So Sue Me!

The Value Implication of Patent Litigation‡

Fred Bereskin*, Po-Hsuan Hsu**, William Latham***, Huijun Wang†

July 23, 2018

Abstract: Using comprehensive patent lawsuit data from 2000 to 2014, we find that firms with

litigated patents and their adversaries experience positive stock returns ranging between 0.25%

and 1.18% in the 10 or 24 days following litigation announcements. In addition, portfolios

consisting of firms involved in patent lawsuits generate risk-adjusted alphas between 0.56% to

0.75% per month in the following year. Further analysis on firms’ future profitability and their

competition suggests that our finding of positive, yet undervalued, effects of patent litigation on

stock prices could be explained by investors’ underreaction.

Keywords: Patent litigation; stock returns; information delay; underpricing

JEL classification: G11, G14, O34

‡ We thank Dan Bereskin, Jim Bessen, Lauren Cohen, Edwin Lai, Haitian Lu, Don MacOdrum, Michael Meurer,

Zhenjiang Qin, Ghon Rhee, Ryan Whalen, and Angela Zhang, as well as seminar participants at the University of

Hong Kong (School of Law), University of Massachusetts Lowell, and the Taiwan Finance Association Annual

Meeting for their valuable comments. We also thank Brian Howard from Lex Machina for preparing and organizing

data for us.

* Trulaske College of Business, University of Missouri, Columbia MO 65211. Email: bereskinf@missouri.edu.

** Faculty of Business and Economics, University of Hong Kong, Pokfulam Road, Hong Kong. Email:

paulhsu@hku.hk. Phone: +852-2859-1049.

*** Lerner College of Business & Economics, University of Delaware, Newark, DE 19716.

Email: latham@udel.edu. Phone: (302) 831-6846.

† Lerner College of Business & Economics, University of Delaware, Newark, DE 19716. Email: wangh@udel.edu.

Phone: (302) 831-7087.

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1. Introduction

Patent litigation is an important issue for firms, managers, and shareholders in today’s

business environment where intellectual property plays a critical role. Compared to the 1980s and

earlier, firms are now more likely to be involved in patent litigation, either as plaintiffs or

defendants (Bessen and Meurer, 2013; United States Government Accountability Office, 2013;

Cohen, Gurun, and Kominers, 2016a,b, 2018). It is commonly perceived that the more litigious

environment has increased firms’ risk of being exposed to significant direct and indirect litigation

costs, which may retard their exploitation of intellectual property and reduce their long-term

growth.1 Recent research in patent litigation focuses on both the role and impact of non-practicing

entities (NPEs, sometimes referred to as “patent trolls”) in today’s legal environment (Bessen,

Ford, and Meurer, 2011; Cohen, Gurun, and Kominers, 2016b, 2018), as lawsuits initiated by NPEs

have increased faster than those initiated by practicing entities (PEs) in recent decades. Unlike

NPEs, which mainly target cash-rich firms and value out- of-court settlements, PEs are generally

more motivated by product competition (Cohen, Gurun, and Kominers, 2018). Moreover, while

NPEs primarily aim to share the profits of targeted defendants, PEs seek to eliminate (potential)

competitors, regardless of their profits. In this paper, we focus on stock market reactions to the

announcements of patent lawsuits that do not involve NPEs; these events offer asset pricing

implications of patent-related industry competition.

We first collect patent lawsuits involving public firms from the Lex Machina database,

which covers patent litigation cases filed since 2000. The Lex Machina database is regarded as the

most comprehensive database of U.S. patent lawsuits and has been used in many recent studies

(Akcigit, Celik, and Greenwood, 2016; Allison, Lemley, and Schwartz, 2015, 2018; Cohen, Gurun,

and Kominers, 2016b, 2018). We then combine the patent litigation data with the CRSP/Compustat

database to examine the stock market’s daily and monthly reactions to announcements that firms

1 For example, the American Intellectual Property Law Association (2015) reports that patent litigation tends to be

costly and time-intensive. Specifically, the median patent infringement suit experiences litigation costs ranging from

$100,000 (when less than $1 million is at risk) to $5 million (when more than $25 million is at risk); additionally,

amounts at risk are associated with greater hours required to litigate. Aside from these direct costs, we also note

potential indirect costs including reduced pledgeability (Chava, Nanda, and Xiao, 2017; Mann, 2017) or the reduced

selling price of intellectual property (Lev, 2001). Akcigit, Celik, and Greenwood (2016) note that 20% of domestic

patents are sold. Other indirect costs include distractions to management, difficulties in ensuring long-run

commitments with suppliers and customers (Tucker, 2013), difficulties in attracting financing (Feldman, 2014), and

delays in implementing innovation and marketing strategies.

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are litigants (either plaintiffs or defendants) in patent lawsuits between 2000 and 2014. We find

that patent litigation leads to positive announcement returns, both for plaintiffs and defendants. On

announcement of patent litigation, litigants experience cumulative abnormal returns (CAR)

ranging from 25 to 118 basis points in various short-term announcement windows, various models,

and different samples.2 For example, plaintiffs provide average returns in excess of the CRSP

value-weighted market index of 52, 114, 55, and 118 basis points in the [-1, 10], [-1, 24], [-3, 10],

and [-3, 24] windows, respectively, around litigation announcement dates (denoted by 0).

Defendants provide average returns in excess of the CRSP value-weighted market index of 33,

109, 34, and 110 basis points in the [-1, 10], [-1, 24], [-3, 10], and [-3, 24] windows, respectively.

Although this phenomenon is slightly larger for plaintiffs, it is significantly positive for both

plaintiffs and defendants. This finding might appear puzzling at first glance, as it challenges the

general impression that patent litigation harms firm value. More interestingly, we find significantly

positive correlations in the cumulative abnormal returns of plaintiff-defendant pairs, which

challenges the belief that patent litigation is a zero-sum game for plaintiffs and defendants.

Our investigation based on comprehensive post-2000 patent litigation cases provides up-

to-date empirical evidence and offers new insights with respect to the effects of patent litigation

on stock prices, which differs from prior event studies based on pre-2000 press data that show

negative announcement effects of approximately 2-3% of firm value (e.g., Bhagat, Brickley, and

Coles, 1994; Lerner, 1995; Bhagat, Bizjak, and Coles, 1998). Moreover, using a sample of patent

lawsuits from 1984-1999 with a 25-day window around the filing dates of cases, Bessen and

Meurer (2012) estimate average losses of 0.5% for defendants.

The following example illustrates how patent litigation can enhance stock prices for

defendants. In April 2008, Seagate sued STEC for patent-infringement; the suit pertained to patents

relating to solid-state drives (SSDs). Although one might expect litigation to be value-destroying

in this case, the lawsuit provided an opportunity for STEC to establish the value and strength of

2 We consider the following four short-term announcement windows: [-1, 10], [-1, 24], [-3, 10], and [-3, 24], around

litigation announcement dates (denoted by 0). We also consider the following three CAR types: CAR1 is defined as

the cumulative daily returns in excess of the CRSP value-weighted market index in the period; CAR2 is defined using

the CAPM model with market beta estimated during the window of trading days -171 through -22 with a minimum of

100 non-missing observations (as in Bhagat, Brickley, and Coles, 1994); and CAR3 replaces the CAPM model in

CAR2 with the Fama-French four-factor model (including the market factor, SMB factor, HML factor, and MOM

factor). In addition, we consider two samples: the first one includes all patent litigation cases, and the second one

includes patent litigation cases with both a public plaintiff and a public defendant.

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its SSD patents. In particular, its CEO indicated that “STEC will take appropriate action to protect

its interests, including seeking the invalidation of Seagate’s patents” (Cheung, 2008). Consistent

with the positive effects of the litigation, the [-1, 1], and [-1, 10] cumulative abnormal returns for

STEC around the lawsuit filing were 4.04% and 27.6%, respectively. The lawsuit was dropped in

the following year, with no licensing agreement or financial exchange. Around this time, STEC’s

CEO expressed how the events around this agreement helped establish the validity of its SSD

patents: “We have always contended that SSD does not borrow from existing hard-drive

technology but rather offers an all-together new approach to storage. We view the dismissal as a

vindication of our technology” (Mearian, 2009).

