Partisanship, Christianity, and Women in the Legislature: Determinants of Parental Leave Policy in...

Partisanship, Christianity, and Womenin the Legislature: Determinants ofParental Leave Policy in U.S. States∗

Sarah Williamson, Georgetown University

Matthew Carnes, Georgetown University

Objective. Although the United States 1993 Family and Medical Leave Act is consid-ered meager by international standards, significant variation in family leave policiesexists across U.S. states. This article develops a political theory—driven by mobilizedinterest groups—to explain variation in the duration and pay level of state parentalleave policies. Methods. Employing four different measures of family leave policy, weuse ordinary least squares and logit models to test the effects of partisanship, womenin the legislature, and evangelical populations on parental leave outcomes. Results.We find that states with a pattern of Democratic Party controlled legislatures andhigh percentages of legislative seats occupied by women see more generous parentalleave protections, while states with large populations of evangelical Christians see lessgenerous policies. Conclusion. Family leave policies are the product of a battle be-tween competing visions of the family and the state, shaped by partisanship, gender,and religion.

Parental leave protections in the United States are extremely weak by inter-national standards. In early 2011, Human Rights Watch released a scathingreport entitled “Failing its Families,” which detailed “America’s deficient work-family policies” and indicted them as “a human rights concern” (HRW,2011:5). In discussing the report, The Washington Post called the UnitedStates “an outlier” and its family policy “an embarrassment” (Crary, 2011).

However, these harsh criticisms obscure as much as they reveal. Parentalleave policy varies enormously across the United States. On the low end of thespectrum, 30 states earn the low international ranking highlighted by HumanRights Watch, providing no protection for private-sector employees beyondthe internationally meager guarantees of the Family and Medical Leave Act(FMLA) of 1993. Yet as of 2005, 20 states provided some additional protec-tions without pay, and California stood out as the only state that provided

∗Direct correspondence to Matthew Carnes, Department of Government, GeorgetownUniversity, Intercultural Center 681, Washington, DC 20057 〈[email protected]〉. Aprevious version of this article was presented at the annual meeting of the Midwest PoliticalScience Association, March–April 2011. We acknowledge the helpful comments of AndrewKarch, Michelle Swers, Clyde Wilcox, and two anonymous reviewers. We will share all dataand coding with those wishing to replicate the study.

SOCIAL SCIENCE QUARTERLYC© 2012 by the Southwestern Social Science AssociationDOI: 10.1111/j.1540-6237.2012.00882.x

2 Social Science Quarterly

paid leave. Since then, two additional states, Washington and New Jersey, havepassed paid family leave laws, but Washington’s policy remains unfunded and,at the time of writing, has yet to go into effect. What explains this strikingvariation in family leave provisions across states?

Despite observable differences in parental leave benefits across states, ex-isting literature on family policy focuses almost exclusively on childcare, anddiscussion of parental leave has been largely absent from both domestic andcomparative literature. This article investigates the causes of divergence amongparental leave policies in U.S. states. It advances a political theory based onthe influence of partisanship, evangelical Christianity, and women’s participa-tion in state legislatures, which is tested on a data set including all 50 states.Patterns of Democratic Party control and higher percentages of women instate legislatures are found to increase parental leave protections, while largeevangelical Christian populations have a dampening effect on the generosityof policies.

This article begins with an exploration of the variation in family leavepolicies across the United States, highlighting the major features of existingfederal legislation and the considerable departures from it in a subset of cases.Next, we articulate a political theory of family leave protection, drawn fromthe larger welfare state literature but adapted to the particular context of theUnited States and the specific features of family leave policy. Third, we test ourhypotheses empirically using a data set on family leave policy provisions acrossstates, and undertake a series of robustness checks using alternative methodsand measures of family leave policy. Finally, we conclude by suggesting theimplications of our work for future development of family leave policy in theUnited States, and for study of the welfare state more generally.

Parental Leave in the United States

Federal parental leave policy in the United States provides for 12 weeksof unpaid leave. This places the United States at the extreme low end ofthe international family policy spectrum. In comparison, Germany offers 14weeks of paid leave; the United Kingdom provides 52 weeks of leave, with39 weeks paid leave; and Sweden guarantees a world-leading 480 days of paidleave, shared between both parents.

