Gay Rights in Congress © The Author 2016. Published by Oxford University Press on behalf of the American Association for Public Opinion Research. All rights reserved. For permissions, please e-mail: journals.permissions@oup.com GAY RIGHTS IN CONGRESS PUBLIC OPINION AND (MIS)REPRESENTATION KATHERINE KRIMMEL* JEFFREY R. LAX JUSTIN H. PHILLIPS Abstract Public majorities have supported several gay rights policies for some time, yet Congress’s response has been limited. We document and analyze this tension through dyadic analysis of the opinion–vote relationship on 23 roll calls between 1993 and 2010, revealing a nuanced picture of responsiveness and incongruence. While constitu- ent preferences influence white male Democrats, black lawmakers and white female Democratic lawmakers generally support gay rights and Republicans consistently oppose them, regardless of constituent prefer- ences. Moreover, changes in constituent opinion typically fail to engen- der vote changes. In sum, despite a degree of responsiveness to opinion, we find there is a persistent bias against constituent will on LGB rights. Scholars have long argued that public opinion tends to shape government pol- icy (e.g., Page and Shapiro 1983; Stimson, MacKuen, and Erikson 1995). The potential for policy change in the realm of lesbian, gay, and bisexual (LGB) rights is therefore thought to depend on cultivating public support. Yet, support has not translated straightforwardly into policy gains on LGB rights issues. National and subnational public opinion data show that Americans favor a variety of legal protections for LGB individuals (Egan and Sherrill 2005; Brewer 2008), yet these protections have not always been adopted. It is there- fore difficult to gauge what kind of congressional action—if any—to expect in Katherine Krimmel is an assistant professor of political science at Boston University, Boston, MA, USA. Jeffrey R. Lax and Justin H. Phillips are associate professors of political science at Columbia University, New York, NY, USA. For helpful comments, the authors thank Robert Erikson, John Kastellec, Kelly Rader, participants at the Law and Institutions Conference at Emory University, and seminar participants at Columbia University, the University of Michigan, and Yale University. They also thank Andrea Gardner for outstanding research assistance, and the Human Rights Campaign for sharing their scorecards. *Address correspondence to Katherine Krimmel, Boston University, Department of Political Science, 232 Bay State Road, Boston, MA 02215, USA; e-mail: kkrimmel@bu.edu. Public Opinion Quarterly doi:10.1093/poq/nfw026 Public Opinion Quarterly Advance Access published June 29, 2016 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from mailto:kkrimmel@bu.edu?subject= http://poq.oxfordjournals.org/ response to the continuing liberalization of opinion on LGB rights. We need to know not only whether, but to what extent, public opinion incentivizes support from different types of legislators in roll-call votes on LGB issues. We study the match and mismatch between opinion and policy on LGB rights, engaging in a deep descriptive case study tied to larger theoretical ques- tions about the role of public opinion in the political process. We develop new tools and approaches to examine how public preferences relate to congres- sional action on LGB rights. Specifically, we analyze roll-call votes on all five major gay rights issues addressed by Congress from the early 1990s to the present: same-sex marriage, adoption, hate crimes, employment nondiscrimi- nation, and military service. We develop an extension of multilevel regres- sion and poststratification (MRP) for this project, to estimate opinion on these issues over time by state and congressional district, and consider the extent to which constituent preferences connect to the votes of members of Congress (MCs). While we find some evidence of responsiveness to opinion, we also document significant breaks, gaps, and lags. Our findings point to a more nuanced understanding of the power and limits of majority will. Moreover, the MRP extension we present herein expands the range of surveys that can be used to estimate district opinion, facilitating future studies of responsiveness. Our case study is also important in its own right: battles over LGB rights have played a central role in America’s ongo- ing “culture wars,” and the policy outcomes of these battles affect the lives of millions of Americans. Our analysis also contributes to long-standing debates over the relative roles of top-down and bottom-up forces in producing civil rights gains, and the appropriate standard of review for LGB rights in federal courts. Theoretical and Empirical Foundations Scholars of Congress have long argued that MCs’ desire for reelection moti- vates them to consider their constituents’ preferences in formulating policy positions (Mayhew 1974; Arnold 1990). Many studies find a positive corre- lation between such preferences and roll-call votes.1 These correlations are stronger on salient matters (Burstein 1981; Page and Shapiro 1983) and moral- ity policy (Mooney and Lee 1995; Lax and Phillips 2012). This suggests we should expect strong opinion effects on LGB rights issues, akin to Miller and Stokes’s (1963) “instructed delegate” model of representation in which MCs know the preferences of the median constituent and act accordingly. Scholars have also shown that the median constituent does not always get her way (Bishin 2000, 2009; Hacker and Pierson 2005). As Fenno (1978) 1. See Miller and Stokes (1963) and, more recently, for example, Clinton (2006); Kousser, Lewis, and Masket (2007); Bafumi and Herron (2010); and Kastellec, Lax, and Phillips (2010). Krimmel, Lax, and PhillipsPage 2 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ famously articulated, MCs have several constituencies to consider. Within an MC’s geographic constituency lie her reelection constituency (electoral supporters), primary constituency (most active and strong supporters), and personal constituency (closest allies and advisors). When their preferences conflict, she may need to prioritize. For Republican lawmakers, religious conservatives create pressure to vote against LGB rights. Correspondingly, LGB individuals and their allies have become a notable constituency of Democratic lawmakers (Fetner 2008). MCs are also likely to face pressure from party leaders. Starting in 1992, the national parties have staked out increasingly divergent positions on LGB rights. These positions play a significant role in party branding (Bishin and Smith 2013).2 Thus, we anticipate partisan differences in responsiveness. Interest group and party pressures should lead Republicans to be less sensitive than Democrats to liberalizing public preferences on gay rights.3 