1 The Impact of Adoption of Wrongful Discharge Laws on Union Density: 1983-2014 Job Market Paper 31 October, 2016 Eric Hoyt, University of Massachusetts Amherst Abstract: Union membership has declined precipitously in recent decades following the rise of retaliatory firing and other harsh union avoidance tactics. This paper utilizes variation in the time and location (i.e. state) of adoption of wrongful-discharge court doctrines in the U.S. over the 1980s and 1990s as a source of numerous natural experiments regarding the impact of employment protection on union membership and union coverage outcomes. This analysis finds that adoption of one wrongful discharge law in particular, the implied contract doctrine, is associated with small but consistent increases in union density, driven by increases in union membership and coverage levels. The impact is greatest for the private sector, manufacturing industry, and younger male and female workers with less than college education. This paper proposes that the implied contract doctrine boosts union density by lowering the cost of seeking union membership and coverage. The findings also suggest that employment protection policies may foster greater equity in labor market outcomes by age and sex, as well as serve as a counterweight to the pressures undermining union density over this period. Keywords: labor unions, union density, wrongful discharge laws, employment protection Acknowledgements: I am grateful for the guidance of Professors Fidan Kurtulus, Gerald Friedman, and Eve Weinbaum as well as for the numerous helpful comments of professors and fellow graduate students at the University of Massachusetts Amherst. Author Contacts: [email protected], 200 Hicks Way University of Massachusetts Amherst, Amherst, MA 01003, (413)-362-4517
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The Impact of Adoption of Wrongful Discharge Laws on Union Density: 1983-2014
Job Market Paper
31 October, 2016
Eric Hoyt, University of Massachusetts Amherst
Abstract: Union membership has declined precipitously in recent decades following the rise of
retaliatory firing and other harsh union avoidance tactics. This paper utilizes variation in the
time and location (i.e. state) of adoption of wrongful-discharge court doctrines in the U.S. over
the 1980s and 1990s as a source of numerous natural experiments regarding the impact of
employment protection on union membership and union coverage outcomes. This analysis
finds that adoption of one wrongful discharge law in particular, the implied contract doctrine, is
associated with small but consistent increases in union density, driven by increases in union
membership and coverage levels. The impact is greatest for the private sector, manufacturing
industry, and younger male and female workers with less than college education. This paper
proposes that the implied contract doctrine boosts union density by lowering the cost of seeking
union membership and coverage. The findings also suggest that employment protection policies
may foster greater equity in labor market outcomes by age and sex, as well as serve as a
counterweight to the pressures undermining union density over this period.
Keywords: labor unions, union density, wrongful discharge laws, employment protection
Acknowledgements: I am grateful for the guidance of Professors Fidan Kurtulus, Gerald
Friedman, and Eve Weinbaum as well as for the numerous helpful comments of professors and
fellow graduate students at the University of Massachusetts Amherst.
Author Contacts: [email protected], 200 Hicks Way University of Massachusetts
Amherst, Amherst, MA 01003, (413)-362-4517
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I. Introduction
In the United States, wrongful-discharge common law doctrines have been growing in
importance as state courts have followed a trend of precedent-setting employment law rulings.
These decisions, though occurring with significant variation between states in timing of adoption
and broadness of applicability, have nonetheless altered the basic institutional foundation
underlying the U.S. labor market since the late 19th
century, the common-law doctrine of
employment-at-will. Employment-at-will is the legal principle which allows employers to
dismiss their employees for any reason or no reason at all.
A vibrant economic literature has arisen in recent decades concerning the impact of wrongful
discharge laws, and employment protection legislation more generally, on employment, wages,
productivity, the use of temporary workers, job flows, and hiring rates, among other labor market
outcomes. A contemporaneous literature has emerged within the labor movement seeking to
understand the dramatic fall in union membership and union economic influence in the U.S.
economy which began in the years immediately following WWII and has accelerated in recent
decades. This work is the first to investigate the impact of wrongful-discharge laws on union
membership and coverage outcomes, adding a previously unexamined dimension to the literature
on the labor market ramifications of wrongful discharge laws, while contributing a rigorous
analysis of the role played by employment law changes in mediating union membership trends to
the literature on union decline.
The literature on the labor market effects of wrongful discharge laws is founded on the
theoretical model articulated by Lazear (1990) and Blanchard and Katz (1997), which predicts
that employment protection policies, by increasing firing costs, have an indeterminate effect on
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employment levels. That is, since employment protection makes it more difficult for employers
to fire workers, it slows both the rate of dismissal and the rate of hiring. The theory predicts that
in the short-run employment levels may increase, decrease, or be unaffected, depending on the
particular labor demand conditions in the economy. However, in the long run, it is predicted, as
Autor et. al. (2006) and others have noted, that employment levels and wages may fall if rising
labor costs caused by employment protection policies are not offset by productivity gains.
One method of analysis, adopted most notably by Autor, Donahue III, and Schwab (2006),
Miles (2000), Dertouzos and Karoly (1992, 1993), and Kugler and St. Paul (2004), and which is
the method used in this paper, utilizes variation in the time and location (i.e. state) of adoption of
wrongful-discharge court doctrines over the 1980s and 1990s as a source of numerous natural
experiments regarding the impact of employment protection on labor market outcomes. Autor
et. al. (2006), with monthly outcome variables for employment, wages, and explanatory
variables for policy adoptions, come to the conclusion that while adoption of two of the three
main wrongful discharge policies has no statistically significant effect on employment outcomes,
adoption of the implied contract doctrine, , which creates an exemption to employment-at-will
through counting employers’ verbal and written promises of job protection as contractually
binding, has moderate and consistent negative effects on employment. The authors find no
significant effect on wages. Further, they find that the effect of the implied contract doctrine is
felt strongest in the short-term by those most likely to switch jobs often, such as women and less
educated workers. They found that the long-run impact was felt strongest by older and more
highly educated workers. They also found stronger negative impacts of the implied contract
doctrine on manufacturing as opposed to nonmanufacturing employment.
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Kugler and Saint-Paul (2004) reveal disemployment results consistent with an explanation
based on asymmetric information and employee-employer matching. They find evidence that
wrongful discharge doctrines motivate a process of adverse selection in which employers
discriminate in hiring against the unemployed. They argue that employers, since they have
imperfect information about employees, assume that workers who are already unemployed are
more likely to be “lemons”, or bad investments from the perspective of the firm, because their
status might reflect non-observable qualities such as a lack of motivation or work effort. The
authors predict that, after firing costs rise following the adoption of wrongful-discharge laws,
employers may become even less likely to hire unemployed workers. They find that this effect is
lessened for union members, whose status as unemployed is mediated by seniority and less likely
to reflect unobservable worker characteristics. Kugler’s results are similar to Autor et. al (2006)
in that they points toward stronger disemployment effects for precarious labor market
participants.