It is also worth noting that a plaintiff in patent litigation is not necessarily the party that

challenges others’ infringements. In cases of declaratory judgments (DJ), the alleged infringer is

the plaintiff, and the patent owner claiming to own patent rights is the defendant. When claiming

for a declaratory judgment, the plaintiff asks the courts to provide clarity (e.g., ruling that there is

no infringement, ruling that the patent owner’s claim is invalid or unenforceable). Thus, we study

both parties—plaintiffs and defendants—due to the unique role that declaratory judgments play in

patent litigation.3

To further understand the value implication of patent litigation, we create a monthly-

rebalanced stock portfolio consisting of firms experiencing patent litigation in the past 12 months

(“litigation portfolio” henceforth), and find that this portfolio generates excess returns as high as

1.02% per month with a t-statistic of 2.1.4 We then consider various risk factor models and find

that the litigation portfolio generates alphas (i.e., risk-adjusted returns) of 56-75 basis points per

month (with t-statistics above 3.0) after we control for relevant risk factors (e.g., size, value,

momentum, profitability, investment factors), for factors related to R&D and patents, and for

mispricing factors. When we focus on the portfolio consisting of defendants only, we obtain

3 9.9% of our cases are declaratory judgments, and our results are robust to excluding these cases from our analysis. 4 This approach is often known as a “calendar-time portfolio” and differs from an “event-time portfolio.” We do so

because stocks typically are involved in multiple patent litigations in the same period. The latter approach treats each

stock involved in each litigation case as an event, and forms a portfolio based on all events. Thus, statistical inferences

based on event-time portfolios will be subject to issues including cross-correlation of event returns and overweighing

stocks with multiple cases.

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consistent results. Therefore, the significantly positive returns on the litigation portfolio cannot be

attributed to conventional risks and factors.

It is also possible that our findings are driven by particular industries. That is, certain fast-

growing, high-tech industries might experience both high stock returns and more patent lawsuits.

To examine this possibility, we implement the following matched-samples analysis: for every

treated firm involved in patent litigation, we find a matched firm that is in the same industry and

has similar firm characteristics, but has not been involved in patent litigation in the prior year; we

then form a control portfolio consisting of those matched firms, so we may examine the difference

in the returns generated by both the litigation portfolio and our control portfolio. The returns of

the litigation portfolio consistently outperform those of the control portfolio, suggesting that our

baseline finding cannot be attributed to industry-specific causes.

To further examine if the predictive power of patent litigation on future stock returns is

distinct from that of existing return predictors, we implement monthly Fama-MacBeth regressions

for all CRSP/Compustat firms by controlling for various firm characteristics, including size, book-

to-market ratio, lagged stock returns, momentum, R&D expenditure, patents, operating profit, cash

flow, financial constraints, leverage, profit margin, and industry concentration, as well as industry

fixed effects. We find a significantly positive coefficient of 0.2%-0.3% on the indicator variable

associated with a firm that experiences at least one patent lawsuit over the past 12 months. This

result, which suggests that the litigation-return relation we document is different from other

documented patterns with respect to return predictability, calls for a deeper analysis.

We propose four possible explanations for the intriguing pattern that firms involved in

patent litigation experience significantly positive stock returns. First, these firms share a common

exposure to an unknown or unspecified systematic risk associated with patent competition and/or

litigation, and thus provide higher expected stock returns as compensation for bearing higher risk

exposure. Second, systematic risk related to financial constraints (Li, 2011) could exist, such that

more financially constrained litigants are then associated with higher expected returns. Third, firms

that are rich in cash or appear promising in future cash flows are more likely to be sued; thus, our

findings simply capture the pattern that cash-rich litigants provide higher subsequent stock returns.

Lastly, these firms may, on average, benefit from being involved in patent litigation, and this

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benefit may be consistently undervalued by the stock market due to the complexity of patent

lawsuits, investors’ limited attention or ambiguity aversion to litigation, and/or systematic

underpricing.5

Our evidence is not consistent with our first explanation (i.e., firms share a common

exposure to an unknown or unspecified systematic risk associated with patent competition and/or

litigation). Whereas a risk-based explanation suggests a price drop following the announcement of

patent litigation (as investors discover new risk exposure and start to discount future cash flows to

a greater degree), we find positive abnormal stock returns in the [-1, 1] and [-3, 3] windows around

the announcements. Moreover, if the risk-based explanation holds, then the return on the litigation

portfolio is driven by heterogeneous exposure to the new risk, and can thus serve as a mimicking

factor for the risk. We implement Fama-MacBeth two-pass regressions to test if the return on our

litigation portfolio is priced in the returns of individual stocks (see Cochrane, 2001). However, we

do not find that the sensitivity of individual stocks’ returns to the litigation portfolio return (i.e.,

the beta to the hypothetical litigation risk) is associated with significantly positive slopes in the

cross-section (i.e., the price of risk for the hypothetical litigation risk).

Further tests show that our baseline finding cannot be explained by systematic risk related

to financial constraints or by litigants pursuing cash-rich firms. We do not find a stronger litigation-

return relation in financially constrained firms that are more subject to systematic risk.6 Nor are

our results concentrated in cash-rich firms, suggesting that firms’ cash balances are unlikely to be

driving our results.

5 The literature shows that investors consistently undervalue R&D-intensive firms owing to concerns about technical

uncertainty associated with R&D activities, which leads to underpricing and return predictability (see, e.g., Lev and

Sougiannis 1996; Aboody and Lev, 2000; Chan, Lakonishok, and Sougiannis 2001; Lev, Sarath, and Sougiannis 2005).

Eberhart, Maxwell, and Siddique (2004) present evidence consistent with investor underreaction to R&D increases.

Cohen, Diether, and Malloy (2013) and Hirshleifer, Hsu, and Li (2013) both argue that investors have limited

processing power for complex innovations in patents and thus undervalue innovative efficient firms. Moreover,

theoretical models show that investors are more skeptical of investment opportunities when they perceive greater

uncertainty (see, Dow and Werlang 1992; Chen and Epstein 2002; Cao, Wang, and Zhang 2005; and Bossaerts et al.

2010). Moreover, the psychology literature finds that individuals tend to interpret signals more skeptically and with

greater risk when they have less processing fluency with those signals (e.g., Alter and Oppenheimer 2006; Song and

Schwarz 2008, 2009). 6 Mezzanotti (2017) finds that patent litigation’s reduction in a firm’s innovation patterns is driven by exacerbated

financial constraints.

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We then turn to our fourth explanation and investigate the potential benefits of patent

lawsuits for both defendants and plaintiffs. First, for both parties, an allegation of infringement

signals the value of patents and provides the public with a gauge of the potential value at stake.

For example, more valuable patents tend to experience more litigation and attract additional

citations following litigation (Lanjouw and Schankerman, 2001). Indeed, previous evidence shows

that the number of patent citations received is associated with greater patent value (Harhoff, Narin,

Sherer, and Vopel, 1999; Sampat and Ziedonis, 2005; Kogan et al., 2017); as a result, the publicity

associated with patent litigation delivers more information to investors (albeit with a potential

delayed reaction). It is difficult for investors to effectively value or ascertain the validity of patents,

and the litigation process helps clarify patent value (Marco and Vishnubhakat, 2013).7 Using

patent lawsuits, investors obtain more information about litigated and other associated patents and

assess their value more accurately. For example, Kiebzak, Rafert, and Tucker (2016) note that

venture capital investment typically increases with patent litigation (before eventually declining),

consistent with the net positive benefit of patent litigation in certain situations.

Moreover, patent lawsuits attract media attention and thus highlight barriers to entry by

discouraging future competitors, such as smaller firms with less extensive patent portfolios (Choi,

1998; Shapiro, 2000; Hall and Ziedonis, 2001; Bessen and Meurer, 2013; Cohen, Gurun, and

Kominers, 2018); indeed, prior research (e.g., Cohen, Gurun, and Kominers, 2016a,b, 2018;

Caskurlu, 2017) has highlighted how patent litigation changes firms’ behaviors. The filing of

patent litigation also informs investors of corporate managers’ determination to enforce intellectual

property rights against infringement or to confront groundless charges, which benefits

shareholders (Agarwal, Ganco, and Ziedonis, 2009). For example, in writing about Eastman

Kodak’s efforts to protect and monetize its patent portfolio, Mattioli (2010) notes that “aggressive

litigation has become an increasingly important part of Kodak’s corporate strategy.” Consequently,

there are certain positive aspects to litigation exposure that can potentially increase firm value,

7 The litigation process helps clarify (over the course of the legal proceedings) a firm’s patent enforcement ability and

rights. Litigation can thus help initially signal patent value and resolve associated uncertainties, albeit with a delay.