The MC’s own views and background can also matter. Elites who are members of historically oppressed groups might set aside public opinion unfavorable to LGB rights, drawing analogies between their own civil rights struggles and the fight for gay rights. The National Association for the Advancement of Colored People (NAACP) has supported gay rights since the debate over open military service reached the national stage in 1993, even over the objections of many of its members and sometimes black public opinion more broadly (cf., Edsall 1993; Conant 2010; Wallsten 2012), linking the battle for LGB rights to the struggle for African American civil rights (cf., Robinson 2012). Similarly, the National Organization for Women (NOW) has supported the cause since 1971, when it expanded its mission to include lesbian rights. Another reason the median constituent may not get her way is stickiness. Besides potential status quo biases, MCs might not wish to “flip-flop,” even if public opinion has changed. Lawmakers could also be unaware of opinion change, though this is less likely for salient policies. Most existing work on public opinion and LGB rights compares state-level opinion to state policy adoption (e.g., Haider-Markel 2001; Haider-Markel and Kaufman 2006; Lax and Phillips 2009b).4 We cannot assume Congress will match state-level patterns. On the one hand, congressional votes are more visible to the public, so federal lawmakers may be more sensitive to 2. For example, the 1992 Republican platform opposed including sexual preference in federal civil rights statutes, legal recognition of same-sex relationships, adoption of children by gay and lesbian couples, and open inclusion of gays and lesbians in the military; the Democratic platform called for civil rights protections for gays and lesbians and an end to discrimination in the Defense Department. 3. This parallels Brady and Schwartz’s (1995) finding that concern about primary election con- stituencies constrained Republicans’ responses to liberalizing opinion on abortion in the 1970s and 1980s. 4. Lax and Phillips (2009b) showed that clear supermajority support for some policies failed to spur changes in state law. Gay Rights in Congress Page 3 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ public preferences. On the other hand, party pressures are perhaps greater in Congress. Moreover, state-level studies connect opinion to system-level out- comes, rather than individual lawmakers’ votes. The few existing studies of congressional action on gay rights tend to employ coarse measures of constituent preferences and legislative behavior. To capture the former, scholars create indices of pro-gay opinion by averaging constituent preferences across several issues; to capture the latter, they create indices aver- aging roll-call votes and/or cosponsorships (Lewis and Edelson 2000; Haider- Markel 2001; Oldmixon and Calfano 2007). This inhibits precise analysis, since surveys consistently document much greater support for some gay rights policies than others. Further, indices of opinion and policy lack a common metric, severely constraining the inferences one can draw. Researchers can show the degree and direction of the correlation between constituent ideology and roll-call voting, but, unlike our dyadic analysis, cannot tell whether MCs follow their median constituent, whether policy is over- or under-responsive to opinion, whether opinion or ideology is the key, how responsiveness varies across policies, or whether opinion change results in policy change. Notably, Bishin and Smith (2013) depart from the trend of using indices to measure opinion, using MRP to calculate district-level opinion on DOMA. They find that MCs consider opinion generally, and pay particular attention to important subconstituencies, consistent with findings by Bishin (2000, 2009). However, public opinion on this set of votes was exceptionally low (on aver- age, 29 percent of constituents supported same-sex marriage at this time), and a broader analysis is needed. Data and Methods ROLL-CALL VOTES We evaluate the opinion-vote relationship on 23 roll-call votes across the five issue areas considered by Congress between 1993 and 2010. We use survey questions on the issue being voted upon around the time of (almost always before) the vote (see table 5 in the supplementary data online for details). Adoption: There were two House votes on amendments to the Washington, DC appropriations bill to prohibit unrelated couples in Washington, DC from adopting a child (one passed in 1998; one failed in 1999). Same-sex marriage: There were three proposals. (1) The Defense of Marriage Act (DOMA) defined marriage as a union between one man and one woman, so that the federal government could not recognize same-sex marriages and no state would be required to recognize those from out of state. It passed both chambers by wide margins. (2) The Federal Marriage Amendment (FMA) aimed to amend the Constitution to define marriage as a union between one man and one woman, but failed to receive the requisite supermajority in the House in 2004, and failed cloture votes in the Senate in 2006. (3) An Krimmel, Lax, and PhillipsPage 4 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/lookup/suppl/doi:10.1093/poq/nfw026/-/DC1 http://poq.oxfordjournals.org/ amendment to the Health Care and Education Reconciliation Act, suspending the issuance of marriage licenses to same-sex couples in DC, was rejected by the Senate in 2010. Gays in the military: There were four failed votes in 1993 and three successful votes in 2010. Of the failures, two tried to codify a full ban on military service by gays and lesbians, and two aimed to allow the president to decide the issue.5 In 2010, the House voted twice and the Senate once to repeal DADT, the pol- icy prohibiting the military from asking recruits about their sexual orientation but allowing the military to discharge gay service members. Jobs: ENDA sought to prohibit employment discrimination on the basis of sex- ual orientation. It was defeated by one vote in the Senate in 1996, and passed the House in 2007. A 1998 effort to defund President Clinton’s executive orders prohibiting discrimination in the federal civilian workforce failed in the House. Hate crimes: There were votes in both chambers in 2000 and 2009 on a pro- posal to expand existing hate crimes protections to include sexual orientation. In 2000, the measure passed but died in conference committee. In 2009, the bill was signed into law. OPINION ESTIMATION: MR. P GOES TO WASHINGTON To estimate opinion for each roll-call vote in our analysis, we use multilevel regression and poststratification (MRP). This technique, first presented by Gelman and Little (1997), uses national surveys and advances in Bayesian sta- tistics and multilevel modeling to generate opinion estimates by demographic- geographic subgroups. MRP can produce accurate estimates of public opinion by state and congressional district using as few data as in a single national survey and fairly simple demographic-geographic models (Park, Gelman, and Bafumi 2006; Lax and Phillips 2009a, 2013; Warshaw and Rodden 2012). MRP proceeds in two stages. The first estimates a multilevel model of individual survey response as a function of a respondent’s demographic characteristics as well as her state and (where appropriate and available) her congressional district. Our models use gender, race, age, and education.6 We include state- and district-level variables that should be correlated with sup- port for gay rights.7 Finally, we control for slight differences across polls and thus variation in question wording. 