Autor (2003), while linking the adoption of wrongful discharge laws with greater use of
temporary help services, notes that states with slower union decline also witnessed greater
utilization of temporary workers and suggests that this reflects employers’ efforts to avoid costs
of union wage premia. Without discounting this highly reasonable line of argument, my analysis
contributes another angle to the correlation between wrongful discharge doctrines, unions, and
greater utilization of temporary workers; that is, my paper suggests that slower union decline
may itself be partially caused by passage of wrongful discharge laws.
My analysis adds to this literature by showing that adoption of one wrongful discharge law in
particular, the implied contract doctrine, is associated with modest but consistent increases in
union membership and coverage density. The greatest increase in union membership and
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coverage rates occurs for the private sector (0.84% and 1.03%), manufacturing industry (1.39%
and 1.59%), and younger male workers with less than college education (1.91% and 1.97%).
Significant increases in union membership and coverage levels in the private sector (5.37% and
6.34%), manufacturing industry (8.06% and 9.15%), and for male workers with less than college
education (11.34% and 11.38%) appear to be the main driver of the rise in union density. The
implied contract doctrine is also associated with increases in union density for young female
workers with the less than college education (0.74% and 1.74%), but this result appears
explained by an only marginally significant 10.19% increase in the level of union coverage and a
nonsignificant increase in union membership levels of 4.55%.
One could reasonably suspect that these results point towards reverse causality because
greater union growth or decline over this period might increase the probability of states’ adoption
of wrongful discharge doctrines. Strong and growing unions could be more likely to demand job
security throughout the labor market, and large losses in collectively bargained job security in a
state could spur demands for broad employment protection. However, Autor (2003) and Miles
(2000) find that neither above average union growth or union decline is associated with greater
probability of states’ adoption of wrongful discharge doctrines.
My results seem to contradict previous findings regarding little or no impact of wrongful
discharge laws on the employment levels of unionized workers, given they are already covered
under employment protections within collective bargaining agreements. However, this analysis
proposes that the implied contract doctrine boosts union density, membership, and coverage
levels by lowering the costs to workers of seeking union membership and coverage, rather than
through an impact on employment levels. Some evidence emerges of statistically significant
disemployment effects for nonmembers and noncovered workers within particular industry and
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demographic subgroups, but none of these correspond, except perhaps in the case of the
construction industry, to the instances of union density growth found within this analysis. This is
especially important since an increase in union density could occur if the numerator, consisting
of the level of employed union members (or covered workers), remains relatively fixed, while
the denominator, consisting of the sum of the level of employed union members and
nonmembers (or covered and noncovered workers), declines as result of employment losses for
nonmembers (or noncovered workers).
In the context of large and precipitous decline in union membership and coverage, and given
the significant role of retaliatory firings in propelling this development1, the finding in this paper
on union density within the private sector, manufacturing industry, and among less educated and
younger men and women suggests that employment protection policies may serve as a
counterweight to the pressures undermining union membership. Since the private sector and
manufacturing industry experienced some of the largest declines in union membership and
coverage over this period, and younger men and women have been historically underrepresented
among union members, these findings suggest that employment protection may also be one way
to grow union membership within, as well as beyond, its traditional base.
II. Institutional Background of Wrongful Discharge Laws
Before proceeding further, it is worthwhile to comment briefly on the institutional
background of wrongful-discharge doctrines and union density in the United States. There have
been three primary wrongful-discharge doctrines that have proliferated in recent decades: the
1 Researchers have found that since the 1980s nearly 1-in-4 union elections is marred by retaliatory discharge, and
pro-union workers face a 2percent-3percent chance of suffering retaliatory discharge (Schmitt and Zipperer 2007,
LaLonde and Meltzer 1991, and Weiler 1983). Even following successful union representation elections, research
shows that only 48percent of organized workplaces have a collective bargaining agreement one year following the
election (Bronfenbrenner 2009).
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implied contract doctrine, the public policy doctrine, and the covenant of good faith and fair
dealing.
The implied contract doctrine states that language in employee manuals and oral promises
made by supervisors detailing a specific duration of employment or standard procedures for
dismissal can override the default of employment-at-will. In some states, the implied contract
doctrine arises in the absence of written or oral promises even when employers’ actions establish
a repeated past practice of discharging only for justified reasons (Hirsch, Secunda, and Bales
2013). Toussaint v. Blue Cross Blue Shield of Michigan is frequently cited as one of the best
cases, though not the first, to articulate the implied contract doctrine. Charles Toussaint worked
for many years for his employer Blue Cross Blue Shield. When he started the position, he was
given told that his employment was secure as long as he preformed his job well. After he was
fired without an explanation, Toussaint sued for wrongful discharge. The Michigan Supreme
Court ruled that his employer’s statement, as well as language in the company’s employee
handbook, constituted legally enforceable guarantees of dismissal for “good cause” (Muhl 2001).
The implied contract doctrine, as it has developed in practice, is the most broadly applicable of
wrongful discharge laws, since it applies to every discharge within a workplace, creating a “good
cause” standard rather than prohibiting particular “bad faith” discharges. The damages awarded
to litigants are contractual, meaning employers owe the employee the salary and wages they
would have paid that employee had they continued working since the date of discharge, less any
amount of wages and salary the employee was able to earn in new employment. In many late-
adopting states, and in certain states that initially adopted strong interpretations of the implied
contract doctrine, courts have weakened this policy by sanctioning employers’ use of disclaimers
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in employee handbooks which counteract implied contractual obligations by explicitly
announcing to employees their status as at-will.
The public policy doctrine is an exception to the at-will rule that arises when an employee is
discharged after invoking a publicly protected right or obligation, such as filing a worker’s
compensation claim or attending jury duty. The doctrine also arises when an employee is fired
for refusing an employer’s request to violate public policy, ranging from a clear infraction like
investment fraud to broader like endangering public health (Hirsch, Secunda, and Bales 2013).
The first case to illustrate the public policy doctrine was decided in August of 1959 in California.