Two laws comprise the entirety of the U.S. family leave policy. First, thePregnancy Discrimination Act (PDA) of 1978 requires that employers treatpregnant women as they would treat any other sick or disabled employee,specifically in the use of sick days and disability benefits (Grant, Hatcher, andPatel, 2005). The law prohibits firing and refusing to hire or promote a womanbecause of her pregnancy, but does not mandate job protection during timetaken off to care for a new child (Grant, Hatcher, and Patel, 2005). Second,the FMLA of 1993 guarantees job protection by mandating 12 weeks of

Determinants of Parental Leave Policy in U.S. States 3

FIGURE 1

Parental Leave Policies Across U.S. States

unpaid parental leave for eligible women and men (Grant, Hatcher, and Patel,2005).1

Of the 50 states, 30 provide only the FMLA-mandated minimum of 12weeks’ protected parental leave. However, 20 states have passed additionallegislation, either prior to or following the FMLA, that guarantees significantlymore protection. These measures range from longer job-protected leave, tosubsidies for at-home infant care for low-income parents, to provisions forthe use of accrued paid leave (vacation or sick days) at the birth of a child,and, in a few cases, paid parental leave. Figure 1 depicts this variation acrossstates, employing data adapted from the National Partnership of Women andFamilies (Grant, Hatcher, and Patel, 2005).2 States that provide no coveragebeyond the FMLA minimum are scored as zero. The remaining states clusterat three score levels, with one group around 20 points (including Iowa, NewYork, Tennessee, and Wisconsin), another around 40 to 45 points (includingMassachusetts and Rhode Island), and a third group around 60 points (Hawaii,Washington, and Oregon).

1The FMLA’s stringent eligibility requirements and limited coverage—which demand thata parent have worked for the same employer for at least one year and have worked at least1,250 hours during the year previous to taking leave, and that apply the law only to employerswith 50 or more employees—effectively exclude roughly half of all U.S. families (Waldfogel,2001).

2As we explain in the section entitled Empirical Testing, we have modified the NPWFparental leave data to include only provisions that apply to the private-sector employees.


Three states stand out as leaders. California was the first in the countryto guarantee paid parental leave. It provides six weeks of partial (55 percent)pay replacement (capped at $987), funded through payroll taxes, to care for anewborn child, a newly adopted child, or a newly placed foster child, as wellas an ill family member, including a partner who is disabled by or recoveringfrom pregnancy (Grant, Hatcher, and Patel, 2005). The Washington StateFamily Leave Act provides a constant $250 per week for up to five weeks fornew parents; it was passed in May 2007, but at the time of writing had yetto come into effect due to lack of a funding mechanism. New Jersey’s FamilyLeave Insurance, passed in 2007 and implemented in 2009, provides six weeksof two-thirds wage replacement (capped at $554).

Seventeen remaining states offer policies beyond the FMLA but have notreached the level of providing paid leave. New Mexico, the lowest rankingof this middle group, expands upon the FMLA by offering an At-HomeInfant Care program that aids low-income parents staying at home in orderto care for their young children (Grant, Hatcher, and Patel, 2005). The mostgenerous of the mid-range states is Oregon, which mandates 12 weeks ofjob-protected leave to care for a spouse or recover from one’s own healthcondition, including pregnancy, as well as 12 weeks to care for a new childwith an additional 12 weeks to care for an ill child. The remaining states fallbetween these two models, offering a variety of expansions on the FMLA,such as broader coverage or more flexible uses for leave.

Theory: Political Determinants of Family Leave Policy

We build on previous academic studies of welfare states to develop a po-litical theory of family leave policies in the United States, highlighting thepolitical conflict between social interests in society (Esping-Andersen, 1990;Wennemo, 1992; Fleischman and Moyer, 2009). We treat individual U.S.states as our unit of analysis. By comparing the U.S. states, we control forshared national-level political and economic institutions, and examine state-level political factors in order to account for the observable divergence inparental leave policy. We focus on three key independent variables: Demo-cratic Party power, the influence of conservative evangelical Christianity, andthe share of legislative seats held by women.

Partisanship

First, we argue that partisanship may shape variation in family leave policy.In international comparisons, leftist parties have been consistently associatedwith expansion of social welfare spending and services (Huber and Stevens,2001). Importantly, this effect is most pronounced when leftist parties govern


for a sustained period of time; short stints in government are not associatedwith significant policy differences.