5. We interpret a “yea” as a vote to allow gays to serve openly in the military, since this was President Clinton’s position. 6. Gender uses two categories: male and female. Race uses three categories: black, Latino, and white or other. Age uses four: 18–29, 30–44, 45–64, and 65+. Education uses five: less than high school, high school degree, some college, college degree, and postgraduate degree. 7. State effects are modeled as a function of region, percent African American, percent Mormon or Christian Evangelical, and percent voting Democratic in the prior presidential election. District effects are modeled similarly (except for district religion data, which are not available), and are grouped by state effect. Gay Rights in Congress Page 5 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ We then use this model to “predict” opinion for each demographic-geo- graphic type of respondent (e.g., the probability that a college-educated black female in New York of age 30–44 supports same-sex marriage). Then, the opin- ion estimates for each type are weighted (poststratified) by their percentages in actual populations (in the state or district) to form the state or district estimate. Population frequencies come from Public Use Micro Data Samples supplied by the Census Bureau, with conversions to congressional district using the Missouri Census Data Center’s Geographic Correspondence Engine (geocorr2k).8 “CROSS-LEVEL MRP” One challenge in generating estimates by congressional district is that polling data for some issues do not include district identifiers, preventing the direct use of district-level predictors in the modeling stage even if we have them at the poststratification stage. Normally, if we used a variable such as presiden- tial vote measured at the district level, we would predict response for a given hypothetical respondent type using that hypothetical respondent type’s district* the estimated coefficient on district-level presidential vote. We always have the former term (by public record) but sometimes lack the latter term. In these cases, one simple way forward would be to use state-level values and correla- tions for things like presidential vote, with the recognition that the poststrati- fied estimates will then vary only because the demographic composition of each district varies. Fortunately, there is another way forward, which does not ignore district- level correlations, and outperforms the simpler solution described above. This modification to basic MRP runs the modeling stage at the only possible level of aggregation, the state level, but imputes the needed coefficient on (say) presidential vote to be the coefficient on district-level presidential vote as if we had originally done the analysis at the district level. Here, we use state-level (as opposed to district-level) values for presidential vote share in the first-stage response model. Then, in the prediction stage, we multiply the resulting coef- ficients on these state-level variables not with the state-level values but with the congressional district values for presidential vote share. We do the same with percent black. This assumes that the “effect” of aggregate presidential vote on individual response is the same whether measured at the state level or district level, or at least that an estimate from the former is a good estimate of the latter.9 To validate this, we take policies that do have congressional district identi- fiers in the survey data and estimate district opinion using (1) a standard MRP that makes use of the congressional district identifiers; (2) an MRP that uses 8. We use distinct poststratification files for the periods before and after the 2000 national redis- tricting, after the 2003 Texas redistricting, and after the court-required changes to the 2003 Texas redistricting. 9. Ardoin and Garand (2003) call this top-down estimation. Krimmel, Lax, and PhillipsPage 6 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ state-level data for presidential vote and percent black in both the response model and in prediction; and (3) an MRP that uses state-level data for presi- dential vote and percent black in the response model, but district-level values for these variables in the prediction and poststratification phase (our mod- ification). Figure  1 plots estimates of district-level opinion for three issues using survey data that include congressional district (cd) identifiers (on the y-axis) against similar estimates that do not make use of these identifiers (on the x-axis). The top panel uses state-level presidential vote and share black in both the response model and the prediction phase; the bottom panel uses state-level values of these variables in the response model, but district-level values in the prediction phase. The 45-degree line is shown. Our modified MRP strongly improves the accuracy of estimates, compared to using only state-level information in both the response model and prediction, producing estimates of district-level opinion that are very similar to the estimates we get when district identifiers are available. Figure  1. Validating Cross-Level MRP. This figure plots estimates of district-level opinion for three issues using survey data that includes con- gressional district (cd) identifiers (on the y-axis) against similar estimates that do not make use of these identifiers (on the x-axis). The top panel uses state-level presidential vote and share black in both the response model and prediction phase; the bottom panel uses state-level values of these variables in the response model, but district-level values in the prediction phase. The 45-degree line is shown. Gay Rights in Congress Page 7 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ DATA SUMMARY Table 1 displays summary statistics for our 23 roll-call votes and related opin- ion. Our estimates are coded in the pro-gay direction: higher values indicate higher support for gay rights. Results RESPONSIVENESS AND CONGRUENCE