An employee of the Teamsters labor union was terminated immediately after having provided
truthful testimony to a state governmental body on internal corruption within the union, violating
instructions given by his employer to deny such accusations under oath. The court ruled in his
favor when he challenged the dismissal, establishing a general principle that has been developed
further in court rulings (Muhl 2001). The doctrine permits tort damage awards for successful
litigants. These damages can be much larger than the contract damages of the implied contract
doctrine, as they can also include punitive damages and legal fees, for instance, in addition to the
employees’ lost wages and salary. The per litigant cost to employers is larger than for the
implied contract doctrine. Tort damages are included as an additional disincentive to employers
since the harm inflicted is both to the individual discharged employee as well as to the broader
community. Since cases involving the public policy doctrine are less common than those arising
under the implied contract doctrine, given they pertain only to particular instances of discharge
for “bad cause”, it is likely that the overall cost of the doctrine to employers is smaller than that
of the implied contract doctrine.
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Finally, the good faith doctrine, if followed to the letter, requires that every discharge
decision be justified. However, in practice the good faith doctrine has developed, like the public
policy doctrine, only to prohibit particular cases of morally reprehensible termination. For
instance, one of the first court decisions to adopt the good faith doctrine was Fortune vs.
National Cash Register Co. in Massachusetts in 1977. Fortune was a sales employee who just
after finalizing a deal entitling him to a substantial commission was discharged for a period of
time during which he would have received payment. He was rehired shortly afterwards. The
company structured its commission payouts to accrue to its sales personnel in two installments:
when a deal was first signed with a customer, and then upon final delivery of the goods. The
company discharged the salesperson without justification after the deal was signed but before the
delivery of goods, leaving the employee with less of the commission owed to him. Fortune sued
for wrongful termination, the Massachusetts Supreme Court affirmed that that he was discharged
in bad faith, and he was awarded damages. Cases involving the good faith doctrine, like the
public policy doctrine, are relatively rare since they prohibit specific instances of “bad cause”
discharge. While tort damages were permitted in most early interpretations of the doctrine, the
policy has developed in practice to limit litigant awards to contract damages in most adopting
states. Thus, since cases involving the good faith doctrine are less common than the implied
contract doctrine yet entail the same per litigant costs to employers, it seems likely this doctrine
has the smallest overall cost to employers (Hirsch, Secunda, and Bales 2013).
III. Data Sources
This paper uses state court decisions from 1970 through 2014 to construct explanatory
variables on wrongful discharge doctrine adoptions. I employ the method of Morris (1995),
which was adopted by Autor et. al. (2006) for the years 1970 through 1999, by coding the first
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state Supreme Court or Intermediate Appellate court ruling that demonstrates a clear acceptance
of each of the three wrongful discharge doctrines.2 While Autor et al. (2006) utilize this data to
construct variables on a monthly basis from the year 1978 through 1999, this paper, in order to
match union membership and coverage data available only at the yearly level for all states, and
only from 1983 onwards, constructs annual policy adoption variables for 1983 until 2014, the
most recent year of available union data. The data set over this period includes the sustained
(i.e. at least five year long) adoption of the implied contract doctrine by 32 states, the public
policy doctrine by 26 states, and the good faith doctrine by 6 states. Table 1 includes a complete
list of all wrongful discharge doctrine recognitions by state as of 2014, the last year of the study,
as well as the closest year3 to the date of policy adoption or repeal. Maps 1 through 3 give a
geographic snapshot of state recognition of the three wrongful discharge doctrines as of 2014.
Table 1 and Maps 1 through 3 show the complete list of recognitions, adoptions, and repeals
stretching back to 1960, long before the period of analysis in this paper. They show a total of 41
2 While I first drew solely upon the legal adoption data for 1970 through 1999 constructed by Autor et. al. (2006)
and are publicly accessible on David Autor’s MIT faculty website,
http://economics.mit.edu/faculty/dautor/data/autdonschw06, and were first accessed on November 14, 2012, I
updated the legal adoptions to 2014 based on my own analysis of the case law. My analysis conforms to Morris
(1995), but differs from Autor et. al. (2006), in counting the implied contract doctrine as being adopted without
subsequent repeal in Arizona from 1983 onwards. My coding differs from Autor et. al. (2006) in not coding
Louisiana as having adopted the good faith doctrine in 1998. Autor et. al. (2006) may have misclassified Barbe v.
A.A. Harmon & Co. (1998) as an adoption of a good faith exception to employment at-will when the text of the case
decision shows that it only created a good faith covenant in the provision of employee bonuses. Finally, my reading
of the case law and secondary literature made clear that the New Hampshire Supreme Court, in its Centronics V.
Genicom (1989) decision, readopted the good faith doctrine. This finding contradicts Morris(1995) and Autor et. al.
(2006) as they seem to have both missed this development in their legal coding.
3 In order to approximate the detail of Autor et. al. (2006) and Morris (1995)’s data regarding month of policy, I
lump adoptions into the calendar year of passage if occurring from January through July, and into the following
calendar year for policy adoptions occurring during July through December. For instance, the adoption by New
Hampshire of the implied contract exception in September of 1988, because it fell in the second half of the calendar
year and was technically closer to January 1989 than January 1988, was classified as occurring in the year 1989. As
one sees, for instance in Appendix A of Autor et. al. (2006), the actual date of the court case decision in this case is
in August of 1988 in New Hampshire, however I follow the method of Autor et. al. (2006) in classifying passage as
taking effect in the first full month following the court decision, September 1988, for this and all adoptions when
observed at the monthly-level of analysis.
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adoptions of the implied contract doctrine, 43adoptions of the public policy doctrine, and 13
adoptions of the good faith doctrine as of 2014. Only Florida, Georgia, Louisiana, and Rhode
Island never adopted any wrongful discharge doctrines. Also, no change occurs in states’
recognition of wrongful discharge doctrines from the adoption of the good faith doctrine in
January 1994 through the end of the analysis in December 2014.
Yearly state-level union membership and coverage outcome variables for the years 1983 to
2014 were constructed from the Current Population Survey Outgoing Rotation Group monthly
files following the methodology of economists Barry Hirsh and David Macphearson (2003).