Related to this point, Marco and Vishnubhakat (2013) find that stock market reactions associated with the resolution

of patent uncertainty are comparable to those of initial patent grants: each are approximately 1.0%-1.5% excess returns.

Their findings imply that the uncertainty of patent validity is economically important, and can affect litigation behavior.

Moreover, Graham and Vishnubhakat (2013) note that the effects of legal uncertainty regarding patent validity are

particularly severe in emerging technologies and in areas with rapid growth in patenting activities.

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even for defendants.8 We recognize, of course, that patent infringement does not always result in

litigation; in particular (especially regarding how our results relate to defendants’ returns), an

infringing defendant might find it optimal to settle prior to a lawsuit. This observation helps explain

why there are positive returns associated with defendants who choose to defend themselves against

alleged infringements.9

To empirically test our explanation of the market’s underreaction to the potential benefits

of patent litigation, we implement further analyses on the subsequent profitability, competition,

and information delay (i.e., the price of a stock is adjusted more slowly). First, we find that firms

experiencing patent litigation tend to be more profitable in the long run: firms involved in patent

litigation are associated with a 3.0% increase in average return on equity (ROE) over the

subsequent five years. This finding indicates that the benefits associated with patent lawsuits, on

average, exceed the associated costs. Similarly, and reflecting increased barriers to entry associated

with patent litigation, we find that litigants experience lower industry competition (measured by

the Similarity and HHI of Hoberg and Phillips (2016)) following litigation.

Consistent with our findings being driven by investors’ delayed reactions to the benefits

from litigation announcements, we find a stronger litigation-return relation among the types of

firms with greater information delay. Specifically, we provide double-sorts using both stock return

R2 (Hou, Peng, and Xiong, 2013) and Hou and Moskowitz’s (2005) price-delay measure. Across

all models, we find that abnormal returns are significantly larger among firms with greater

information delay (lower stock return R2 or higher price delay): specifically, firms with greater

information delay experience monthly excess returns and alphas that are 65 to 107 basis points

higher than those of firms with less information delay.

Consequently, to the extent that it takes time for investors to fully appreciate patent-related

news, our findings offer new evidence on patent litigation to the asset pricing literature. In

8 To the extent that the litigation will result in a trial (as opposed to a settlement), the market values the ability to better

evaluate management with information revealed over the course of a trial and its judgment. For example, Haslem

(2005) notes that litigation resolution by court decisions dominates out-of-court settlements (from shareholders’

perspectives), and attributes his results to the valuable information associated with court judgments. 9 In evaluating patent disputes, it is important to be aware of their endogenous nature (Bessen and Meurer, 2006; 2013).

For example, Bessen and Meurer (2013) note that “the rate of lawsuit filing depends as much on the frequency of

disputes as the frequency of bargaining breakdown.” Also, Lemley, Richardson, and Oliver (2017) use survey data to

show that about one-third of alleged infringement results in litigation. This indicates that litigation is not an infrequent

result of negotiations between litigants to address potential infringement.

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particular, we demonstrate the presence of the market’s delay in reacting to benefits associated

with patent litigation; indeed, litigation is an important event that helps the market understand or

re-evaluate the value of a firm’s patent portfolio.10 We also provide intriguing evidence and new

insights into the effects of patent litigation on shareholder value. Although prior studies highlight

the pernicious effects of patent litigation, we show that the net effect of patent litigation is positive:

by increasing the market value (or simply the awareness) of firms’ intellectual property and by

deterring potential competition, litigation’s net effect is positive, despite the direct and indirect

costs of addressing a lawsuit. Our evidence does not negate the significant direct and indirect costs

associated with patent litigation, as discussed in Footnote 1. Rather, we present evidence showing

that the effect of litigation is positive, despite these considerable costs. Our investigation based on

post-2000 cases based on court records provides up-to-date empirical evidence and offers new

insights on patent litigation’s effect on firm value, which is different from prior event studies based

on pre-2000 data that are based on news articles rather than court records (e.g., Bhagat, Brickley,

and Coles, 1994; Lerner, 1995; Bhagat, Bizjak, and Coles, 1998). More recent papers have

suggested a smaller magnitude in value loss associated with patent litigation. 11

Other recent studies have also explored how the patent system affects firms. Many recent

studies have focused on the negatives associated with the current patent system, especially

associated litigation costs and effects on innovation (Heller and Eisenberg, 1998; Jaffe and Lerner,

2004; Lemley and Shapiro, 2007; Ewing, 2011; Cohen, Gurun, and Kominers, 2018). Indeed,

citing a recent letter to Congress by 51 prominent economics and legal scholars in innovation and

intellectual property law, Asay et al. (2015) assert the harmful effects of patent litigation not only

on innovation, but also on R&D and venture capital investment as well.12 By examining the

10 Our study is related to the broader literature on mispriced intangibles. Aside from the innovation and R&D literature,

other examples of mispriced intangibles include corporate governance (Gompers, Ishii, and Metrick, 2003), executive

perks (Yermack, 2006), social norms (Hong and Kacperczyk, 2009), and employee satisfaction (Edmans, 2011). 11 For example, Bessen, Meurer, and Ford (2011) estimate average losses of 0.37% associated with lawsuits by non-

practicing entities. Similarly, using a sample of lawsuits from 1984-1999 with a 25-day window around the filing

dates of cases, Bessen and Meurer (2012) estimate average losses of 0.5% for defendants. In contrast, our sample

begins in 2000. Additionally, Feldman (2014) describes the extent to which the increased rate of patent assertion

claims is very different from earlier periods: previously, litigation served as a last resort and was harmful to both firms

(since a defendant would use its patent portfolio to retaliate against the plaintiff). 12 In contrast, Farre-Mensa, Hegde, and Ljungqvist (2017) present causal evidence that patent approvals are associated

with start-up success. They note that patents help start-up firms address information asymmetries in four ways: (1) by

providing evidence that a firm can monetize its invention; (2) by increasing a firm’s ability to disclose details of its

invention; (3) by allowing those details to be conveyed with improved certification; and (4) by helping firms to signal

their quality. We recognize that the effects and determinants of patent litigation depend on the unique circumstances

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frequency of litigation, one of the most controversial aspects of the current patent system, we use

this paper to offer new insights into the appropriateness of future reforms. Recognizing that policy

makers may wish to consider reforming the patent approval system in ways that address frequent

litigation, we argue that our study’s findings suggest that lawsuits may indeed have a positive

effect with respect to promoting intellectual property protection; in turn, this positive effect

benefits firm value.

We organize our paper as follows. In Section 2, we present our sample construction and

summary statistics. In Section 3, we describe our main results for stock market reactions to patent

litigation. In Section 4, we examine possible explanations for our empirical results. We conclude

this paper with Section 5.

2. Sample and summary statistics

To construct our data, we collect patent lawsuits related to public firms’ patents by

combining the patent database of public firms (available until 2014) and the Lex Machina database

for patent litigations (available since 2000). We begin by constructing a list of 1,609,059 patents

that were granted to public firms from 1983 to 2014 by combining the NBER patent dataset

(originally developed by Hall, Jaffe, and Trajtenberg (2001)), the patent dataset of Kogan et al.

(2017), and the Google Patent database. The updated NBER patent dataset contains the patent

number, application date, grant date, and Compustat firm identifier GVKEY of the patent assignee

(i.e., the firm that owns the patent) of all utility patents granted to public firms from 1976 to 2006.

Kogan et al.’s patent database includes the patent number, application date, grant date, and Center

for Research in Security Prices (CRSP) firm identifier (PERMNO) of the patent assignee of each

utility patent granted to public firms in 1926–2010. Lastly, we use the Google Patent database to

extend the patent data to all patents granted by 2014.13

of the litigants. For example, Chien (2009) describes litigation by two large firms as a “sport of kings,” litigation by

an independent inventor against a large firm as “David v. Goliath,” and litigation to exploit a defendant’s financial

distress as “patent predation.” 13 To find the GVKEY and/or PERMNO for assignees that own patents granted in 2011–2014, we first collect the

company names and locations of the patent assignees that are public firms and that receive at least one patent in the

period 1976-2010. The company names are from the CRSP and Compustat databases. We then develop a matching

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We start our patent list with the 1983 grant year because patents were valid for 17 years

from the grant date at that time, and the Lex Machina database is available from 2000 (since June

8, 1995, patents are valid for 20 years from their application date). Then, we use this list of

1,609,059 patents in our search in the Lex Machina database, which includes patent litigations

announced since 2000. In establishing its database, Lex Machina cleaned and verified daily

updates from the United States federal court system, all United States district courts, the United

States International Trade Commission database (EDIS), and the USPTO. Lex Machina is

regarded as the most comprehensive database of U.S. patent litigation since 2000. Other recent

studies have used data from Lex Machina (Akcigit, Celik, and Greenwood, 2016; Allison, Lemley,

and Schwartz, 2015, 2018; Cohen, Gurun, and Kominers, 2016b, 2018). We derive a list of 9,343

patent litigations (“Case_IDs”) in which the involved patents were granted to at least one public

firm, which includes those without court decisions or out-of- court settlements.