Literature specific to the United States has reached mixed results on the im-pact of partisanship on state political outcomes. Erikson, Wright, and McIver(1989) find an apparent lack of correlation between the relative strength ofstate parties and the policies they enact. They argue that this disconnect oc-curs because both major parties respond to public opinion within the stateand moderate themselves accordingly (Erikson, Wright, and McIver, 1989).However, many recent studies have shown significant effects of partisanshipon substantive welfare outcomes, such as the percentage of citizens attainingaccess to insurance (Cummins, 2011). Of particular importance to our study,Brown (1995) finds that Democratic Party control of state governments leadsto more generous welfare provisions, at least in the subset of cases where partyand class identification overlap. Thus, we formulate our first hypothesis (H1):in states with a pattern of consistent Democratic Party control of government,parental leave policy will be more generous than in states without such a pattern.

Evangelical Christians

Cultural norms and social values can also affect parental leave policy. TheChristian Democratic parties of Western Europe—motivated by traditionalChristian values—discourage policies that facilitate mothers working outsideof the home (Huber and Stevens, 2001). In the United States, evangelicalChristians, who place particular emphasis on personal conversion and reli-gion’s role in the public sphere, hold a similarly strict view of the family;they seek to have this view preserved in law for both moral and materialreasons. First, on moral grounds, they argue that family leave policies thatmake it easy for women to work outside the home negatively impact both thefamily and American society. Second, they oppose the expansion of familypolicies because they themselves will derive no material benefit from them(since evangelical mothers are expected to work in the home).

While evangelical Christians do not constitute a political party in the UnitedStates, they do form a significant voting bloc that is “more conservativethan other Americans” (Wilcox, 1990). Seventy-six percent of evangelicalChristians report that religion is important or very important in their livesand has a strong connection to their political opinions (Green, 2007). Thesereligious communities have a significant effect on the social and political viewsof their adherents; recent work has shown that evangelicals display significantopposition to state sanctioning of contentious public issues, including theteaching of evolution (Haider-Markel and Joslyn, 2008) and the legalizationof same-sex marriage (Fleischmann and Moyer, 2009).

Thus, through their strict views of family and the saliency of their religious-political views, evangelical Christians have the potential to limit progressiveparental leave policies. This leads to our second hypothesis (H2): in states


with large populations of evangelical Christians, parental leave policy will be lessgenerous than in states without large evangelical populations.

Women in State Legislatures

Family leave policy may expand due to the direct action of women aslegislators. Survey data indicate that American women care more about theissues of paid family leave and paid sick days than any other issue, including theeconomy (Appelbaum and Milkman, 2009). As members of state legislatures,we expect that women will carry similar preferences into their legislative actionsand promote family leave policies. Past literature on women in the legislaturehas produced mixed results but finds that women legislators tend to be moreliberal than men, and that this gender gap is most significant for Republicanwomen (Swers, 1998).

Swers (2002) has shown that female legislators are more likely than theirmale counterparts to support women’s issues in a variety of ways. They moreoften participate in floor debates, advocate the inclusion of women’s issues incommittee legislation, and sponsor and co-sponsor legislation that particu-larly affects women. These differences are especially apparent around “issuesthat reflect women’s traditional role as caregiver” (Swers, 2001:218). Thomas(1991) finds that such legislation is particularly likely to pass in states in whichwomen comprise more than 20 percent of the legislature.

This evidence of the impact of female legislators on women-, child-, andfamily-related legislation leads to the third hypothesis (H3): in states withgreater proportions of women in the state legislatures, parental leave policy will bemore generous than in those with lower proportions.

Empirical Testing

We employ a series of econometric models to examine the relationshipbetween our hypothesized political variables and the variation in parentalleave policy across states.

Variables

Dependent Variable. Given the complexity and multidimensionality of fam-ily leave policies, we choose to draw our main measure of the generosity ofparental leave protections from the National Partnership for Women and Fam-ilies (NPWF)’s 2005 publication, “Expecting Better: A State-by-State Analysisof Family Leave Programs.” This publication presents an index that quantifieseach state’s family leave programs by awarding points based on the presence ofvarious types of parental leave protections in 2005. The NPWF bases its scor-ing on benefits afforded to employees of both the private and public sectors,


but only a very few states provide additional benefits to the latter. In order toensure the comparability of scores across the states, and to avoid possibly con-founding factors in the politics of public-sector employment, we adjusted thefamily leave scores to include only benefits offered to private-sector workers.3

Independent Variables. The strength of the Democratic Party was mea-sured by the total number of years between 1991 and 2005 that the partycontrolled both houses of the state legislature. Huber and Stevens (2001) sug-gest that a party must have a history or pattern of legislative dominance inorder to influence the direction of welfare policy, so we take the 15-year figureto represent the achievement (or lack) of such dominance.4 We choose notto analyze the partisanship of governors because they lack legislative author-ity, and because earlier work reveals that “partisan control of governorshipsover the long term bears little relationship to other indicators of partisanship”(Erikson, Wright, and McIver, 1989). The data were drawn from Gerber andHopkins (2011).