Monthly microdata data was extracted and then averaged, first over all individuals in each state
for each month, and then over each year in each state, for the number of employed wage and
salary workers over aged 16 and below 65 years of age, the subset of these workers who are
union members, and the subset of these workers who are covered by union collective bargaining
agreements. These figures were constructed for total state workforces, for public and private
sectors, for major industrial categories, and for eight sex-age-educational attainment
demographic categories. This data was then used to construct yearly state union membership
and coverage density figures for each state over the period 1983 to 2014, per the following
formula:
% Memst = (Membersst/Employmentst)100
% Covst = (Coveredst/Employmentst)100
This study utilizes the union membership rate and union coverage rate, defined as the %age of
employed wage and salary employees who are union members and %age who are covered by a
union collective bargaining agreement, in order to capture the broadest possible window into the
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impact of wrongful discharge laws on union density. The subscripts s and t on the union density
measurers, as in the econometric specifications in the following section, indicate the relevant
state and year for each observation. While multiple attempts were made to extract and construct
figures by race and ethnicity, zero-valued union membership and coverage variables, even using
the broadest (i.e. non-white vs. white or nonhispanic vs. hispanic) demographic categories,
persisted for several states during multiple years over the period 1983 to 2014. In this paper,
outcome variables for the levels of nonmembership and noncoverage are constructed from the
union membership and coverage level data by subtracting from the total, sectoral, industrial, and
demographic group aggregate employment levels the respective levels of union membership and
coverage for each category.
IV. Econometric Models
Firstly, to give a picture into the dynamics of the impact of wrongful discharge doctrines on
union membership and coverage outcomes over the year surrounding adoption in treatment
relative nonadopting control states, this paper utilizes, before delving into the full fixed effects
specification, a more base line differences-in-differences regression with yearly leads and lags
from two years before to two years after adoption, which is nearly identical to the dynamic
specification included in Autor et. al. (2006). The model is
Yst = αs + δt + Σ2
τ= -2 γτ Ls,t-τ + δt x Regions+ ԑst (1)
Yst is the outcome variable which, in this specification, captures the % change in union
membership and union coverage rates in a state in each year. The variable αs is a state intercept
that captures idiosyncratic, unobserved, and pre-existing features of each relevant state which
might impact the change in union outcomes irrespective of the year of observation. δt represents
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year fixed effects which control for yearly state-invariant shifts in the regression intercept,
accounting for common national shocks to union outcomes. δt x Regions is a term composed of
fixed effects representing the four U.S. Census regions interacted with the year fixed effects in
order to control for passing regional shocks to union membership and coverage outcomes. The
dummy variable Ls,t switches on from zero to one only during the year a state court adopts a
wrongful discharge doctrine, and permits the contemporaneous estimation of the impact of each
doctrine on union density in adopting states relative to nonadopting states. The parameter of
interest γτ gives the first-year impact of the adoption of wrongful discharge laws on the average
change in union membership and coverage outcomes between adopting and nonadopting states.
The specification is weighted by the share of national population aged 16 to 64 in each state in
the given year, and uses Huber-White robust standard errors. Identification of γτ comes from
within-region variation in union membership and coverage outcomes between adopting and
nonadopting states.
This specification is used to capture the impact of wrongful discharge laws on union density
for young males with less than college education, to give a representative view of the dynamics
surrounding policy adoption. A much broader investigation of outcomes by sector, industry, and
demographic groups is reserved for regression of the full fixed effects specification. In order to
match the results for the full fixed-effects regression, the overall sample window of adoptions
analyzed was restricted to 1985 to 2011. By omitting analysis on adoptions from before 1985, as
well as data on one and two year lags for adoptions before 1985, like with regressions using
equation 2, the overall number of adoptions actually studied becomes 19 implied contract
doctrine adoptions, 21 public policy doctrine adoptions, and 6 good faith doctrine adoptions.
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The treatment observations in the full fixed effects difference-in-differences regression
analysis, following equation 1 of Autor et. al. (2006), consist of the five years surrounding policy
adoption of one of the three wrongful discharge doctrines, from two years before to two years
after the year of policy change. The control groups are composed of the states which do not
adopt any of the three wrongful discharge laws over the five years surrounding each adoption in
a treatment state. States which adopt a wrongful discharge doctrine before or after the five-year
period surrounding policy change in a treatment state enter the control group for that adoption.
The five-year treatment and control window is used instead of a panel composed over the entire
sample period to account for possible serial correlation in yearly state union membership and
coverage outcome data, as well as to identify discrete changes in the outcome variables
attributable to policy adoption.
The full fixed effects difference-in-differences regression specification is:
Treatst is an indicator for the period 2 years before to 2 years after the year of adoption of a
wrongful-discharge law in treatment state s. Postst is an indicator for the period from the year of
adoption4 to 2 years after the year of adoption for all states, both adopting and nonadopting.
TreatstPostst is an interaction term representing the three year period following policy adoption,
in the treatment state only, beginning with the first year of adoption. The coefficient of interest
β2 is an estimate of the pre-post change in the outcome variable in adopting states relative to the
corresponding change in non-adopting states. Postpostst is a dummy variable that turns on for
4 This analysis does not retain the “doughnut hole” restriction of Autor et. al. (2006); that is, it does not omit data for
the first twelve months immediately following policy adoption to account for a response period during which
employers become aware and begin to react to the policy change. My analysis is yearly so this restriction based on
implementation delay over the first few months did not seem appropriate.
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each state that reenters the control group 3 years after the year of policy adoption to account for
any bias caused by the enduring effect of the policy adoption after the five-year treatment
window expires. These window lengths are the same as the windows in Autor et. al. (2006) for
consistency.5
Yst, the outcome variable6, is identical to that in equation 1, with the exception that it is also
represents the approximate %age change in the level of union membership (i.e. the change in the
natural logarithm of the level of union membership), the approximate %age change in the level
of nonmembership, the approximate %age change in the level of union coverage, and the
approximate %age change in the level of noncoverage over sectors, industries, and sex-age-
education groups. The variables αs, δt , and δt x Regions are defined identically as for equation
5 To illustrate the legal doctrine adoption variables, consider the following example. Massachusetts adopted the
implied contract doctrine in 1988, while neighboring Rhode Island did not adopt any wrongful discharge law over
the period studied. Massachusetts is a treatment state, so the variable Treatst takes on the value one for the years
1986, 1987, 1988, 1989, and 1990, and takes on the value of zero in years before 1986 and after 1990. Since Rhode
Island did not adopt any wrongful discharge law over the five years surrounding adoption of the implied contract
doctrine in Massachusetts, Treatst always remains zero in this state. In this case, Rhode Island is included within the
control group. The variable Postst takes on the value one in Massachusetts and Rhode Island, and in all other states,
for all years that this policy remains in effect in Massachusetts, beginning with the year of adoption, which in this
case is the year 1988 to the end of the dataset in 2014. Treatst* Postst is an interaction term representing the three
year period, in the treatment state only (i.e. Massachusetts), beginning with the year of adoption of the implied
contract doctrine in 1988. In other words, Treatst* Postst takes on the value one only in Massachusetts and only for
the years 1988, 1989, and 1990. Maine, another neighboring state, also did not adopt any wrongful discharge laws
over the five years surrounding 1988, so it is included within the control group. However, Maine adopted the
implied contract doctrine in 1978. In this case, the variable PoststPostst switches from zero to one for Maine starting
the year after the treatment window for its adoption ended, which is 1981, and takes on the value one for the rest of
sample period, that is, until 2014. If Massachusetts becomes eligible as a control for a later state adoption, its
PoststPostst variable, which switches on beginning in 1991, would take on the same role as it plays in Maine in this
example. The illustration described here is summarized in Table 2. One important consequence of this five year
treatment window structure for the fixed effect specification is that it omits analysis of policy adoptions occurring
from 1983 and 1984, and from 2012 to 2014.