When evaluating the effects of patent litigation, it is important to recognize the differences

between practicing entities (PEs) and non-practicing entities (NPEs, or “patent trolls”).14 Evidence

suggests that PEs’ patent litigation has not increased at the same rate as that of NPEs. A great deal

of litigation with PEs involves inadvertent infringement (Bessen and Meurer, 2013) and, more

generally, is driven by the actual infringement itself, as opposed to being driven by exploiting the

profitability of potential lawsuits (Cohen, Gurun and Kominers, 2016b, 2018). Since our study

focuses on litigation between PEs, we exclude all lawsuits related to patent trolls. We manually

identify litigants that are not manufacturers, universities, government entities, or non-profit firms,

and also identify those whose firms’ activities are classified as intellectual property consulting

according to Google search websites. By merging this list with the sample of public patent trolls

used by Bessen, Ford, and Meurer (2011), we construct a list of NPEs that we use to filter out

NPE-related cases.15 For brevity’s sake, we use the term “patent litigation” to describe a firm

involved in a litigation case that is related to a public firm’s patents, but that is unrelated to NPEs.

algorithm that matches the name and location of each patent assignee that appears in 2011–2014 to the name and

location in the list of assignees in 1976-2010. As a result, we construct a database of patents granted to U.S. public

firms from 1976 to 2014. 14 Nevertheless, Bessen, Neuhausler, Turner, and Williams (2015) find no differences in announcement effects for

PEs and NPEs (after they include their control variables). 15 An alternative approach, to which our results are robust, is to omit cases filed in the Eastern District of Texas (a

venue that is often preferred by patent trolls).

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For each patent litigation case unrelated to NPEs, we manually match the plaintiff and

defendant names to the corresponding GVKEY and PERMNO in the CRSP and Compustat

databases. Doing so yields 1,216 unique public firms involved in a total of 4,721 patent litigation

cases in our sample period. We then collect the financial data of these firms from CRSP and their

accounting data from Compustat. We exclude financial firms with SIC codes between 6000 and

6999 and utility firms with SIC codes beginning with 49.

To show how litigation intensity varies over time in our sample, the upper panel of Figure

1 provides the number of patent litigation cases with publicly-traded plaintiffs in each year

(Plaintiff), the number of cases with publicly-traded defendants in each year (Defendant), and the

number of cases with any publicly-traded litigants in each year (Both).16 We observe continuous

growth in patent litigation for most of our sample period: the number of patent litigation cases

gradually increases to 2007 and then dips in 2008 (perhaps consistent with the effects of the

financial crisis). The number peaks in 2012 with 535 observations in the Both group, and then

gradually declines. The decline in recent years may be attributed partly to the Leahy-Smith

America Invents Act, which reduces patent litigation (e.g., Cohen, Gurun, and Kominers, 2018).

Since a case may involve multiple public firms and their stocks, we also plot the number of stocks

involved in patent litigation (i.e., stock-case observations).17 The lower panel of Figure 1 provides

the annual number of all stock-case observations in each group (Plaintiff, Defendant, and Both); it

shows a mostly similar pattern, and peaks in 2012 (with a value of 2,027 in the Both group).

Since we also have patent numbers of each litigation case, we find that, in 82.3% of sample

cases, all litigated patents belong to plaintiffs. In addition, in 16.5% of sample cases, all litigated

patents belong to defendants. Lastly, both plaintiffs and defendants own parts of litigated patents

in 1.1% of sample cases. Moreover, as we note in the introduction, our sample includes declaratory

judgments (9.9% of our sample cases), where the alleged infringer is the plaintiff and asks the

court to provide clarity regarding whether infringement is occurring.

16 When both the plaintiff and defendant of a litigation case are public firms, this case is counted in all three groups. 17 A firm involved in multiple cases will be counted as having multiple stock-case observations.

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3. Empirical Results

3.1. Announcement returns around disclosure

We first examine the stock market’s reaction to the disclosure of patent litigation in our

sample, based on Lex Machina from 2000 through 2014. In Table 1, we provide the cumulative

abnormal returns (CAR) of a stock in an event window [-n, m] starting from n days before the

disclosure day (day 0) to m days after the disclosure day. We consider three ways to calculate CAR.

Specifically, CAR1 is defined as the cumulative daily returns in excess of the CRSP value-

weighted market index in the period; CAR2 is defined using the CAPM model with market beta

estimated during the window of trading days -171 through -22 with a minimum of 100 non-missing

observations (as in Bhagat, Brickley, and Coles, 1994); CAR3 replaces the CAPM model in CAR2

with the Fama-French four-factor model including the market factor, SMB factor, HML factor,

and MOM factor to account for market beta, the beta to the size premium, the beta to the value

premium, and the beta to momentum premium.

In Panels A1 and A2, we focus on the following four event windows: [-1, 10], [-1, 24], [-

3, 10], and [-3, 24]. We select these windows to more effectively capture the effect of announced

litigations, since Bessen and Meurer (2012) point out the delayed announcement and reporting of

patent litigation, and express the concern of lawsuits not being disclosed (either widely or at all)

around smaller event windows.18 For example, -3 indicates that the period starts three days before

disclosure and 10 indicates that the period ends ten days after disclosure. For each disclosure, we

categorize each stock into either the plaintiff or defendant group, and report the mean and median

of cumulative abnormal returns and the number of stocks (NSTOCK) in Table 1. For each litigant

type (Plaintiff, Defendant, and Both), we also provide the number of patent litigation cases

(NCASE). It is noteworthy that one case may involve multiple public firms on both sides, so

NSTOCK is larger than NCASE.

We provide the mean, median, and statistical significance for CAR in different samples in

Table 1: Panel A1 includes all litigation cases, and Panel A2 includes the litigation cases with both

18 Bessen and Meurer (2012) focus their analysis on the [-1, 24] window, and Bessen, Neuhäusler, Turner, and

Williams (2015) consider the [-1, 3] and [-1, 23] windows.

13

a public plaintiff and a public defendant.19 We find that both plaintiffs and defendants consistently

experience significant positive cumulative abnormal returns around the disclosure dates of patent

litigation. Examining CAR1 in Panel A1, we note that the plaintiff group provides cumulative

abnormal returns of 52 basis points in the [-1, 10] window, 114 basis points in the [-1, 24] window,

55 basis points in the [-3, 10] window, and 118 basis points in the [-3, 24] window. The magnitudes

are comparable when we use CAR2 (with values of 52, 111, 55, and 114 basis points over the [-1,

10], [-1, 24], [-3, 10] and [-3, 24] windows, respectively) and CAR3 (with values of 43, 95, 43, and

95 basis points over the [-1, 10], [-1, 24], [-3, 10] and [-3, 24] windows, respectively). These values

are all significantly positive, suggesting that the stock market evaluates the initiation of patent

litigation as positive news.

The defendant group also experiences significantly positive cumulative abnormal returns.

Specifically, this group experiences cumulative abnormal returns (CAR1, CAR2, or CAR3) of 25-

33 basis points in the [-1, 10] window, 83-109 basis points in the [-1, 24] window, 25-35 basis

points in the [-3, 10] window, and 83-110 basis points in the [-3, 24] window. All of these values

are significantly positive (with the exception of the first and third windows for CAR3), suggesting

that the stock market reacts positively to the event of being sued.

We combine all defendant and plaintiff stocks in each event window in the column labelled

“Both”, and we find that this group provides cumulative abnormal returns (CAR1, CAR2, or CAR3)

of 33-42 basis points in the [-1, 10] window, 90-113 basis points in the [-1, 24] window, 33-45

basis points in the [-3, 10] window, and 90-115 basis points in the [-3, 24] window. All of these

estimates are significantly positive.