The size of the evangelical Christian population as a percentage of totalpopulation in each state is drawn from John C. Green (2007).5 The per-centage of women in state legislatures is collected over the same time frameas Democratic Party control, from 1991–2005. The data are taken from theCenter for American Women and Politics, part of the Eagleton Institute ofPolitics at Rutgers, The State University of New Jersey. Summary statisticsfor all of the variables employed in the econometric analysis, including thecontrols introduced below, are presented in the Appendix.

Models of Family Leave Policy Variation

We present a series of ordinary least squares (OLS) regression models belowto test the effect of Democratic Party control (H1), the size of the evangelicalChristian population (H2), and the percentage of women in the state legisla-ture (H3) on our main measure of family leave policy. Table 1 presents thesemodels.

Model 1 is our base model, testing only our three principal explanatoryvariables. All three of our key hypotheses—concerning Democratic Party

3Additional testing, not reported below, finds that the raw-NPWF scores and our adjustedprivate-sector scores behave similarly and produce nearly identical results.

4With its unicameral, nonpartisan state legislature, Nebraska is coded as lacking DemocraticParty control.

5Collecting data on church membership is extremely complicated, since denominationshave different standards for what constitutes membership and categorizing denominationswithin religious traditions can be the subject of theological controversy. We employ datadrawn from Green (2007) due to his focus on the influence of faith in politics. His dataset is constructed from a variety of sources, including the American Religious Landscapestudies, the Pew Research Center, the National Election Studies, the General Social Survey, the2000 Religious Congregations and Membership Survey, the American Religious IdentificationSurvey, and other state-level surveys.


TABLE 1

Determinants of Family Leave Policy

(1) (2) (3) (4) (5)Basic Legislative Economic Education Ideology

Variables Model Professional Controls Controls Controls

Years Democ. 2.098∗∗∗ 2.128∗∗∗ 2.001∗∗∗ 2.325∗∗∗ 2.140∗∗∗

(0.513) (0.503) (0.545) (0.633) (0.552)Percent

evangelical−1.002∗∗∗ −0.811∗∗ −0.925∗∗ −1.030∗∗ −0.931∗∗

(0.327) (0.341) (0.404) (0.397) (0.407)Women in

legislature1.073∗∗ 1.133∗∗ 1.002∗∗ 0.902 0.946(0.492) (0.484) (0.491) (0.615) (0.586)

Log GSP percapita

14.63(50.78)

Unemployment 1.324(3.307)

Union members 1.142∗

(0.646)Urban population −0.292

(0.281)Leg. profession. 43.28

(26.03)High school −0.318

(1.229)College degree 3.594∗

(1.974)Advanced

degree−4.432

(3.243)Moral 12.13

(8.870)Individual 5.288

(9.576)Traditional 2.172

(11.01)Constant 6.729 −7.490 −59.53 10.51 3.848

(16.52) (18.33) (220.4) (83.53) (19.76)Observations 50 50 50 50 50R-squared 0.415 0.449 0.482 0.462 0.441

OLS models. Standard errors in parentheses.∗∗∗p < 0.01, ∗∗p < 0.05, ∗p < 0.1.

control, evangelical Christian population, and women in the state legislature—are supported, displaying statistically significant effects in the expected di-rections. Thus, we find that greater Democratic dominance of, and largerpercentages of women in, the state legislature is correlated with more gener-ous state-level family leave policy; conversely, a larger evangelical Christianpopulation is correlated with less generous policy.

Next, we control for a variety of factors that may explain variation in familyleave policies. Given the small number of observations on our dependentvariable, we sequentially introduce sets of control variables to our base models.