6 The outcome variables for the full fixed effect specifications are investigated over a slightly broader range of
economic categories than in Autor et. al. (2006), expanding beyond an industrial breakdown of manufacturing and
nonmanufacturing to include construction, service, and transportation-communication-utilities industries, and by
including analysis at the level of economic sector. The eight sex-age-educational attainment demographic categories
are nearly identical to those used by Autor et. al. (2006), categorizing young workers as aged 16 to 39 as opposed to
18 to 39 in order to conform to the method of union membership and coverage data construction of Hirsch and
Macphearson (2003).
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1. The term αs * Treatst is a state-time specific intercept that accounts for changes in union
membership and coverage rates, and the levels of union membership, nonmembership, coverage
and noncoverage attributable to idiosyncratic, unobserved, and pre-existing features of the
relevant treatment state over the respective five-year treatment window in that state. As with
equation 1, each regression is weighted by the share of national population aged 16 to 64 in each
state in the given year, and Huber-White robust standard errors are used to account for within-
state error correlations. Identification of β2 comes from variation in union membership and
coverage outcomes between adopting and nonadopting states within the same census regions.
This specification gives an estimate of the causal effect of policy adoption controlling for a
variety of state, regional, national, and time-specific characteristics that may influence union
membership and coverage outcomes.
V. Results
A. Event Study Difference-in-Differences Regression
Figure 1 illustrates the dynamic effects, obtained using equation 1, of state adoption of the
wrongful discharge doctrines on union membership and coverage outcomes from two years prior
to three years post in adopting states relative to nonadopting states, and normalized to zero for
the year of adoption. These regressions, in contrast to the broader investigation of an extended
list of outcome variables used in the full fixed-effects analysis, are restricted to results for young
males with less than college education in order to highlight the dynamics surrounding policy
adoption, since union outcomes for this demographic group witnessed the most significant
impact of policy adoption.
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Young less educated male union membership and coverage rates grew substantially to a peak
of 1.30% to 1.19%, for union membership rates and coverage rates respectively, over the two
years following the year of adoption of the implied contract doctrine. No clear trend emerges for
either the public policy or good faith doctrine adoptions. Young less educated male union
density appears to decline somewhat following good faith doctrine adoption, though there is no
steady pretreatment trend and significant year-to-year variation in union density. However, these
findings should be taken with some caution given the small size (6 treatment states) of the good
faith adoption sample over this period and the outsized role of large membership and coverage
declines, as well as the role played by possible confounding events, at least in one state,
Arizona7.
B. Fixed Effects Difference-in-Differences Regression
The fixed-effect full difference-in-difference regression specification from equation 2
reveals8 that, in the case of the implied contract doctrine, policy adoption is associated with
increases in aggregate private sector union membership and coverage rates of 0.84% and 1.03%,
as seen in column 3 of the first rows of Panels A and D of Table 2. The increase in union
coverage rate appears largely driven by a respective increase in the level of union membership
7 A closer investigation reveals large and disproportionate changes among the five sustained good faith doctrine
adoptions over the five years 1983 to 1988 surrounding good faith doctrine adoption in 1985 in Arizona, with a
nearly 30percent decline in union membership and coverage levels and a roughly 20percent increase in
nonmembership and noncoverage levels. Perhaps helping to explain this result, from 1983 to 1986 in Arizona the
copper industry experienced an historic strike of miners and mill workers that resulted in the replacement of nearly
all strikers and the decertification of their unions (Kingsolver 1989).
8 Sample statistics tables containing the mean, standard deviation, and number of observations for all fixed effects
regression samples of union outcome variables were constructed but omitted from this paper to save space. They are
available from author upon request. These tables are divided by subsamples for the implied contract, public policy,
and good faith doctrine combined treatment and control samples. The sample means for all outcome variables
across all three doctrine regressions approximate the national average levels of union membership and coverage and
union density over the years studied.
19
and coverage of 5.37% and 6.64%, visible in column 6 of the first row of Panels B and E of
Table 2. While there is evidence of a marginally significant 11.07% decline in private sector
union coverage levels associated with adoption of the good faith doctrine, in column 6 of the last
row of Panel E of Table 2, this does not exhibit any significant impact on union density, and is
not reflected in a statistically significant decline in union membership in column 6 of the last row
of Panel B of Table 2. As noted previously, these large declines could be at least in part related
to the disproportionately large declines in union membership and coverage levels in Arizona,
though the majority of the six good faith doctrine treatment states exhibit union decline well
above the national average in the years surrounding policy adoption.9
Turning to state industry groups in Table 3, analysis reveals that the implied contract doctrine
is associated with modest but statistically significant increases in manufacturing union
membership and coverage rates of 1.39% and 1.59 %, visible in columns 1 and 6 of the first row
of Panels A and B. These findings appear largely driven by a highly significant increase in the
level of manufacturing union membership and coverage of 8.06% and 9.15%, shown in columns
1 and 6 of the first row of Panels C and D. In columns 2 and 7 of the first row of Panels A and
B, one sees that adoption of the implied contract doctrine is associated with 0.38% and 0.68%
increases in nonmanufacturing union membership and coverage rates. Among the three major
9 If this high profile instance of union membership and coverage level decline and nonmembership and noncoverage
level growth does not in itself account for the disproportionate impacts for Arizona as a whole, the resolution of
such a large and public labor dispute could nonetheless set the tone of union membership development throughout
the Arizona labor market over these years. Delaware, New Hampshire, and Wyoming also exhibited union
membership and coverage level declines in the range of 10percent to 20percent per year over the five years
surrounding good faith doctrine adoption in each state, well above the national average for those years. On the other
hand, Idaho and Nevada, the remaining two good faith adopting states within the 1985 to 2011 sample window,
exhibit union decline at about the national average and slightly below the national average, respectively, over the
five years surrounding good faith doctrine adoption. These above average declines in other states might be
explained by an uptick in plant closings in Delaware and New Hampshire, and employment fluctuations connected
to swings in mineral export prices and long-run mechanization in extractive industries in Wyoming. Nevada’s
slower union decline could be explained by the growth of a highly organized service sector in Las Vegas over this
period.