If MCs act as instructed delegates on gay rights issues, their roll-call votes should be both highly responsive to and congruent with constituent prefer- ences. That is, a strong positive correlation should exist between constituent opinion and the probability of a pro-gay vote. A congruent vote is one aligned with majority constituent opinion. RESPONSIVENESS Each graph in figure 2 takes one roll-call vote and plots the probability of an individual legislator casting a vote in favor of LGB rights against our estimates of opinion. Responsiveness to public opinion is strong if the logit curve is steep and positively sloped. For each of our 23 roll-call votes, the probability of an MC casting a pro-gay vote is indeed positively correlated to the level of public support for gay rights in the MC’s home district or state. Bivariate regressions show that the slopes of all of the logit curves are statistically sig- nificant at the 95 percent level, and that the slopes vary across policies. CONGRUENCE The opinion–vote relationship appears weaker when it comes to actual con- gruence, and is often biased in one direction or the other. Consider the maps of majority opinion and roll-call votes on DADT and ENDA in figure 3. There are far more conservative votes than there are conservative constituencies. We can see this in figure 2 as well. The dotted line extending from the x-axis indicates the 50 percent opinion level, and the line from the y-axis indicates a 50 percent pro-gay vote probability. The y-value at which the logit curve inter- sects the vertical dotted line is the predicted probability of a pro-gay roll-call vote when public support is 50 percent. The x-value at which the horizontal dotted line intersects the curve is the level of public support needed for the predicted probability of a pro-gay vote to reach 50 percent. In a system of perfect majoritarianism, the regression curves would be very steep at 50 per- cent opinion and pass through the crosshairs in the middle of each graph. This would yield perfect congruence. For some votes (cf., “FMA2006senate”), the curve comes close to this “majoritarian ideal,” but we do not always observe majoritarianism. Krimmel, Lax, and PhillipsPage 8 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ T ab le  1 . O p in io n a n d C on gr u en ce b y R ol l- C al l V ot e V ot e M ea n pr o- ga y op . ( % ) M in . pr o- ga y op . ( % ) M ax . pr o- ga y op . ( % ) P ro -g ay op in io n m aj or it ie s (% ) P ro -g ay vo te (% ) C on gr ue nc e (% ) N et l ib er al v ot e bi as H ou se N et l ib er al v ot e bi as S en at e A D O P T IO N 19 98 ho us e 44 21 63 35 46 69 43 A D O P T IO N 20 00 ho us e 45 21 66 35 50 69 59 D A D T 19 93 ba n se n at e 4 6 2 6 6 6 4 0 63 61 18 D A D T 19 93 b ox er se n at e 4 6 2 6 6 6 4 0 36 77 -8 D A D T 19 93 hu nt er ho us e 48 21 76 45 66 63 79 D A D T 19 93 m ee ha nh ou se 48 21 76 45 40 71 -3 0 D A D T 20 10 ho us e 59 39 79 87 54 68 -1 27 D A D T 20 10 m ur ph yh ou se 59 39 79 87 59 65 -1 46 D A D T 2 01 0 se n at e 58 45 70 88 68 74 -2 1 D C M A R R IA G E 2 01 0 se n at e 4 6 23 6 4 4 4 61 74 12 D O M A 19 96 ho us e 29 12 47 0 17 83 68 D O M A 19 9 6 se n at e 2 8 13 4 4 0 14 86 13 F M A 20 04 ho us e 56 38 77 80 45 62 -1 51 F M A 2 0 0 4 se n at e 55 42 73 78 51 59 -2 8 F M A 20 06 ho us e 54 31 80 65 44 73 -9 5 F M A 2 0 0 6 se n at e 53 4 0 67 58 49 8 0 -9 H A T E 20 00 ho us e 69 47 90 98 54 56 -1 87 H A T E 2 0 0 0 se n at e 67 51 82 10 0 58 58 -4 1 H A T E 20 09 ho us e 71 49 92 10 0 59 59 -1 72 H A T E 2 0 0 9s en at e 69 51 83 10 0 69 69 -2 8 JO B S1 9 95 se n at e 56 29 76 72 4 8 6 4 -2 4 C o n ti n u ed Gay Rights in Congress Page 9 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ V ot e M ea n pr o- ga y op . ( % ) M in . pr o- ga y op . ( % ) M ax . pr o- ga y op . ( % ) P ro -g ay op in io n m aj or it ie s (% ) P ro -g ay vo te (% ) C on gr ue nc e (% ) N et l ib er al v ot e bi as H ou se N et l ib er al v ot e bi as S en at e JO B S 19 98 ho us e 62 32 87 85 59 71 -1 13 JO B S 20 07 ho us e 73 47 92 99 56 56 -1 80 M ea n 54 33 74 64 51 68 -7 3 -1 2 N o te .— T he f ir st t hr ee c ol um ns s um m ar iz e op in io n by d is tr ic t or s ta te . T he f ou rt h is t he p er ce nt ag e of c on st it ue nc ie s w it h pr o- ga y op in io n m aj or it ie s. T he f if th an d si xt h co lu m ns s ho w p er ce nt ag es o f pr o- ga y ro ll -c al l vo te s an d co ng ru en t vo te s, r es pe ct iv el y. T he f in al c ol um ns a re t he n et n um be r of p ro -g ay v ot es , b y ch am be r, lo st d ue t o in co ng ru en ce . T he re i s a la rg e ra ng e in o pi ni on a cr os s st at es a nd d is tr ic ts . S en at e vo te s ar e in i ta li cs . T ab le  1 . C o n ti n u ed Krimmel, Lax, and PhillipsPage 10 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ Figure 2. Basic Relationships. Each graph plots the probability of a pro- gay vote from a logistic regression curve (the dark line) given state or dis- trict opinion (lighter lines are lowess curves). Each x- and y-axis runs from 0 to 100 percent for opinion and the probability of a pro-gay vote, respec- tively. Opinion in states/districts whose MC cast a pro-gay (anti-gay) vote are plotted in a “rug” on the top (bottom) axis. Dotted lines show the 50 percent marks in opinion and vote probability. Panels are ordered by the position of the curve relative to the 50 percent crosshairs (top to bottom, left to right). Gay Rights in Congress Page 11 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ Of course, some degree of incongruence is expected. An MC may not know or care about the difference between 48 and 52 percent support. Mismatches between opinion and voting near the majority threshold are not necessarily of academic interest either. When large supermajorities are needed to bring about a 50 percent chance of a pro-gay vote (i.e., if the logit curve is shifted far to the right of the crosshairs), however, there are significant biases in policymaking Figure  3. Maps of Opinion and Voting on “DADT2010house” and “JOBS2007house.” Krimmel, Lax, and PhillipsPage 12 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ that cannot be explained by uncertainty or dismissed as trivial.10 For exam- ple, in the case of “JOBS2007house,” constituent opinion needs to be 71 per- cent before the MC has a 50 percent probability of casting a liberal vote. For “HATE2009house,” constituent opinion needs to be 68 percent. Consequently, congruence for both is relatively low—only 56 percent for “JOBS2007house” and 59 percent for “HATE2009house.” In contrast, congruence for the 2006 Senate vote on the Federal Marriage Amendment (with a responsiveness curve that passes through the crosshairs) is a whopping 80 percent. Overall, we find that 68 percent of the 6,435 terminal roll-call