One possible concern with the results reported in Panel A1 is if public firms

disproportionately benefit from patent litigation by opportunistically suing smaller, private firms,

or have financial or other advantages when they are sued by smaller, private ones. We examine

this concern in Panel A2, where we restrict our sample to cases with both a public plaintiff and a

public defendant. After imposing this filter, we still find significantly positive CARs in most cases.

In fact, we find that the abnormal returns become even higher: for each of the 36 CARs calculated

19 When there is more than one plaintiff (defendant), we require at least one plaintiff (defendant) to be publicly listed,

in forming the sample used in Panel A2.

14

in the panel, the value exceeds the corresponding value in Panel A1. These results do not support

the argument that public firms have advantages in patent litigation against smaller, private firms,

and suggest that the significantly positive announcement returns are a general pattern existing in

different samples.

At first glance, the significantly positive announcement returns experienced by both

plaintiff and defendant groups might appear to be surprising for two reasons. First, the direct and

indirect costs associated with patent litigation are non-trivial and would potentially lead one to

expect negative announcement returns. Second, patent litigation is frequently treated by financial

analysts and media as a zero-sum (or even negative-sum) game for plaintiffs and defendants.

Consequently, the arguably intriguing pattern that we identify—both sides’ stock prices reacting

positively to the initiation of patent litigation—calls for further explanation.

Some might argue that disclosure of patent litigation might make investors aware of

specific systematic risk associated with the litigants and that such risk is resolved quickly within

10 or 24 days. These stocks might thus be expected to provide higher returns during that narrow

event window as compensation for the short-term risk. However, if the announcements make

investors aware of some risk exposure, those investors should discount the stock price immediately

on announcement dates. To examine this possible explanation, we examine the average returns in

the short windows [-1, 1] and [-3, 3], which we show in Panels B1 and B2 of Table 1. Panel B1

includes all litigation cases, and Panel B2 includes the litigation cases with both a public plaintiff

and a public defendant. Our finding of no significant price drops in either of these windows does

not support the explanation of newly discovered (and soon resolved) risk exposure. In contrast, it

support Bessen and Meurer’s (2012) finding of limited disclosure/reporting immediately around

litigation filing dates.

In Panel C, we examine the relation between the CARs of the plaintiffs and defendants to

evaluate whether a positive market reaction for a defendant (i.e., news that it intends to challenge

the lawsuit) is associated with offsetting reactions to the plaintiff. The sample in Panel C includes

all cases with both a public plaintiff and a public defendant, and shows that the correlation between

plaintiffs’ and defendants’ CARs is, in fact, positive. The correlation coefficients (CAR1, CAR2,

or CAR3) are 2.99%-3.47% for the [-1, 10] window, 6.48%-8.27% for the [-1, 24] window, 4.67%-

15

5.70% for the [-3, 10] window, and 7.62%-8.32% for the [-3, 24] window, all statistically

significant except for the first window. These significantly positive correlations indicate that the

positive CARs do not generally come at the expense of the opposing party in patent litigation;

rather, they indicate that the market reacts positively to both plaintiffs and defendants, even though

the lawsuit places them on opposing sides.

3.2. Portfolio analysis for the returns in the subsequent 12 months

Having documented a pattern of positive daily stock returns, we now examine monthly

stock returns using a calendar-time portfolio approach. At the end of every month, t, from January

of 2000 to November of 2014, we create a portfolio for firms that were involved in patent litigation

within the most recent 12 months (t-11 to t) and for their opponents. We use such a calendar-time

portfolio instead of an event-time portfolio, as the latter treats each stock involved in each litigation

case as an event and forms a portfolio based on all events. Since a stock can often be involved in

multiple patent litigations in the same period, an event-time portfolio will be subject to issues that

include a cross-correlation of event returns and overweighting stocks with multiple cases. We hold

this portfolio for the next month t+1, calculate its equal-weighted portfolio return at the end of the

month, and then adjust/rebalance the portfolio in a similar way, based on patent lawsuits

announced from t-10 to t+1 at the end of month t+1.20 Similar to Table 1, we report the portfolio

results in two panels: Panel A includes all litigation cases, and Panel B includes the litigation cases

with both a public plaintiff and a public defendant.

In Panel A of Table 2, we present the average monthly excess returns (i.e., monthly returns

in excess of the one-month T-bill rate) of the constructed portfolio (“litigation portfolio”

henceforth) and the alphas based on various linear factor models to control for various systematic

20 Because the number of stocks involved in the portfolio is small (212, on average), we focus on equal-weighted

portfolio returns to ensure that the effects of patent litigation on all stocks are presented, following Loughran and

Ritter (2000), Chan, Lakonishok, and Sougiannis (2001), Eberhart, Maxwell, and Siddique (2004), and Lyandres,

Sun, and Zhang (2007). If a firm is involved in two cases as a plaintiff and in one case as a defendant, then it is weighed

by three in the litigation portfolio. The results based on the value-weighted portfolios are robust, albeit with weaker

statistical significance.

16

risk or other potential explanatory factors.21 In Model 1, we present the average monthly excess

return, which is as high as 102 basis points per month with a t-statistic of 2.07. Such an estimate

is of reasonable statistical significance, especially given that our sample period is only 179 months

long (February 2000 to December 2014).

In order to consider various systematic risk or other potential explanatory factors in Models

2 to 8, we conduct time-series regressions by regressing the monthly excess returns of the portfolio

on various combinations of return factors, and then report the estimate and statistical significance

of the intercept term (“alpha”) in Panel A. In the Fama-French four-factor model (Model 2), the

alpha is 60 basis points per month with a t-statistic of 3.99. The six-factor model of Fama and

French (2015) that includes the profitability factor RMW and the investment factor CMA (Model

3) generates an alpha of 57 basis points per month with a t-statistic of 3.47. These findings suggest

that the return of the litigation portfolio cannot be explained by the size effect, value effect,

momentum effect, profitability effect, or the investment effect.

In Models 4 to 6, we add each of the following innovation-related factors to Model 2: an

R&D factor, a patent factor, and an innovative efficiency factor. The R&D factor is the difference

between the monthly returns of the high R&D portfolio and that of the low R&D portfolio. The

high (low) R&D portfolio consists of firms with R&D capital (the R&D expenditure over the

rolling five-year period, with 20% annual depreciation) divided by the market value of equity in

the top 20% (bottom 20%), following Chan, Lakonishok, and Sougiannis (2001) and Lev, Sarath,

and Sougiannis (2005).22 The patent factor is the difference between the monthly returns of the

high patent portfolio and those of the low patent portfolio. The high (low) patent portfolio consists

of firms with the number of patents granted over a rolling five-year period (with 20% annual

depreciation) divided by total assets in the top 20% (bottom 20%).23 The innovative efficiency

factor is the difference between the monthly returns of the high efficiency portfolio and those of

the low efficiency portfolio. The high (low) efficiency portfolio consists of firms with the log of

21 Our one-month T-bill rate and Fama and French factors—MKT, SMB, HML, UMD, RMW, and CMA—are from

the Kenneth French data library. The Hou, Xue, and Zhang q-factors—IA and ROE—are from Lu Zhang. The

Stambaugh and Yuan mispricing factors—MGMT and PERF—are from Yu Yuan's website. We thank Kenneth

French and Yu Yuan for making the data publicly available, and thank Lu Zhang for sharing the data with us. 22 R&D capital in year t is defined as the sum of [1- 0.2τ] times R&D expense in years t-τ (with τ ranging from 0 to

4), following Chan, Lakonishok, and Sougiannis (2001). 23 We scale the number of granted patents by total assets following Hall, Jaffe, and Trajtenberg (2005) and Noel and

Schankerman (2013).

17

the number of patents granted over a rolling five-year period (with 20% annual depreciation) minus

the log of one plus the industry-adjusted R&D expenditure over a rolling five-year period (with

20% annual depreciation) in the top 20% (bottom 20%).24 We find positive coefficients on the

R&D factor and the patent factor. Nevertheless, the alphas from the models that include

innovation-related factors range between 56 and 58 basis points per month and remain statistically

significant (with t-statistics above 3), suggesting that the return of the litigation portfolio cannot

be explained by R&D- or patent-based factors.

In Models 7 and 8, we consider two new factor models and find even higher alphas: the q-

theory factor model of Hou, Xue, and Zhang (2015) and the mispricing factor model of Stambaugh

and Yuan (2017). The q-theory factor model includes a market factor, a size factor, an investment

intensity factor, and a profitability factor that are derived from a q-theory model; the mispricing

factor model includes a market factor, a size factor, a management factor, and a performance factor.