20
industrial subdivisions of nonmanufacturing listed in Table 3 (i.e. construction, transportation-
communication-utilities, and service industries), construction industry union density increases
seem to drive the overall increase in nonmanufacturing union density, with a 2.76% and 2.3%
increase in union membership and coverage rates, respectively, visible in columns 3 and 9 of the
first row of Panels A and B. This result itself, unlike the situation for manufacturing industries,
appears driven not by statistically significant increases in the level of union membership and
coverage, but by 9.73% and 9.3% declines in the level of nonmember and noncovered
employment in columns 3 and 9 of the first row of Panels E and F of Table 3. While seeming to
show evidence of standard disemployment effects as demonstrated by Autor et. al. (2006),
among others, this result is not obvious since there is no reason one would expect nonmember
and noncovered employment within the construction industry to be composed of a higher
percentage of workers with less attachment to the labor market than, say, the service industry,
which does not exhibit comparable declines.
The impact of the good faith doctrine adoption may be reflected in the significant 12.6% and
9.83% respective declines in manufacturing union membership and coverage levels, shown in
columns 1 and 6 of the last row of Panel C and D of Table 3. These large declines could at least
in part be related to the outsized role of high profile union membership declines in Arizona,
Wyoming, Delaware, and New Hampshire, attributable to the small sample size and events
confounded with the year of policy adoption.
Analysis of the eight sex-age-educational attainment demographic groups in Tables 4 and 5
suggests that the implied contract doctrine is associated with a roughly 1.91% and 1.97% highly
significant increases in union membership and coverage rates for young male workers with less-
than-college education, as shown in the column 1 of the first row of Panel A in both tables. This
21
growth appears driven by a large and statistically significant roughly 11.34 and 11.38% increase
in the level of union membership and coverage for both young less educated males, as seen in
column 1 of the first row of Panel B in Tables 4 and 5. In the case of young females with less
than college education, union membership and coverage rates also increase by 0.74% and 1.14%
following adoption of the implied contract doctrine, as visible in column 5 of the first row of
both tables. However, only in the case of union coverage is it clear that this increase in density is
driven by a marginally significant 10.19% increase. This discrepancy between young less
educated female union membership and coverage level growth could suggest that, for certain
demographic groups, employment protection policies might increase free-riding behavior in
terms of increased utilization of union coverage without increased participation in union
membership participation. One could imagine that groups which have the most time
commitments outside of work, such as women who often bear the bulk of parenting
responsibilities, might be more inclined to free-ride, out of sheer necessity, on the benefits of
active union membership without participating themselves.
The implied contract doctrine also appears associated with a sizeable and marginally
significant 11.16 %and 10.13% decline in college educated older female membership and
coverage levels, visible in column 8 of the first row of Panel B in Tables 4 and 5, and a 8.82%
and 9.30% decline in college educated older female nonmember and noncovered employment
levels. However, this result is somewhat perplexing given that no significant or similarly-sized
disemployment effect emerges for less-educated younger and older females, whether union
members and covered or nonmembers and noncovered, who one might expect to have even
lower attachment to the labor force, and because the decline is greater for union members and
covered workers than for nonmember and noncovered workers, who one might expect to be
22
more precarious. Finally, the good faith doctrine adoption appears associated with a reallocation
of employment away from younger males and females with less than college education, whether
unionized or nonunionized, and toward older male college education nonmember and
noncovered employment. As noted previously, these declines could be confounded by
employment trends in the small sample of largely nonrepresentative states.10
VI. Conclusion
The implied contract doctrine does appear to increase unionization, and this growth seems to
occur in those (i.e. private), industries (i.e. manufacturing), and demographic groups (i.e. young
men and women with less-than-college education) where unions have both a strong base, as well
as potential to grow. Younger workers, especially younger women, without college education
have been underrepresented compared to older and more educated workers. While having the
most to gain from union membership, they are often too precarious to take the risks necessary to
form or join unions, or to secure a first collective bargaining agreement. Policies which lower
the costs to seeking union membership and union coverage could nudge these workers into
action, boosting union membership and coverage outcomes for themselves and the industries and
sectors in which they reside.
Past economic research has linked employment protection to increased labor productivity and
worker morale, as well as to slowed employment and job flows, small but consistent declines in
employment, and increased use of temporary workers. These new findings demonstrate, at least
10
One can imagine that such steep union decline in the highly organized manufacturing industries in Delaware and
New Hampshire, in which younger male and female workers, union and nonunion alike, predominate, along with
disorganization of the heavily unionized copper sector in Arizona, could spur erosion of union membership and
coverage across demographic groups, while paving the way for expansion of employment of older college educated
male nonmember and noncovered workers, a demographic who would be likely to staff new positions within
nonunion mineral extraction and processing industries.
23
in the context of the U.S., that wrongful discharge doctrines may also promote a main vehicle
workers have for voice and power at their jobs, union representation and collective bargaining.
At the very least, one could argue that these policies, passed during a period of steep union
decline and deindustrialization in the United States, helped stem losses in union membership and
coverage. Given current debates regarding the persistent and growing inequality following the
Great Recession, employment protection and unions could be a mutually reinforcing set of tools
to foster growth and fairness in labor markets and the economy at large.
24
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Muhl, Charles J. 2001 “The Employment-At-Will Doctrine: Three Major Exceptions.”