votes in our analysis are congruent.11 By roll-call vote, congruence ranges from 56 to 86 percent. By issue area, congruence is highest on same-sex marriage (74 per- cent) and lowest on hate crimes (61 percent).12 SYSTEMATIC MISREPRESENTATION While incongruence of different types could theoretically cancel out, it does not. Only 552 of 2,089 (26 percent) incongruent votes are in the liberal direction. Votes against constituent preferences tend to be conservative votes here. The final col- umns of table 1 show the net liberal vote bias—the number of liberal incongruent votes minus the number of conservative ones. In the House, the greatest benefit the pro-gay side ever received from incongruence amounted to 79 votes, while they lost more than 150 votes four times. These mismatches between opinion and voting are often consequential. Under constituent opinion majorities, four roll-call votes would have flipped in the pro-gay direction (“FMA2004house,” “FMA2006house,” “FMA2006senate,” and “JOBS1995senate”), and three would have flipped the other way (“DADT1993senate,” “DADT1993hunterhouse,” and “DCMARRIAGE2010senate”). DIFFERENCES BY PARTY, RACE, AND GENDER Now consider “raw” voting records by MC, shown in figure  4. Each graph plots mean pro-gay opinion against the career percentage of pro-gay votes cast by each MC. The top-left panel captures the positive overall relationship 10. On non–civil rights issues, the same concerns would arise about liberal bias (i.e., instances in which the logit curve is shifted far to the left of the crosshairs). As Madison articulates in “Federalist #10,” however, there are legitimate reasons for MCs to ignore majority opinion that would oppress minority rights. 11. If we consider roll-call votes only where the size of the opinion majority is greater than 60 or 70 percent, then congruence rises to 78 and 86 percent, respectively. 12. To put this in perspective, Lax and Phillips (2009b) find a similar 62 percent level of congru- ence between opinion and policies (not votes) at the state level, which is significantly higher than the 48 percent congruence level that Lax and Phillips (2012) find over a much larger set of policy types. Matsusaka (2010) finds a 57 percent congruence level for a subset of those policies. Finally, Monroe (1998) finds a 55 percent match between national policies and national opinion majorities over a wide set of issues. Gay Rights in Congress Page 13 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ between opinion and roll-call voting. However, this aggregate panel masks significant variation by party and race, shown by the other three panels. Democratic non-black (i.e., white, Hispanic, or other race) MCs drive the aggregate relationship between voting and opinion; neither Democratic black MCs nor Republican MCs show much of a relationship between opinion and voting. Black MCs are concentrated at the top of the graph, and, comparing the flat lowess curve in the top-right graph to the steep curve in the lower left, there is a much weaker relationship between opinion and voting for black Democrats than for non-black Democrats. Democratic female MCs vote Figure 4. Pro-Gay Voting Record Given Opinion, By Party and Race. The unit is the member of Congress, plotted by mean pro-gay votes and mean pro-gay opinion. The size of the points shows the number of votes represented (from 1 to 14). Republicans are shown with squares, white Democrats with circles, and black Democrats with triangles. Lowess curves are displayed. Krimmel, Lax, and PhillipsPage 14 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ similarly to black MCs (see figure  8 in the supplementary data online).13 In sum, black MCs and Democratic female MCs tend to support LGB rights, even when their constituents do not. Republican MCs are far less responsive to the liberalization of opinion on LGB rights than Democrats. Two-thirds of Republicans in our sample have never cast a pro-gay roll-call vote, regardless of opinion.14 These results suggest that public opinion “matters” for roll-call voting only in a broad sense; the success of proposals to extend LGB rights will depend in large part on the composition of Congress. VOTE-SWITCHING We go further by focusing on MCs who cast multiple votes on the same issues over time, to see if opinion change matters (we might not expect votes to change if opinion has not). Since support for gay rights has been steadily increasing over time, we are most interested in studying the extent to which MCs shifted from opposition to support for gay rights. So, we analyze the 687 (of 1,453) pairs in which the first vote was against gay rights. Figure 5 plots pro-gay constituent opinion at the time of the first vote against pro-gay opinion at the time of the second vote. Each point represents a legisla- tor who voted twice on a particular issue. The key area is the top-left quadrant. Here, we have MCs whose constituents did not support gay rights at the time of the first vote, but did at the time of the second. In a world of perfect majori- tarianism, all of the dots in this quadrant would be black (i.e., every legislator should switch his vote). In reality, only 21 percent in this quadrant switched their votes (14 of 23 Democrats switched, while only 2 of 55 Republicans did). White male Republicans who started out with an anti-gay vote in an anti-gay district, whose district shifted to being pro-gay, had only a 4 percent chance of switching to the pro-gay position in the second vote. Similarly situated white male Democrats had a 65 percent chance of switching. Vote switching is related to opinion, in that switchers saw an 8 percent increase in pro-gay opinion between votes on average, while non-switchers saw a 1 percent increase. However, switching is uncommon overall, occur- ring in only 6.3 percent of vote pairs (91 switches), and is particularly rare among Republicans. Almost 40 percent of the Democrats whose constituents switched from anti-gay to pro-gay failed to follow as well. Uncertainty cannot explain all of this resistance—while some of the points in the graph’s top-left quadrant are clustered around the 50 percent mark, many are not (the same is true for the top-right quadrant). Thus, some degree of turnover in both parties may be necessary for LGB rights measures to succeed in Congress. 13. Female Republican MCs are between Democratic and Republican male MCs. See figure 8 in the supplementary data online. 