Legislative Professionalism. Model 2 adds a measure of legislative pro-fessionalism to our analysis. Since the passage of new laws requires time andeffort by lawmakers, we expect that states that have legislators that meet regu-larly and have greater staff and other resources will be more likely to introduceand pass new laws, including family leave policies. We draw our measure oflegislative professionalism from Squire (2007). The results for our principalhypotheses remain unchanged, and while the coefficient for legislative pro-fessionalism is positive, as expected, it does not reach conventional levels ofstatistical significance.

Economic and Demographic Controls. Model 3 adds a series of economicand demographic controls. The first of these, logged gross state product (GSP)per capita, is included with the expectation that states that are wealthier mayprovide more expansive social policy measures. The second economic control,unemployment, may increase pressures on the state to diminish burdens onemployers (such as family leave benefits), in order to promote hiring. Ifthis is the case, we would expect that higher unemployment would decreasethe likelihood of passage of family leave benefits in a state. Alternatively,unemployment may have a positive effect on the generosity of parental leave,especially if voters begin to push for family leave as a means to job security.Fiscally constrained states may also see expanded family leave as a way tokeep parents employed (after their leave) rather than risk adding them to theunemployment rolls. We draw our average measure of unemployment fromthe U.S. Census Bureau, Statistical Abstract of the United States (2009).

We also include the level of unionization in the state, as the welfare stateliterature (Huber and Stephens, 2001; Esping-Anderson, 1990) has found astrong relationship between the presence of labor unions and the extensionof social policies, including family benefits, in cross-national samples. Wemeasure the level of unionization in each state by the percentage of employedworkers who are union members. The data are drawn from a 2002 data setcompiled by Barry T. Hirsch and David A. Macpherson, who in turn drawon the Current Population Survey (CPS) Outgoing Rotation Group EarningFiles.

Finally, we include the level of urbanization because Erikson, Wright, andMcIver (1993) find that states that are more urbanized are more likely to exhibitincreased spending. Urbanization, measured as the percent of the populationliving in urban areas, was taken from the 2000 Census of Population andHousing (U.S. Census Bureau, 2000).

As can be seen in Model 3, only one of these additional controls is statisticallysignificant at the 10 percent level—the level of union membership in the state.States that have greater unionization of their workforce have more generousfamily leave policies. Wealth, unemployment, and urbanization do not havesignificant effects. Our principal variables continue to be supported by thisanalysis.


Educational Controls. In Model 4, we introduce three measures of edu-cational attainment. Research on U.S. social policy has shown that states withhigher levels of education tend to be more liberal on social issues (Fleischmanand Moyer, 2009), and more generous in their social policies. We include dataon maximum educational attainment at three levels drawn from the 2011Statistical Abstract of the United States: high school, college, and advanceddegrees. In states where there are larger numbers of (exactly) college-educatedcitizens, family leave benefits are more generous. Interestingly, the pattern ofhigher benefits does not continue as education levels increase. In states withhigher numbers of advanced degree recipients, the coefficient on family leavebenefits is negative, but it does not reach conventional levels of statisticalsignificance.

Our principal findings are weakened by the inclusion of these educationalvariables, such that the percentage of women in the legislature is no longerstatistically significant. The coefficient retains its sign and value, but it is esti-mated less precisely. The other two hypotheses continue to produce statisticallysignificant results.

Ideology Controls. We introduce in Model 5 three indicators of state-level ideology. Ideology is the most frequently cited cause of state-level policyvariation, other than those we have already considered. We draw our measuresfrom Elazar (1972), who distinguishes three trends in American state-levelpolitical culture. Citizens in moralist states rely on the government to bean active, positive force in society and politics often revolves around socialissues. In states with an individualist political culture, the public generallyprefers government involvement only where absolutely necessary. In stateswith traditionalist political cultures, the government is seen as a vehicle formaintaining a hierarchical status quo.

Our results indicate that none of the state-level ideology measures has astatistically significant impact on family leave policy variation. Both our mea-sure of Democratic control of the legislature and evangelical population retaintheir values and significance in the presence of the state-level ideology mea-sures; however, women in the legislature narrowly misses achieving statisticalsignificance in this model.

Conclusion. The analysis in Table 1 provides strong support for our theo-retical argument that family leave benefits are a function of competing interestsbetween Democratically controlled legislatures, in which women play a sig-nificant role, on the one hand, and evangelical Christian populations, on theother.

