchanged when using these alternative news measures for PBE disclosures. The bundled PBE announcements are defined as those occurring within one day surrounding management earnings forecast dates or earnings announcement dates. There are 444 (1577) observations that are bundled with management earnings forecasts (earnings announcements), accounting for only 1.83% (5.84%) of my disclosure sample, for the period 2002–2011 (2002–2012).²³ I also obtain the same inferences for the hypotheses H1 and H2 if I tease out all the bundled observations for the hypothesis tests.