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27
28
29
Figure 1. --- Union Membership and Coverage Rates for Male Workers Aged 16 to 39 With Less
Than College Education Before and After Adoption of Wrongful Discharge Laws: Yearly Leads
From 2 Years Before to 3 Years After Adoption
Panel A. --- Implied Contract
100 x (Members/Employment) 100 x (Covered/Employment)
Panel B. --- Public Policy
100 x (Members/Employment) 100 x (Covered/Employment)
Panel C. --- Good Faith
100 x (Members/Employment) 100 x (Covered/Employment)
Note: Each figure represents a separate weighted OLS baseline difference and differences regression with two years of leads and lags in which the dependent variable is 100 times the number of
male less than college educated employed wage and salary workers between 16 and 39 years of age who are union members or covered by collective bargaining agreements over all male less
than college educated aged 16 to 39 employed wage and salary workers in 50 U.S. states. The number of union members, covered workers, and total wage and salary employees were constructed
from Current Population Survey Outgoing Rotation Group monthly earnings files for the years 1983 to 2014. All models include state main effects and indicators for each year in the sample, as
well as interactions between four Census-region dummies and calendar year dummies. Models are weighted by state’s share of national population aged 16-64 in each year using CPS earnings
weights. Huber-White robust standard errors were used to allow for unrestricted error correlations across observations within states.
-3
-2
-1
0
1
2
3
-2 -1 0 1 2
Point Estimate
Robust 90 Percent Confidence Interval
Year Relative to Adoption -3
-2
-1
0
1
2
3
-2 -1 0 1 2
Point Estimate
Robust 90 Percent Confidence Interval
Year Relative to Adoption
-3
-2
-1
0
1
2
3
-2 -1 0 1 2
Point Estimate
Robust 90 Percent Confidence Interval
Year Relative to Adoption -3
-2
-1
0
1
2
3
-2 -1 0 1 2
Point Estimate
Robust 90 Percent Confidence Interval
Year Relative to Adoption
-3
-2
-1
0
1
2
3
-2 -1 0 1 2
Point Estimate
Robust 90 Percent Confidence Interval
Year Relative to Adoption
-6
-4
-2
0
2
4
6
-2 -1 0 1 2
Point Estimate
Robust 90 Percent Confidence Interval
Year Relative to Adoption
30
TABLE 1. --- Recognition of Wrongful Discharge Laws by State as of December 31, 2014 and
Closest Calendar Year to Date of Policy Adoption and Repeal
State Implied Contact Y/N Adopt Repeal
Public Policy Y/N Adopt Repeal
Good Faith Y/N Adopt Repeal Readopt
Alabama………………….
Alaska………………….....
Arizona…………………..
Arkansas………………….
California…………………
Colorado…………………
Connecticut……………....
Delaware………………….
Florida……………………
Georgia…………………..
Hawaii…………………...
Idaho……………………..
Illinois……………………
Indiana…………………..
Iowa………………………
Kansas……………………
Kentucky…………………
Louisiana…………………
Maine…………………….
Maryland…………………
Massachusetts…………….
Michigan………………….
Minnesota…………………
Mississippi………………..
Missouri………………….
Montana…………………..
Nebraska………………….
Nevada……………………
New Hampshire…………..
New Jersey……………….
New Mexico………………
New York…………………
North Carolina……………
North Dakota……………..
Ohio………………………
Oklahoma…………………
Oregon……………………
Pennsylvania……………..
Rhode Island………………
South Carolina……………
South Dakota……………..
Tennessee…………………
Texas……………………..
Utah……………………….
Vermont…………………..
Virginia……………………
Washington……………….
West Virginia……………...
Wisconsin…………………
Wyoming………………….
Total
Y 1988
Y 1983
Y 1983
Y 1984
Y 1972
Y 1984
Y 1986
N
N
N
Y 1987
Y 1977
Y 1975
Y 1988
Y 1988
Y 1985
Y 1984
N
Y 1978
Y 1985
Y 1988
Y 1980
Y 1983
Y 1992
N 1983 1988
Y 1987
Y 1984
Y 1984
Y 1989
Y 1985
Y 1980
Y 1983
N
Y 1984
Y 1982
Y 1977
Y 1978
N
N
Y 1987
Y 1983
Y 1982
Y 1985
Y 1986
Y 1986
Y 1984
Y 1978
Y 1986
Y 1985
Y 1986
42 43 1
N
Y 1986
Y 1985
Y 1980
Y 1960
Y 1986
Y 1980
Y 1992
N
N
Y 1983
Y 1977
Y 1979
Y 1973
Y 1986
Y 1981
Y 1984
N
N
Y 1982
Y 1980
Y 1976
Y 1987
Y 1988
Y 1986
Y 1980
Y 1988
Y 1984
Y 1974
Y 1981
Y 1984
N
Y 1985
Y 1988
Y 1990
Y 1989
Y 1975
Y 1974
N
Y 1986
Y 1989
Y 1985
Y 1984
Y 1989
Y 1987
Y 1985
Y 1985
Y 1979
Y 1980
Y 1990
43 43 0
N
Y 1983
Y 1985
N
Y 1981
N
Y 1980
Y 1992
N
N
N
Y 1990
N
N
N
N
N
N
N
N
Y 1978
N
N
N
N
Y 1982
N
Y 1987
Y 1974 1981 1990
N
N
N
N
N
N
N 1985 1989
N
N
N
N
N
N
N
N
N
N
N
N
N
Y 1994
11 12 2 1 Notes: Based on author’s analysis of case law and secondary literature. “Y” denotes that a policy is recognized and “N” denotes that it is not recognized by a state as of December 31, 2014.
Adoptions and repeal dates are rounded to the calendar year in which they occurred if the first full month following policy change was from January to May of that year, and are rounded forward
to the following calendar year if the first full month following policy change is from June to December of the original year.