The substantive effects of each of these variables are large. Using the Clarifystatistical package in Stata (Tomz, Wittenberg, and King, 2001), when holdingall other variables at their mean, we find that a one standard deviation increasein our years Democratic variable (from 6.52 to 12.6 years, out of 15 years)raises the family leave score by nearly one-half standard deviation—from the


TABLE 2

Robustness Checks

(6) (7)Variables OLS Tobit

1976–2005 1976–2005

Years Democratic 1.204∗∗∗ 2.303∗∗∗

(0.274) (0.653)Percent

evangelical−1.081∗∗∗ −1.961∗∗∗

(0.318) (0.687)Women in

legislature1.429∗∗ 2.776∗∗

(0.549) (1.180)Constant 2.060 −40.31

(15.50) (34.65)Observations 50 50 (30 observations

censored)R-squared 0.437 0.094

OLS and Tobit models. Standard errors in parentheses.∗∗∗p < 0.01, ∗∗p < 0.05, ∗p < 0.1.

mean score of 17.88 to 30.53. In Figure 1, this corresponds to moving astate’s score from that of Tennessee to nearly that of Montana. Similarly, a onestandard deviation increase in the evangelical population (from 25.33 to 35.67percent) corresponds to a drop of over 10 points to 7.48 points, or from thatof Massachusetts to Tennessee. Finally, a one standard deviation increase inpercentage of women in the legislature (from 21.32 to 28.22 percent) resultsin a one-third standard deviation increase, in the family leave score, to 25.32(roughly a move from Louisiana to Rhode Island).

Robustness Checks. Table 2 presents two additional robustness checkson our findings. Since many of the provisions that exceed the FMLA wereinstituted prior to 1990, we lengthen the timeframe over which we measuretwo of our key independent variables—years of Democratic control of thelegislature and women in the legislature—to 30 years, extending from 1976to 2005. Model 6 shows that our findings continue to be supported when thislonger timeframe is considered.6 The longer term impact of both Democraticcontrol and women in the legislature is positive and statistically significanton family leave benefits, and the percentage of evangelicals in the state has anegative impact.

In addition, because many states do not offer any benefits beyond theFMLA, there might be concern that our results are driven by the large numberof states that receive a score of zero in our adjusted-NPWF data set. To addressthis issue, Model 7 presents a Tobit model, which censors the observations

6Unfortunately, state-level data on evangelical church membership are not available for thisearlier period, so we must continue to use the Green data for 2000.


that scored zero and models the remaining observations, and again uses datadrawn from the longer 1976–2005 time period. The results are consistentwith all our previous findings.

Alternative Measures of Family Leave Policy Variation

Thus far, our analysis has used the NPWF-derived scores, a continuousmeasure, as a dependent variable. However, our significant findings couldbe an artifact of the construction of the NPWF scores, which allot varyingnumbers of points to states for each of their additional family leave policies.While this provides much finer grained variation for our analysis, there is adanger that it overstates some policies and undervalues others. For this reason,we conduct additional analyses on a series of binary codings of family leavepolicy, which simply consider whether or not a state has passed family leavepolicies of its own.

First, we examine the determinants of the presence or absence of familyleave policies in each state prior to the passage of the national FMLA. Thisallows us to explore the causes of family leave adoption prior to the federalintervention in the 1993 FMLA, which might have served to raise the salienceof family leave policy and shaped the kinds of proposals that were made.Furthermore, recent research has highlighted the possibility that states mayfunction as “the primary sites of [policy] innovation” (Kail, Quadagno, andDixon, 2009).

Second, we examine the factors contributing to variation in the policiesoffered in 2000 that extend beyond the provisions of the FMLA. These allowus to test how states responded in the presence of the FMLA as a focusingdevice.

Finally, we examine behavior during an interesting natural experiment: in1999, President Bill Clinton gave instructions to the Department of Laborto investigate allowing states to use unemployment insurance (UI) to fundpaid parental leave, which became known as “Baby UI.” States varied in theirreaction to this development, with some rapidly extending paid family leavebenefits and others refusing to implement any payments.

We test our three main hypotheses on these binary outcomes across thestates. We do not include other controls, due to the small number of ob-servations in our data set and the lack of contemporaneous data on somecovariates. The results are presented as three additional models in Table 3.Since the dependent variables in these models were measured in differentyears, we employ contemporaneous data on the independent variables: wemodel pre-FMLA policies (measured in 1993) with legislative data coveringthe period from 1976 to 1993, and we model FMLA-exceeding and “BabyUI” measures (both evaluated in 2000) with data from 1976 to 2000.