The former and the latter generate alphas of 75 and 72 basis points per month with t-statistics of

3.74 and 3.33, respectively. These results suggest that the return of the litigation portfolio cannot

be attributed to investment-based risk or common mispricing factors.

Our results remain similar in Panel B of Table 2, in which we restrict our sample to lawsuits

where both the plaintiff and defendant are public. The excess returns and alphas of the litigation

portfolio are comparable to (and slightly larger than) the counterparts in Panel A, which confirms

an intriguing finding in Table 1: market reactions seem to be even more positive when the opposing

party in the litigation is another public firm. This panel thus confirms that our results are not driven

by large public firms’ advantages in patent litigation against smaller, private firms. As we observe

consistent results across different case samples in both Tables 1 and 2, we will only report the

results based on all litigation cases for the rest of our analyses for brevity.

An important implication of Table 2 is that the positive short-term stock returns that we

report in Section 3.1 are persistent and do not revert in the future. In fact, firms involved in patent

litigation not only experience short-term stock price appreciation, but also provide significantly

24 The construction of the innovative efficiency factor is motivated by Cohen, Diether, and Malloy (2013) and

Hirshleifer, Hsu, and Li (2013).

18

greater stock returns in the following year. Consequently, a positive litigation-return relation

receives further support and calls for further explanation.

We further examine the robustness of Table 2 by using alternative definitions for our

sample of relevant firms. In Panel A of Table 3, we include firms that were involved in any patent

litigation in month t-11 to t but were not involved in any patent litigation in months t-23 to t-12

(the prior year). We consider this alternative approach in constructing our sample in order to

mitigate the concern that some firms may be involved in patent litigation for many years and are

always present in our portfolio. That said, we find that our results are robust when we exclude such

firms: in Panel A, the mean excess returns and alphas of the portfolio range between 56 and 108

basis points per month, with t-statistics ranging from 2.08 to 3.57.

In Panel B of Table 3, we restrict treated firms to only defendants and find that our results

are robust when we focus on such firms: the mean excess returns and alphas of the defendant

portfolio are commensurate to (albeit slightly weaker than) their counterparts in Table 2. This

finding further supports our earlier argument that defendants in patent litigation cases may also

benefit from the announcement of patent lawsuits.

As discussed earlier, it may be inappropriate to separate our sample into the plaintiff and

defendant groups because, with Declaratory Judgments, the plaintiffs are alleged infringers and

the defendants are patent owners claiming to own patent rights. In addition, our analyses thus far

suggest that both parties experience significantly positive price increases in most scenarios. Thus,

in subsequent analyses, we include all plaintiffs and defendants in our test sample.

Another possible concern is if the findings from our portfolio analysis in Table 2 are subject

to industry effects, for some industries may be subject to greater patent litigation risk and also

provide greater stock returns at the same time. To address this concern, we construct portfolios

that include only industry-matched firms. In this approach, we define “treated” firms as those that

experienced patent litigation within the previous 12 months. For each treated firm in each month,

we match it with another firm (i.e., control firm) with the same SIC 2-digit industry code and with

the closest distance in size, book-to-market ratio, momentum, profitability, and investment. At the

end of every month from December of 2000 to December of 2014, we create an equal-weighted

portfolio (“Control”) for these control firms in each month and then track its equal-weighted return

19

in the next month. Similarly, we create an equal-weighted portfolio (“Treated”) for these treated

firms, so we can find matched control firms in each month and then track their equal-weighted

returns in the next month. We also construct a difference portfolio by taking a long position in the

treated portfolio and taking a short position in the control portfolio. We then report the mean excess

returns and alphas of the treated portfolio, the control portfolio, and the difference portfolio. The

difference portfolio generates monthly returns of 30-45 basis points per month, which are

statistically significant at the 5% level in all models.

Consequently, we use Tables 3 and 4 to confirm the robustness of a positive litigation-

return relation by excluding firms that frequently experience patent litigation, by focusing on

defendants, and by controlling for industry-specific issues.

3.3. Fama-MacBeth cross-sectional regressions for returns

In this section, we continue our analysis by using monthly Fama-MacBeth (1973) cross-

sectional regressions for all public firms (except financial and utility firms) in the

CRSP/Compustat database, which enables us to control more directly for characteristics

potentially associated with returns, so we may examine whether the effect of patent litigation on

stock valuation is distinct from that of other firm or industry characteristics. In these regressions,

our focal variable is an indicator variable, DUMMY_CASE, for whether a firm experiences patent

litigation in the prior 12 months, either as a plaintiff or a defendant. The other control variables

include returns over the previous month (RET), the log of the book-to-market ratio (LOGBM), the

log of the market value of equity (LOGME), the log of one plus the R&D expense over the past

five years using an annual discount rate of 0.2, divided by total assets (LOGXRD), the log of one

plus the number of patents granted in the past five years divided by the book value of total assets

(LOGNPATENT), cumulative returns over the past twelve months with a one-month gap (MOM11),

cumulative returns over the past three years with a one-year gap (MOM36), the growth rate of the

book value of total assets (INV), stock operating profit defined by revenue minus the costs of goods

sold, interest expense, selling, general and administrative expense, divided by book value of equity

(OP), cash flow scaled by total assets (CF), the Kaplan-Zingales index (KZ), long-term debt plus

current liabilities scaled by the book value of equity (LEVERAGE), gross profit scaled by sales

20

(PM), and the Herfindahl-Hirschman Index (HHI), based on the sales of shares in industries

defined by 2-digit SIC codes.

In Table 5, we report the summary statistics (mean, median, and standard deviation) of the

related variables for the following four litigation groups: (1) Plaintiff: those in which the firm was

a plaintiff in at least one patent lawsuit and was not a defendant in any litigation in the prior year;

(2) Defendant: those in which the firm was a defendant in at least one patent lawsuit and was not

a plaintiff in any litigation in the prior year; (3) Mixed: those in which the firm played both roles

(i.e., it was a plaintiff in at least one patent lawsuit and was a defendant in at least one patent-

related litigation in the prior year); and (4) No case: those in which the firm was neither a plaintiff

nor a defendant in any patent lawsuit in the prior year. Distinguishing our results in this manner

enables us to evaluate the differences between firms based on their status as litigants.

The median Plaintiff firm experiences two lawsuits per year, the median Defendant firm

also experiences two lawsuits per year, and firms that experience both types of litigation (Mixed)

experience eight lawsuits per year. Consistent with the intuition that larger firms experience

significantly more litigation, the number of cases per year are skewed; for example, Mixed firms

experience a mean of 13 lawsuits per year. We find that median book-to-market (BM) ratio is

larger for firms that do not experience any litigation compared to those that do experience litigation,

and R&D capital stock (XRD) is smallest for firms that do not experience litigation. Firm size (ME)

is consistent with larger firms experiencing more litigation: firms without any litigation are

smallest, plaintiffs are slightly smaller than defendants, and firms that are both plaintiffs and

defendants are the largest. Plaintiffs have more patents (NPATENT) than defendants, consistent

with the former group’s desire to enforce their intellectual property rights. Median returns over the

prior three years (MOM11 and MOM36) and operating performance (OP) are smallest for firms

that do not experience litigation, perhaps consistent with these firms being less likely to be targeted

for litigation (or behaving more conservatively to avoid litigation). Consistent with litigation

targeting relatively cash-rich firms, firms that do not experience any litigation have the lowest

level of cash flow (CF) and the highest measure of financial constraints (KZ) of all the groups.

Similarly, litigation intensity increases with profit margins (PM): firms not experiencing litigation

have the lowest profit margins, and those that are both plaintiffs and defendants have the highest

profit margins. Finally, litigation intensity declines with industry concentration (HHI), since firms

21

that do not experience any litigation have the highest value of HHI, and those that acted as both

plaintiffs and defendants have the lowest value of HHI.