31
TABLE 2. --- Simplified Illustration of the Legal Policy Variables for the Full Fixed Effects
Differences in Differences Regression Using the Adoption of the Implied Contract Doctrine in
Massachusetts in 1988 for Treatment Observations and Rhode Island and Maine for Control
Observations
Panel A. --- Massachusetts (Adopts the Implied Contract Doctrine in 1988)
Legal Policy
Variables 1985 1986 1987 1988 1989 1990 1991 1992
Treatst 0 1 1 1 1 1 0 0
Postst 0 0 0 1 1 1 1 1
Treatst Postst 0 0 0 1 1 1 0 0
Postst Postst 0 0 0 0 0 0 1 1
Panel B. --- Rhode Island (Never Adopts A Wrongful Discharge Doctrine)
Legal Policy
Variables 1985 1986 1987 1988 1989 1990 1991 1992
Treatst 0 0 0 0 0 0 0 0
Postst 0 0 0 1 1 1 1 1
Treatst Postst 0 0 0 0 0 0 0 0
Postst Postst 0 0 0 0 0 0 0 0
Panel C. --- Maine (Only Adopted the Implied Contract Doctrine in 1978)
Legal Policy
Variables 1985 1986 1987 1988 1989 1990 1991 1992
Treatst 0 0 0 0 0 0 0 0
Postst 0 0 0 1 1 1 1 1
Treatst Postst 0 0 0 0 0 0 0 0
Postst Postst 1 1 1 1 1 1 1 1 Note: Constructed from author’s analysis of Autor et. al. (2003). Treatst is an indicator for the period 2 years before to 2 years after the year of adoption of a wrongful-discharge law in
treatment state s. Postst is an indicator for the period from the year of adoption11 to 2 years after the year of adoption for all states, both adopting and nonadopting. TreatstPostst is an
interaction term representing the three year period following policy adoption, in the treatment state only, beginning with the first year of adoption. The coefficient of interest β2 is an estimate of
the pre-post change in the outcome variable in adopting states relative to the corresponding change in non-adopting states. Postpostst is a dummy variable that turns on for each state that
reenters the control group 3 years after the year of policy adoption to account for any bias caused by the enduring effect of the policy adoption after the five-year treatment window expires.
32
TABLE 2. --- Difference-In-Differences Estimates of the Impact of Wrongful Discharge Laws on
State Union Membership and Coverage Outcomes by Sector: Contrasting Outcomes in Years 2
and 3 Following Adoption with Years 1 and 2 Preceding Adoption: 1983-2014
A. Membership Rates B. Membership Levels C. Nonmembership Levels
All Public Private All Public Private All Public Private Doctrines (1) (2) (3) (4) (5) (6) (7) (8) (9)
Public 0.23 -0.37 0.24 -0.84 -1.71 -0.15 -1.66 -0.55 -1.69
Policy
R-Sq
(0.32)
0.99
(0.94)
0.98
(0.32)
0.98
(3.20)
1.00
(4.76)
0.99
(3.71)
1.00
(1.90)
1.00
(2.28)
0.99
(1.94)
1.00
N 465 465 465
Good 0.09 -0.39 -0.26 -3.50 7.38 -11.07+ -2.14 11.10* -4.35
Faith
R-Sq
(0.62)
0.99
(1.16)
0.98
(0.45)
0.98
(6.14)
1.00
(7.85)
0.99
(5.91)
1.00
(3.35)
1.00
(4.65)
0.99
(4.28)
1.00
N 493 493 493 Note: Each figure represents a separate weighted OLS fixed-effects difference and differences regression in which the dependent variable is 100 times the number of employed wage and salary
workers between 16 and 64 years of age in a givens sector who are union members or covered by collective bargaining agreements over all employed wage and salary workers between the ages
of 16 and 64 in a given sector, 100 times the natural log of the number of employed wage and salary workers between 16 and 64 years of age in a given sector who are union members or covered
by collective bargaining agreements over all employed wage and salary workers between the ages of 16 and 64 in a given sector, or 100 times the natural log of the number of employed wage
and salary workers between the ages of 16 and 64 in a given sector who are not union members nor covered by collective bargaining agreements over all employed wage and salary workers
between the ages of 16 and 64 in a given sector in 50 U.S. states. The number of union members, covered workers, nonmembers, noncovered workers, and total wage and salary workers in a
given sector is constructed from Current Population Survey Outgoing Rotation Group monthly earnings files for the years 1983 to 2014. All models include state main effects and indicators for
each year in the sample, as well as interactions between four Census-region dummies and calendar year dummies. Models are weighted by state’s share of national population aged 16-64 in each
year using CPS earnings weights. Huber-White robust standard errors were used to allow for unrestricted error correlations across observations within states.
33
TABLE 3. --- Difference-In-Differences Estimates of the Impact of Wrongful Discharge Laws on State
Union Membership and Coverage Outcomes by Industry: Contrasting Outcomes in Years 2 and 3
Following Adoption with Years 1 and 2 Preceding Adoption: 1983-2014
Panel A.--- Membership Rates Panel B. --- Coverage Rates
N 493 493 Note: Each figure represents a separate weighted OLS fixed-effects difference and differences regression defined identically to those in Table 2 with the exception that investigation is conducted for selected
industries.
34
TABLE 4. --- Difference-In-Differences Estimates of the Impact of Wrongful Discharge Laws on State
Union Membership Outcomes by Sex, Age, and Education: Contrasting Outcomes in Years 2 and 3
Following Adoption with Years 1 and 2 Preceding Adoption: 1983-2014
Good -8.12* -4.81 -1.28 8.97* -7.23** 5.53 -1.60 0.05 Faith R-sq N
(3.49) 1.00
(6.95) 0.99
(7.46) 1.00
(3.41) 0.99 493
(2.49) 1.00
(5.50) 0.99
(5.17) 1.00
(6.34) 0.99
Note: Each figure represents a separate weighted OLS fixed-effects difference and differences regression on union membership outcomes defined identically to those in Table 2, though
investigated for given each of eight age-sex-education demographic category. MHY=males with less than college education between the ages of 16 and 39, MHO=males with less than college
education between the ages of 40 and 64, MCY=males with college or more education between the ages of 16 and 39, MCO=males with more than college or more education between the ages of
40 and 64, FHY=females with less than college education between the ages of 16 and 39, FHO=females with less than college education between the ages of 40 and 64, FCY=females with
college or more education between the ages of 16 and 39, FCO=females with college or more education between the ages of 40 and 64.
35
TABLE 5. --- Difference-In-Differences Estimates of the Impact of Wrongful Discharge Laws on State
Union Coverage Outcomes by Sex, Age, and Education: Contrasting Outcomes in Years 2 and 3
Following Adoption with Years 1 and 2 Preceding Adoption: 1983-2014
Note: Each figure represents a separate weighted OLS fixed-effects difference and differences regression on union coverag outcomes defined identically to those in Table 2, though investigated
for given each of eight age-sex-education demographic category. MHY=males with less than college education between the ages of 16 and 39, MHO=males with less than college education
between the ages of 40 and 64, MCY=males with college or more education between the ages of 16 and 39, MCO=males with more than college or more education between the ages of 40 and
64, FHY=females with less than college education between the ages of 16 and 39, FHO=females with less than college education between the ages of 40 and 64, FCY=females with college or
more education between the ages of 16 and 39, FCO=females with college or more education between the ages of 40 and 64.