14. Figures 9 and 10 in the supplementary data online show a comparison to MC Nominate scores for context. Gay Rights in Congress Page 15 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/lookup/suppl/doi:10.1093/poq/nfw026/-/DC1 http://poq.oxfordjournals.org/lookup/suppl/doi:10.1093/poq/nfw026/-/DC1 http://poq.oxfordjournals.org/lookup/suppl/doi:10.1093/poq/nfw026/-/DC1 http://poq.oxfordjournals.org/ MODELS OF ROLL-CALL VOTES The dependent variable is whether the roll-call vote cast was pro-gay (lib- eral).15 We include indicator variables for Republican, Female, Latino, Black, and Senate. We also include both dimensions of the Poole and Rosenthal meas- ures of MC ideology, DW Nominate 1 and DW Nominate 2. Table 2 displays results from eight model variants, to check robustness across specifications (with further notes in the supplementary data online, pp. iv–v) and to facilitate 15. We performed a similar analysis of congruence. Because the results were substantively identi- cal, we put this in the supplementary data online (see table 4) rather than in the paper. Our discus- sion of congruence in the “Time Trends” section of the paper refers to “bottom line” congruence; that is, we simply look at whether or not MCs voted with the majority in their state or district. The models in the supplementary data online facilitate more complex all-else-equal comparisons. For congruence, however, we believe the bottom line is sometimes more important. Figure 5. Vote Switching. We plot voting behavior for the 687 pairs of votes by the same legislator on the same issue where the initial vote was anti-gay. Each circle is a Republican, each square a white Democrat, and each triangle a black Democrat, filled in when the second vote was pro-gay, and hollow when the second vote was anti-gay. The x-axis shows opinion at time 1 and the y-axis at time 2, with the 45-degree line showing where opinion has not changed. We break the votes into quadrants to show whether opinion at each time was above or below the 50 percent mark. The fraction switching within each quadrant is shown. Krimmel, Lax, and PhillipsPage 16 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/lookup/suppl/doi:10.1093/poq/nfw026/-/DC1 http://poq.oxfordjournals.org/lookup/suppl/doi:10.1093/poq/nfw026/-/DC1 http://poq.oxfordjournals.org/lookup/suppl/doi:10.1093/poq/nfw026/-/DC1 http://poq.oxfordjournals.org/ Table 2. Responsiveness Models Responsiveness regressions (Did the legislator cast a pro-gay vote?) Model 1 Model 2 Model 3 Model 4 b (SE) b (SE) b (SE) b (SE) Opinion 6.2* (0.6) 5.7* (0.5) 4.5* (0.5) 6.0* (0.5) (Slope st. dev) (2.5) (1.9) (1.9) (1.8) Republican –3.7* (0.1) –3.6* (0.1) Female 1.2* (0.1) Fem. * Rep. Latino 0.5 (0.3) Black 0.9* (0.2) Senate 0.3 (0.5) DW Nom1 –5.0* (0.1) Intercept 0 (0.3) 2 (0.3) 0.2 (0.2) 1.7 (0.4) N 6,435 6,435 6,427 6,435 AIC 6,198 3,923 3,454 3,834 Model 5 Model 6 Model 7 Model 8 b (SE) b (SE) b (SE) b (SE) Opinion 4.5* (0.5) 4.3* (0.5) 6.0* (0.5) 3.0* (0.5) (Slope st. dev.) (1.8) (1.8) (1.9) (1.9) Republican .7* (0.2) –3.7* (0.1) –1.1* (0.3) Female 1.1* (0.2) 1.4* (0.3) 1.3* (0.3) 1.4* (0.4) Fem. * Rep. –0.6 (0.3) –0.3 (0.4) –0.7 (0.4) Latino 0.2 (0.3) 0.2 (0.3) 0.5 (0.3) 0.4 (0.3) Black –0.1 (0.2) –0.2 (0.2) 0.5* (0.2) –1.0* (0.3) Senate –0.1 (0.4) –0.1 (0.4) 0.3 (0.5) –0.7* (0.3) Black * Op. –2.0* (0.5) –2.2* (0.5) Latino * Op. 0.2 (0.7) –0.4 (0.7) Republican * Op. –0.8* (0.3) –0.6* (0.3) Female * Op. 0.4 (0.6) 0.8 (0.7) Fem. * Rep. * Op. –0.1 (0.9) –0.5 (1.0) DW Nom1 –5.0** (0.1) –5.7** (0.3) –5.5* (0.3) DW Nom2 –2.2* (0.1) Intercept 0.1 (0.3) –0.2 (0.2) 1.8 (0.4) 1.0 (0.2) N 6,427 6,427 6,435 6,427 AIC 3,405 3,398 3,823 3,062 Note.—Standard errors are shown next to the coefficients. Continuous variables are standard- ized (subtracting the mean and dividing by 2 standard deviations, putting them on the same scale as each other and roughly the same scale as the dichotomous variables). Two-tailed tests are used: *p < .05, **p < .01. Gay Rights in Congress Page 17 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ various “all else equal” comparisons (so raw coefficients must be interpreted with caution). We allow varying intercepts and slopes for opinion. The basic relationship between voting and opinion holds: MCs whose con- stituents demonstrate higher levels of pro-gay support are more likely to cast pro-gay votes, even controlling for party and ideology, which are themselves strongly related to constituent preferences. The effects of opinion remain when we control for Democratic presidential vote share in the state or district, and other similar predictors. At an average value of opinion (in model 1), an additional point of support increases the chance of policy adoption by approxi- mately five percentage points. Party also predicts voting (e.g., models 2, 4, and 7; in models 6 and 8, the Republican coefficient is the effect of party after controlling for Nominate score, a strange all-else-equal comparison). Model 4 shows that blacks and Latinos tend to vote pro-gay relative to whites, controlling for opinion (and not controlling for Nominate). Models 5 and 6 show almost no difference between blacks and whites once we control for Nominate, but this is true only on average, as explained later.16 Regressions confirm that black MCs are more likely to cast pro-gay votes than white MCs (see the positive, significant coefficient on Black in model 4). These models allow for the effect of opinion to vary by MC type. Additional pro-gay support matters less for black MCs than white MCs, as indicated by the negative and significant coefficient on the interaction with opinion in mod- els 7 and 8. For each additional point of policy-specific opinion (in model 7), the probability of a white male Democrat casting a pro-gay vote rises by 5. For white Republicans, the probability only rises by 4, and for black Democrats, it only rises by 3. Table 3 shows the level of pro-gay opinion needed for differ- ent types of MCs to have a 50 percent probability of casting a pro-gay roll-call vote, based on model 7. It is highest for white male Republicans (66 percent support) and lowest for black female Democrats (31 percent support). Overall, we find strong evidence for our hypothesis that support for gay rights should be especially high among MCs belonging to groups that have historically faced discrimination. Though our findings about Latinos depend on model specification, we consistently find that African American and female MCs are especially likely to cast pro-gay votes.17 Turning to differential 16. We generally find no significant differences between the House and Senate (see the supple- mentary data online, pp. iv, vii–viii). 