TABLE 3

Determinants of Non-FMLA Family Leave Measures

(8) (9) (10)Variables Policy Policy Introduction

Before Beyond of PaidFMLA (1993) FMLA Leave Legislation

by 2000 by 2000

Years Democratic 0.161∗∗ 0.111∗∗∗ 0.0764∗

(0.0680) (0.0433) (0.0443)Percent evangelical −0.0862∗ −0.0522 −0.0715

(0.0480) (0.0387) (0.0475)Women in legislature 0.164∗∗ 0.128∗ 0.188∗∗

(0.0774) (0.0664) (0.0750)Constant −3.423 −3.032 −3.641∗

(2.141) (1.930) (2.159)Observations 50 50 50Pseudo R-squared 0.272 0.196 0.254

Logit models. Standard errors in parentheses.∗∗∗p < 0.01, ∗∗p < 0.05, ∗p < 0.1.

1. The first model (Model 8) seeks to explain the presence or absence ofparental leave protections to private-sector employees before the passageof the FMLA. We draw our measure of these protections from CharlesL. Baum II (2003). The results are consistent with our earlier analysis.All three of our main explanatory variables reach statistical significance:as expected, Democratic dominance and the percentage of women in thelegislature are positively correlated with the passage of pre-FMLA familybenefits. The size of the evangelical population is negatively correlatedwith such benefits.

2. Model 9 seeks to understand the determinants of benefits in the year2000 that exceeded the minima required by the FMLA. Data are drawnfrom Wisensale (2001). Again, all of our main coefficients have theexpected sign, but only the Democratic control of the legislature andwomen in the legislature achieve statistical significance. Given the shortwindow that states had to introduce additional legislation beyond theFMLA, we believe these results lend additional credibility to our findingsthus far.

3. Model 10 seeks to predict the introduction of legislation providing forpaid parental leave before 2000 as part of the Clinton “Baby UI” ini-tiative. Wisensale (2006) documents which states responded quickly tothis proposal. The results suggest that Democrats, when in the majority,and state legislatures that had larger shares of women, were able to morequickly introduce legislation to take advantage of the Clinton provision.Both of these coefficients are positive and significantly correlated withthe introduction of UI-financed parental leave policy. The evangelical


variable shows a negative relationship to the family leave expansion ef-forts, but does not achieve statistical significance.

Taken together, these statistical analyses suggest that a history of DemocraticParty control and high proportions of women in state legislatures lead tothe expansion of parental leave policy, while large populations of evangelicalChristians exercise pressure to limit the size and scope of such policy.

Conclusion

This article provides a political explanation for variation in family leavepolicies across U.S. states. Our findings highlight a battle between competingvisions of the family and of the state, with Democratic legislators on one side,favoring more expansive family leave policies, and evangelical Christians onthe other, preferring to limit such measures. In addition, a growing proportionof women in state legislatures play a role in promoting and assuring passageof more generous family leave policies.

We have tested this theory using a variety of measures of family leave policy.Our analysis suggests that Democratic Party control of state legislatures, sus-tained over time, has a statistically significant, positive effect on the generosityof state parental leave policy. While we recognize that some studies have notfound strong partisan effects on policy at the state level, we believe that ourfindings may reflect political dynamics particular to the dependent variable wehave examined: family leave policy. Partisan effects may be more pronouncedin this particular policy realm because of its combined social and economicimplications, but get washed out when other policy outcomes are studied.Thus, our analysis points out the need for more fine-grained measurementsof social policy outcomes in testing partisanship hypotheses.

Our analysis also highlights the role of evangelical Christians in shapingfamily leave policy development. Remarkably, the evangelical Christian popu-lation in a state displays a consistently negative correlation with every measureof family leave policy we employ (although it does not reach conventionallevels of statistical significance in all of our alternative, dichotomous models).This finding helps explain the divergent policy results in California and NorthCarolina, for example. The former is the far-and-away leader in paid familyleave; its strikingly small evangelical population has not generated any mean-ingful opposition to the expansion of family benefits. North Carolina, on theother hand, has seen family policy proposals blocked repeatedly; in that state,evangelicals make up nearly 40 percent of the population (as compared tothe national average of 25 percent). The mechanism producing these resultsneeds to be better specified and tested, but the pattern observed here suggeststhat religious alignments can have a significant effect on even fine-grainedelements of policy design.