In Table 6, we provide the means and t-statistics based on the times series of the

coefficients on all explanatory variables from our Fama-MacBeth cross-sectional regressions. Our

focal variable is DUMMY_CASE, which is an indicator variable equal to one for a firm in month t

if the firm is involved in any patent litigation over the past 12 months (t-11 to t) as either a plaintiff

or defendant, and zero otherwise. In each month t from December 2000 to December 2014, we

conduct a cross-sectional regression by regressing the monthly excess returns of all firms in month

t+1 on DUMMY_CASE of all firms in month t, all control variables known in month t, and industry

fixed effects based on 2-digit SIC codes. These cross-sectional regressions let us estimate the

coefficient on DUMMY_CASE in every month. We then test if the average of monthly coefficients

on DUMMY_CASE is statistically significant by using the time-series of monthly coefficients, and

report the average and t-statistics in Table 6. We apply the same procedure for the coefficients on

control variables. The first four specifications (Models (1) through (4)) do not use industry fixed

effects; the final four specifications (Models (5) through (8)) include industry fixed effects.

We find that the litigation-related variable (DUMMY_CASE) is positive in all specifications,

with a magnitude of 23-33 basis points per month; the t-statistics range between 2.50 and 3.65.

The signs and magnitudes of the coefficients of other control variables are largely consistent with

the literature. Consequently, our baseline findings that firms involved in patent litigation are

associated with significantly higher subsequent stock returns remain robust when we control for

relevant firm- and industry-specific variables. Therefore, firm and industry characteristics cannot

(fully) explain our baseline results.

4. Possible Explanations

We now examine four possible explanations for our baseline finding that patent litigation

positively predicts stock returns. First, systematic risk related to patent litigation might exist, and

firms involved in patent litigation carry higher exposure to such risk, which provides greater

expected returns. Second, systematic risk related to financial constraints might exist, such that

22

financially constrained litigants may be subject to risk to a greater extent. Third, only firms that

are rich in cash are sued and provide higher future stock returns. Finally, substantial benefits

related to patent lawsuits might exist that exceed the associated costs, and these benefits might be

undervalued by the market.

4.1. Systematic risk related to patent litigation

In this section, we provide additional analyses to examine if the return predictability

associated with patent litigation can be explained by systematic risk. Firms involved in patent

litigation are subject to discounts in their stock prices due to greater systematic risk exposure, and

are thus expected to provide higher expected stock returns to investors as risk premia. In Section

3.1, we argue that this explanation is not supported by the observation that the average return of

firms in the three-day window around the disclosure of patent litigation is positive. Nevertheless,

to further examine the risk explanation, we employ a two-pass procedure to test if the return on

the litigation portfolio serves as a risk factor. If the litigation portfolio we construct in Section 3.2

creates significantly positive monthly excess returns and alphas (as reported in Table 2), that

portfolio may be considered as a mimicking portfolio that reflects the risk compensation for

bearing one unit of risk exposure to a systematic risk associated with patent litigation (see Fama

and French (1993)).

In this section, we refer to the monthly returns on the portfolios formed by firms that are

involved in patent litigation (litigation portfolio) as the “litigation factor.” Then, we test if this

factor exists in a linear stochastic discount factor model by implementing a two-pass procedure

(see Cochrane (2001)). First, we conduct a rolling window estimation to estimate the beta

associated with the litigation risk, βi,Litigation, for stock i using its stock returns in the most recent 60

or 12 months. For example, for stock i in month t, we estimate its βi,Litigation,t by regressing its

monthly excess returns on the litigation factor and other factors (MKT, SMB, and HML) from

month t-59 (or t-11) to month t. Then, we conduct a cross-sectional regression for each month; for

each month in our sample period, we regress the monthly excess returns of all stocks on the

litigation betas of all stocks (and other betas, such as market betas) to calculate the coefficient on

the litigation betas for the month. The coefficient on the litigation beta (βLitigation) serves as an

23

estimate of litigation risk premium (known as “lambda”) in the month. Finally, we test the

significance of the litigation risk premium by the time series mean and standard deviation of the

coefficients on litigation betas across all months and report our results in Table 7: a significantly

positive coefficient associated with βLitigation indicates that the litigation factor is positively priced

in testing assets. On the other hand, an insignificant coefficient associated with βLitigation indicates

a lack of evidence that the litigation-return relation is driven by risk.

The results across all panels in Table 7 indicate that the coefficient on βLitigation is

consistently insignificant across various models and specifications.25 In Panel A, we present our

results using 60 months to estimate beta, and we use Panel B to present our results using 12 months

to estimate beta. As we cannot find any significant coefficient on βLitigation that reflects litigation

risk premium in the various models in Table 7, we note that the litigation factor is not priced in the

stock market, which casts doubt on the likelihood that our results are driven by any systematic risk

associated with patent litigation.

4.2. Systematic risk related to financial constraints

Another potential risk-based explanation for our results is that costly patent litigation

affects financially constrained firms in particular, as R&D-intensive firms incur greater systematic

risk when they are under financial constraints (Li, 2011). Costly patent litigation makes these

R&D-intensive firms subject to such risk (Lanjouw and Schankerman, 2004), and this explanation

would predict excess returns to be concentrated in those financially constrained firms. In Panel A

of Table 8, we examine this possibility by creating portfolios sorted by financial constraints, based

on the Kaplan-Zingales index (1997). In particular, we classify all firms experiencing patent

litigation by their financial constraints into three groups: the low group includes firms with

financial constraints below the 30th percentile, the high group includes firms with financial

constraints above the 70th percentile, and the middle group includes firms with financial constraints

between the 30th percentile and the 70th percentile. We then calculate the equal-weighted excess

returns for each group in each month, and report the mean excess returns and alphas (similar to

25 These results use all individual stocks as test assets; however, our results are also robust when we use industries or

portfolios as test assets.

24

Table 4). At the bottom of each panel, we report the mean excess returns and alphas (and the

corresponding t-statistics) of a high-minus-low (High-Low) portfolio that takes a long position in

the high group and a short position in the low group. If the litigation-return relation is caused by

financial constraints, we would expect a stronger predictive ability in the high group (i.e., the high

group will outperform the low group in excess returns and alphas).

We find that the mean excess returns and alphas of the High-Low portfolio range between

2 and 27 basis points per month, although these values are not statistically significant. This finding

indicates that the effect of patent litigation on stock returns does not concentrate on financially

constrained firms. Thus, it is difficult for us to attribute the litigation-return relation to the

systematic risk associated with financial constraints. In Panel B, we show that this pattern holds

when we only consider defendants (since this issue is particularly acute among financially

constrained defendants). Our results remain similar to those provided in Panel A.

Panels A and B of Table 8 thus suggest that the higher stock returns associated with firms

involved in patent litigation cannot simply be attributed to systematic risk associated with financial

constraints.

4.3. Cash holdings

In Panel C of Table 8, we consider an important driver of litigation—cash—that may

explain our baseline finding. Firms with larger cash balances are known to attract patent litigation

since they are able and/or are more willing to settle lawsuits. It is possible that our excess returns

are driven by cash-rich firms that both attract litigation and outperform in stock returns in

subsequent periods.

To evaluate this explanation, we implement one-way sorts based on our proxies for cash

holdings, which is defined as cash and marketable securities scaled by lagged total assets. In each

month, we divide firms that had experienced litigation in the prior 12 months into three groups,

based on their cash holdings. As in Panel A, the three groups are defined by the 30th and 70th

percentiles of cash holding: the low group includes firms below the 30th percentile, the high group

includes firms above the 70th percentile, and the middle group includes firms between the 30th

25

percentile and the 70th percentile, similar to Panels A and B. We present the mean excess return,

the difference in excess return between the low and high groups, and the t-statistic of the difference

in Panel C of Table 8.

We do not find that the positive returns are significantly larger among cash-rich firms,

which indicates that the excess returns are not driven by cash-rich firms being more likely to attract

patent litigation. In Panel D, we provide the results among defendants only, as the impact of cash

on litigation could be more severe among defendants. Our results remain broadly comparable to

those that we provide in Panel C, albeit with statistical significance in Models 3 and 7.

4.4. Potential benefits from patent litigation

Our last explanation is that the benefits associated with patent litigation outweigh the

associated costs, but these benefits are generally undervalued by investors. The benefits should be

particularly intuitive for plaintiffs: they ought to only litigate when doing so increases value, and

the firm is likely to win its lawsuit. In addition, the initiation of patent lawsuits may illustrate to

investors the economic value or unique applications of plaintiffs’ patents. The announcement of a

patent lawsuit also informs the stock market of managerial determination and aggressiveness in

monetizing and protecting its intellectual property (Agarwal, Ganco, and Ziedonis, 2009), which

also positively contributes to firm value.