17. Interacting party and gender, we find that the effect of gender varies by party. This is also clear in figure 8 in the supplementary data online. We do not interact race and party, because there is almost no variation in party identification among African American MCs, and little among Latinos. There does not appear to be a particularly strong interaction between party and race for Latinos. Latino Democrats voted pro-gay 90 percent of the time; Latino Republicans voted pro-gay 33 percent of the time. White Democrats voted pro-gay 82 percent of the time; white Republicans voted in the pro-gay direction 11 percent of the time. Latino Democrats are congru- ent 87 percent of the time; Latino Republicans are congruent 57 percent of the time. In com- parison, white Democrats are congruent 82 percent of the time, while white Republicans are congruent 47 percent of the time. Krimmel, Lax, and PhillipsPage 18 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/lookup/suppl/doi:10.1093/poq/nfw026/-/DC1 http://poq.oxfordjournals.org/lookup/suppl/doi:10.1093/poq/nfw026/-/DC1 http://poq.oxfordjournals.org/lookup/suppl/doi:10.1093/poq/nfw026/-/DC1 http://poq.oxfordjournals.org/ responsiveness, however, race seems more influential than gender. While women are not any more or less responsive to growing support for gay rights than men, changes in opinion have less influence on African American MCs than their white colleagues. This is primarily because black MCs strongly sup- port gay rights, even if their constituents do not. To further examine the LGB rights–civil rights connection, we coded floor speeches on four of our votes to see what kind of arguments MCs made in favor of LGB rights (see pp. v–vi of the supplementary data online). We found that, for each vote, civil rights arguments played a greater role in speeches by black Democrats, compared to white Democrats or Republicans. Older cohorts of black MCs (i.e., those socialized in the civil rights era) are more likely to cast pro-gay votes, relative to younger cohorts of black MCs, conditioned on opinion. We find no similar relationship for white or Latino MCs. We also find that the sec- ond dimension of Nominate, which tended to represent racial justice issues for much of the twentieth century, also influences voting on gay rights issues (see model 8). This is notable, as Poole and Rosenthal (2007) show that most conflict occurs on the first ideological dimension today. Overall, MCs’ preferences with regard to civil rights appear to influence their willingness to support LGB rights, and this connection is particularly strong for African American MCs. TIME TRENDS The snapshot provided thus far obscures important differences over time, illu- minated in figure  6. Reading these panels in order tells the following story: [1] Mean pro-gay opinion increased over time, from around 45 to around 60 percent. [2] The number of pro-gay opinion majorities increased more sharply, from around 35 to 85 percent. [3] However, the percentage of pro-gay roll-call votes cast increased far less dramatically, from 50 to 60 percent. [4] Surprisingly—for now—overall congruence stayed nearly constant (around 70 percent). [5] and [6] But, the nature of incongruence changed drastically. Incongruence, once leaning to the liberal side, now strongly cuts against pro- gay policy, measured either as a percentage of total incongruence (where the Table 3. “Coin Flip” Points “Coin flip” point Black female Democrats 31 (18, 39) Black male Democrats 39 (29, 45) White female Democrats 40 (35, 45) White male Democrats 46 (42, 50) White female Republicans 60 (55, 64) White male Republicans 66 (62, 70) Note.—We used model 7 to calculate the level of pro-gay opinion needed for a 50 percent probability of casting a pro-gay roll-call vote for six types of MCs, ordered from most pro-gay to least (for the average roll call). Confidence intervals are displayed in parentheses. Gay Rights in Congress Page 19 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/lookup/suppl/doi:10.1093/poq/nfw026/-/DC1 http://poq.oxfordjournals.org/ Figure 6. Opinion, Votes, Congruence, and Bias over Time. Averages for each stated quantity are shown over districts or states, as appropriate, over time. The dashed line and dotted lines are lowess curves for all policies and policy area subsets, respectively. Krimmel, Lax, and PhillipsPage 20 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ degree of incongruence is incorporated) or by the net vote bias (under +15 to –25 percent). The predicted conservative vote bias from incongruence now averages 109 votes in the House (i.e., 109 votes are “lost” because MCs are not following constituent opinion) and 25 in the Senate. Breaking this down by party, figure 7 reveals more nuance. Five of the panels parallel panels in the previous figure, and are labeled [P] to indicate as much. They show: [1P] Support for gay rights has grown in both Democratic and Republican constituencies over time. While the parties started roughly in the same place, in terms of opinion in their districts/states, Democrat-represented constituen- cies (DRCs) have grown more liberal at a slightly higher rate than Republican- represented constituencies (RRCs), leading to a small party gap today (roughly five percentage points). [2P] There is a larger and more constant gap in terms of opinion majorities. DRCs went from a rough split between pro-gay and anti- gay majorities to nearly 100 percent pro-gay majorities. Most RRCs have been majority pro-gay since the late 1990s as well (75 percent as of 2011). [3P] Within Congress, we observe a very different pattern. Democrats have steadily voted more pro-gay over time, starting from a relatively high base rate. But, Republicans have remained relatively constant around a much lower rate (less than 15 percent), despite changes in their constituencies. Consequently, while the gap between Democratic and Republican districts has grown only slightly, in Congress it has grown dramatically. (This is consistent with work by Hacker and Pierson [2005], who say increasingly conservative policymaking does not reflect public will.) The final panel focuses more directly on party polarization.18 [4P] Since RRC pro- gay majorities have become far more common, Republican congruence rates have plummeted from 75 to 35 percent. Democratic congruence has increased. [6P] This is partly because DRCs have moved in line with the initial pro-gay voting rates observed, and partly because Democrats have moved to match their increas- ingly pro-gay DRCs by voting along those lines. The initial liberal vote bias by Democrats disappeared by the early 2000s, and they have stayed in line on average since (that is, the remaining incongruence cancels out). Republicans, who started out with balanced incongruence, now show clear conservative bias relative to their RRCs. Congruence remained constant overall because the increase in Democratic congruence balanced the decrease in Republican congruence. The Limits of Responsiveness In his dissenting opinion in Lawrence v. Texas, Justice Antonin Scalia accused his fellow Supreme Court justices of being complicit in a subversion of the 18. The Rice likeness score is the absolute difference between the percentage of yeas cast by each party, subtracted