We also find that the percentage of state legislative seats filled by womenhas a significantly positive effect on the generosity of parental leave protec-tions. Women play an important role in introducing legislation and movingit through committees, and ultimately achieving passage of family leave mea-sures. However, what matters most is numbers: if women do not make up alarge enough proportion of the legislature, passage may prove elusive. NorthCarolina illustrates this point well. In the late 1980s and early 1990s, staterepresentative Annie Brown Kennedy repeatedly introduced a parental leavebill, very similar to the FMLA, in the General Assembly. However, she lackedsufficient support to achieve passage, and the bill failed on four separate oc-casions (Thomas, 1991). North Carolina falls in the bottom half of statesaccording to the number of women in their legislatures.

While the results of this study successfully explain variation in family leavepolicy across U.S. states, we believe that additional research is warranted.First, the number of observations in our quantitative analysis is limited bydata availability. We have relied on a single measure of family leave policies,the modified NPWF score, calculated for the single year of 2005, for ourcentral models. While this measure is particularly well suited to our analysis,as it codes various features of family-leave policy in a consistent manner acrossU.S. states, it suffers from providing only a single cross-section of states andmay ignore or obscure important dynamics across time. Extending this dataset to include such fine-tuned dependent variables for earlier and later periodswould provide additional leverage for testing our hypotheses. Nevertheless,changes in family leave policy tend to be extremely difficult politically andtherefore infrequent, and we believe that our results would be unlikely tochange substantially if data for additional years were available. Indeed, ouranalysis of other dichotomous measures of family leave policy—including thepresence of pre-FMLA policy, programs that go beyond FMLA, and the choiceof states to use unemployment funds to pay family leave benefits—suggeststhat similar dynamics determine family leave policies at various points in time.

Second, our results may be driven in part by the large number of states thatoffer no services beyond the minimum federal requirements. This is one of themain tradeoffs of studying a federal country: many covariates are controlledfor, but variation may be truncated since states may “default” to the federallymandated policy provisions. We have sought to compensate for this by usingdifferent modeling strategies and several measures of policy generosity which,as noted above, result in similar findings, but additional case studies and finer-grained examinations of policy provisions would better test our hypotheses.

Finally, our findings suggest that the advancement of family leave policiesrequires a complex alignment of interests that may prove difficult to achievein practice. Indeed, passage of such legislation requires a lasting partisanorientation in the legislature, as well as the development of social trends thatdo not emerge suddenly, but evolve over generations. To a pessimistic reader,this means that inequality of benefits is likely to persist across U.S. states, andthat federally mandated policies will not change anytime in the near future.


But to the optimistic reader, it suggests that considerable opportunities existat the state level, especially as women achieve greater representation in thelegislature. Indeed, the diversity in policies that presently exist indicates thatadditional change in family leave policy at the state level is not only possible,but likely.

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Appendix

Summary Statistics

StandardVariable Observation Mean Deviation Minimum Maximum

Yrs Dem 91–05 50 6.52 6.09 0 15Yrs Dem 76–05 50 15.8 11.59 0 30Yrs Dem 76–00 50 13.58 10.04 0 25Yrs Dem 76–93 50 10.96 7.62 0 18Percentage

Evangelical50 25.33 10.35 6.7 47

WomLeg 91–05 50 21.31 6.90 6.68 37.75WomLeg 76–05 50 16.94 6.10 5.53 30.01WomLeg 76–00 50 15.82 6.12 4.78 30.26WomLeg 76–93 50 13.62 6.02 2.89 29.92Log GSP Percap 50 4.44 0.08 4.28 4.65Unemployment 50 5.19 1.07 2.8 8.18Union Members 50 13.53 5.87 4.08 27.16Urban Population 50 71.24 14.64 39.19 94.08High School 50 57.36 4.07 51.35 65.09College 50 14.18 2.48 8.21 19.77Advanced

Degree50 7.59 1.84 4.98 12.17

LegislativeProfessionalism

50 0.18 0.12 0.034 0.571

Pre-FMLA Law 50 0.24 0.43 0 1Beyond FMLA 50 0.34 0.479 0 1Intro Paid 50 0.26 0.443 0 1Adjusted NPFW

Score50 17.88 26.88 0 110

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