For defendants, being sued may increase their stock prices for the following reasons. First,

defendants who feel confident that they will win their lawsuits do not choose to settle their issues

out of court. Second, a lawsuit highlights the profits earned (or potentially being earned) from the

defendants’ use of certain intellectual property. In other words, defendants would generally not be

sued if they were not generating (or are not expected to generate) substantial profits from the

alleged patent infringement. Thus, patent litigation provides more information about the (expected)

value of the defendant to the market. Third, more generally, a lawsuit reduces the information

asymmetry associated with the defendant’s profits and/or innovations for investors (because more

information about each firm’s intellectual property is revealed over the course of the lawsuit),

which may also improve stock valuation. Fourth, patent litigation highlights the substantial barriers

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to entry associated with a defendant’s operations and products, which may deter potential

competitors.

To test the explanation based on net benefits associated with patent litigation, we first

examine how patent litigation is associated with future profitability in Table 9. In particular, we

examine firms’ subsequent three-year and five-year average ROE, and perform annual Fama-

MacBeth cross-sectional regressions against DUMMY_CASE (the indicator variable for a firm

being a party in patent litigation within the prior 12 months) and other explanatory variables,

similar to our approach in Table 6, except that we use annual regressions instead of monthly

regressions. We examine the average ROE over the subsequent three-year and five-year periods

since the benefits associated with the patent litigation are likely to be recognized over time, albeit

with a delay. As we do in Table 6, we control for investment (INV), the log book-to-market

(LOGBM), the log market value of equity (LOGME), the log patent output (LOGNPATENT), and

the log R&D investment (LOGXRD). Additionally, to control for persistence in profitability, we

include the prior year’s ROE (ROE). Also, to control for mean reversion in profitability (Fama and

French, 2000), we include the prior one-year change in ROE (ΔROE). We also include industry

fixed effects.

Consistent with our proposition of litigation benefits, we find that the coefficient on

DUMMY_CASE is significantly positive: firms involved in any patent litigation (DUMMY_CASE

=1) are associated with an increase in the mean five-year ROE of 3.0% and an increase in the mean

three-year ROE of 2.4%. The slopes of the other explanatory variables are also as expected, and

experience little change with respect to the two different windows of future profitability. These

results support our argument that patent litigation may be, on average, beneficial to patent litigants

by both signaling the value of patents and the potential of associated profits to the public and

highlighting barriers to potential new competitors.

Another way to assess the increased barriers to entry associated with patent litigation is to

examine future competition at the industry levels. In Table 10, we examine the effects of patent

litigation on future five-year and three-year averages of Similarity and HHI as defined in Hoberg

and Phillips (2016), since both are important measures of competition in an industry (these

measures are based on textual analysis from firms’ 10-K annual filings of their business

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descriptions). Consistently, our results point to the effects of increased barriers to entry following

patent litigation. Firms’ products become more differentiated as Similarity in the following five

years (three years) declines by 0.28 (0.29) around this time. Industries also become less

competitive as defined with the sales-based HHI, with average future HHI increasing by 1.5% in

the subsequent five-year and three-year periods. Consequently, using a range of industry-level

proxies of competition, we find that patent litigation leads to increased barriers to entry.

Tables 9 and 10 thus collectively help identify the mechanisms of our results by providing

evidence of higher future profits for patent litigants, as well as of increased barriers to entry in

their industries.

4.5. Undervaluation due to information delay

After presenting the positive relation between patent litigation and subsequent profitability

and the negative relation between patent litigation and future competition, we argue that investors

are unaware of and thus underreact to the positive value associated with firms’ involvement in

patent litigation. Such underreaction leads to our baseline finding that litigation predicts stock

returns. The details of patents and lawsuits are often complicated in ways that challenge investors,

even institutional ones, to readily analyze and comprehend these details (Hirshleifer, Hsu, and Li,

2013, 2018).26 In Table 11, we examine the information delay argument by using a one-way sorted

portfolio, similar to our approach in Table 8.

To do so, we rely on two information delay measures: stock return R2 as in Hou, Peng, and

Xiong (2013) and the delay in stock-price reactions to new information as in Hou and Moskowitz

(2005). In Panel A, we use the stock return R2 to measure information delay. Following Hou, Peng,

and Xiong (2013), stock return R2 is calculated as the R2 from regressing the stock’s monthly

returns on the contemporaneous returns of the CRSP value-weighted index portfolio and industry

portfolio, based on Fama and French 48 industries. We require a minimum of 24 observations to

estimate R2. Hou, Peng, and Xiong (2013) argue that a low R2 is associated with price inefficiency,

26 See Footnote 3 for a review of prior studies that have provided empirical evidence and explanations for the market’s

undervaluation of innovation-related information.

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in part by showing that a lower R2 is associated with greater medium-term momentum, and also

associated with stronger long-term price reversal.

In Panel B, we use Hou and Moskowitz’s (2005) price delay measure, which reflects the

delay in stock-price reaction to new information and is mostly driven by frictions in investor

recognition. Following Hou and Moskowitz (2005), we estimate price delay by the following two

steps. In the first step, at the end of June in each calendar year, we calculate the “first-stage

individual stock delay” measure. This measure is defined as one minus the R2 from the regression

of an individual stock’s weekly returns on the market portfolio and four weeks of lagged returns

on the market portfolio over the prior year. In the second step, we derive “second-stage portfolio

delay” measures to mitigate the errors-in-variables problem in individual stock regressions.

Specifically, at the end of June of each calendar year, we sort stocks into deciles based on their

market capitalization. Within each size decile, we then sort stocks into deciles based on their first-

stage individual delay measures. We then compute the equal-weighted weekly returns of the 100

size-delay portfolios over the following year from July to June, and we re-estimate the regression

in step one by using the entire past sample of weekly returns for each of the 100 portfolios in the

second stage. We then assign the computed delay measures for each portfolio to each stock within

the portfolio, and defined them as stock-level price delay.

In each month from January 2000 to November 2014, we classify firms that had

experienced litigation in the prior 12 months into three groups based on each of the information

delay proxies: the group consisting of firms with information delay proxies below the 30th

percentile is defined as the “low” group, the group consisting of firms with information delay

proxies above the 70th percentile is defined as the “high” group, and the group consisting of firms

with information delay proxies between the 30th and 70th percentiles is defined as the “middle”

group.27 We present the mean excess return, the difference in excess return between the high and

low groups (High-Low), and the t-statistic of the difference in Table 11. In both panels, we show

that the litigation-return effect is larger among firms with greater information delay; this difference

27 Specifically, when we use stock return R2 as the proxy for information delay, the high group includes firms with the

lowest 30% R2, the low group includes firms with the highest 30% R2, and the middle group includes the remaining

firms. When we use price delay as the proxy for information delay, the high group includes firms with the highest 30%

price delay, the low group includes firms with the lowest 30% price delay, and the middle group includes the remaining

firms.

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is statistically significant across models in Panels A and B based on R2 and price delay,

respectively.

In Panel A, litigants with low return R2 are associated with significantly higher excess

returns than those with high return R2, as shown by the significantly negative differences in the

High-Low row. In particular, the differences range from 81 to 107 basis points per month

(depending on the model used); all differences are statistically significant at the 5% level. In Panel

B, based on the price delay measure of Hou and Moskowitz (2005), we again find that litigants

with high information delay are associated with significantly higher excess returns. The differences

in excess returns between the high and low groups range from 63 to 84 basis points per month, and

are statistically significant in all but two models.

Table 11 thus provides support for our fourth explanation of our results based on

underraction: the benefits of firms experiencing patent litigation are not immediately observed

and/or understood by investors, which results in positive return drifts in both daily and monthly

frequencies.

5. Conclusion

Patent litigation has important and wide-ranging effects on firms, and firms are

experiencing patent litigation with increased frequency. Although there are significant direct and

indirect costs associated with this type of litigation, there are also well-documented benefits,

including increases in patent citations and patent value following litigation, signaling managers’

determination and confidence to the market, increased barriers to entry for more litigation-

intensive industries, reduced information asymmetries for investors, and increased prominence of

firms’ intellectual property.

In this paper, we examine the effects of patent litigation on firms’ stock prices, focusing on

the role of practicing entities. Our empirical results suggest that the positive effects associated with

patent litigation outweigh the associated costs. Consistent with these effects, we show that both

plaintiffs and defendants experience positive stock returns following the announcements of patent

30

litigation. We then show that these firms experience significantly positive returns over the longer

term as well.

Further analyses indicate that firms involved in patent litigation provide higher future

profits and experience less future competit