from 100, measuring similarity of pro-gay voting rates between parties. The Rice cohesion score is voting agreement within each party (absolute difference between the yea and nay votes cast within a party) (Rice 1925, 1928). Cohesion has risen and likeness plummeted, a clear display of polarization. Gay Rights in Congress Page 21 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ Figure  7. Parties over Time. Panels other than the last are numbered in parallel with figure 6, but broken down by party. Democrats are shown with solid triangles and a dashed lowess curve; Republicans are shown with open circles and a dotted lowess curve. The bottom-right panel contains Rice like- ness scores (shown with gray open squares and gray solid line) and Rice cohesion scores. Krimmel, Lax, and PhillipsPage 22 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ democratic process, rushing ahead of public opinion: “So imbued is the Court with the law profession’s anti-anti-homosexual culture, that it is seemingly unaware that the attitudes of that culture are not obviously ‘mainstream.’” He invokes the lack of congressional action on LGB rights as clear evidence that the “homosexual agenda” is counter-majoritarian. We actually find the opposite pathology: there is a persistent conservative bias against constituent will on LGB rights. Our analysis might even under- estimate this bias, because it does not account for agenda control. Pro-gay legislation, even when supported by large opinion majorities, is almost never brought to a vote in the House when Republicans control the chamber. Our results suggest that LGB individuals cannot rely on the political pro- cess to further their rights. Support in roll-call votes requires a large (and likely unachievable) supermajority of public support for many MCs; and those MCs who are relatively impervious to gains in public support for LGB rights often constitute a pivotal voting bloc in Congress. LGB individuals may need to turn to courts to protect their rights, and courts would have the grounds to step in to do so—not being able to resolve the problem using normal political processes is one of the reasons that courts raise the standard of review they apply in cases of potential discrimination. LGB individuals seemingly qualify for “suspect class” status; that is, they are a group in need of particular pro- tection on the basis of political vulnerability (a concept introduced by Justice Stone in United States v. Carolene Products).19 Disaggregating MCs by party, gender, and race illuminates important nuances in the opinion–vote relationship that system-level studies, like those conducted at the state level, cannot capture. Increased polarization at the elite level has inhibited responsiveness to the liberalization of opinion on LGB rights. While black MCs and white female Democratic MC generally cast pro- gay votes regardless of their constituents’ preferences, they cannot compete with Republican MCs in terms of numbers, and the latter have maintained their anti-gay stance, even against strong pro-gay majorities. Altogether, these patterns have led to a large partisan gap in responsiveness, and a growing con- servative bias in policymaking. Our findings resist an easy categorization into top-down or bottom-up explana- tions for policy change. There seems to be a top-down process pulling black and female MCs toward gay rights, and bottom-up pressure from the public pushing 19. When the courts deal with discrimination claims, they can apply different standards of review, from the lowest hurdle, rational basis; to intermediate scrutiny; to strict scrutiny, the highest hur- dle. The rational basis test requires only some rational basis for the differential treatment, with a presumption of constitutionality. Strict scrutiny begins with a strong presumption of unconstitu- tionality, which the government can overcome only if it has a compelling interest in the law, which must also be narrowly tailored to that interest. There are two reasons to apply strict scrutiny: a violation of a fundamental right or if the population affected constitutes a “suspect class.” At present, “suspect classes” include race, religion, and national origin. We do not speak here to the “fundamental right” basis for strict scrutiny, but our results speak to the second basis. Gay Rights in Congress Page 23 of 26 at C olum bia U niversity L ibraries on A ugust 3, 2016 http://poq.oxfordjournals.org/ D ow nloaded from http://poq.oxfordjournals.org/ many white Democratic MCs. However, our analysis of vote change shows that Republican MCs are not the only ones who hold their ground against gay rights even when a majority of their constituents grow supportive—many Democratic MCs fail to switch their votes as well. On the whole, opinion change has limita- tions as a tool for achieving civil rights gains. Much depends on the partisan composition of Congress, and even member replacement among Democrats. Scholars have long argued that it is important for women and minorities to hold office because they should be most willing and able to represent people sharing their demographic characteristics (e.g., Pitkin 1967; Sapiro 1981; Phillips 1995; Mansbridge 1999), and have evaluated the extent to which such descriptive repre- sentation occurs (e.g., Swers 1998, 2002; Harris 2012). We cannot speak directly to this, as openly LGB members of Congress are still too few to study systemati- cally. We can look more broadly at the extent to which MCs who are members of historically underrepresented populations represent members of other histori- cally underrepresented populations. Our findings suggest that descriptive repre- sentation can operate in this broader sense; however, this can come at the expense of classic descriptive representation. Like the NAACP, African American MCs (many representing majority-minority states and districts) have sometimes sup- ported LGB rights over the objections of African American constituents.20 The potential for conflict between narrower and broader views of descriptive repre- sentation in Congress is intriguing and deserves further study. We would also like to see more work on dyadic representation and on stop- pages in the democratic process, to put our extended case study into context. Methodologically, we have extended the reach of the MRP opinion-estimation technique, facilitating this substantive research agenda. Supplementary Data Supplementary data are freely available online at http://poq.oxfordjournals. org/. References Ardoin, Phillip J., and James C. Garand. 2003. “Measuring Constituency Ideology in US House Districts: A Top-Down Simulation Approach.” Journal of Politics 65:1165–1189. Arnold, R. Douglas. 1990. The Logic of Congressional Action. 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