What’s in a (Party) name? Examining preferences,discipline,and social identity in a parliamentary free vote

Abstract

In trying to understand the effects of political parties on shaping the voting behaviour of legislators, research has attempted the difficult task of separating the effects of preferences from rules used by party leaders to enforce discipline. However, little research has explored the prospect that party labels also reflect a social identity that is independent of legislators’ preferences and the rules used by party leaders to enforce discipline. In this study we examine that possibility, employing a data set that permits us to control both for leadership-based effects and legislator preferences on a 2000 free vote dealing with stem cell research. Using the British Representation Studies 1997 – which interviewed Members of Parliament regarding their preferences on several key issues related to the bill – we find significant evidence that party-as-identification plays a role in shaping how legislators vote, even after preferences and discipline are accounted for.

Keywords

free votes party identification political parties social identity

What is a political party? How do parties operate? How do they encourage cohesive behaviour among identifiers? Generations of scholars have devoted considerable attention to these and similar questions. One anomaly in the voluminous literature that has resulted is the disparity between the theories generated at the mass and at the elite levels of analysis. At the mass level, one of the principal explanations for partisanship has centred on the psychology of social identity (see Campbell et al., 1960). At the elite level, in contrast, psychological explanations are essentially absent,¹ at least until quite recently. For instance, among legislators party cohesion is typically explained as a function of shared ideology/preferences (see Krehbiel, 1993) or the mechanisms of party discipline (see Cox and McCubbins, 1993) or a combination of the two (Aldrich and Rohde, 2000).

In this paper, we capitalize on an unusual opportunity to test for the presence of a social identity motivating the behaviour of an important group of political elites, Members of Parliament (MPs) in the House of Commons in the United Kingdom. Specifically, we examine their votes on the Human Fertilisation and Embryology (Research Purposes) Regulations. Since this was a free vote in the Commons, whips were not used and the chamber’s usually strict party discipline was relaxed. We can also control for MPs’ individual policy preferences, using their responses to related questions on the 1997 British Representation Study (BRS). Since both free votes and measures of MPs’ individual inclinations are rare, this vote presents an uncommon chance to test the cohesive clout of party identification among elite policy makers even in the absence of usual party discipline and after controlling for shared policy predilections. While one test can hardly be considered dispositive, our finding that party identification remains significant – statistically and substantively – net of other forces certainly suggests that there is a psychological dimension to partisanship even among elected elites.

Our analysis proceeds as follows. The next section reviews the literature regarding the treatment of party effects. Then we outline our argument regarding the social psychology of intergroup relations in order to understand why party (or any other group-based) identification may lead members to act cohesively net of their personal preferences. We then move to discuss the context surrounding the Human Fertilisation and Embryology (Research Purposes) Regulations. Following that, we describe the details of the research design and discuss the results of our analysis. Finally, in a concluding section, we discuss the implications of the findings for the literature regarding legislative voting behaviour.

The roots of ‘party’ and their effects on legislative voting behaviour

While the ‘textbook’ view of the American Congress developed in the decades after World War II often held that parties played only an insignificant role (see, most prominently, Mayhew, 1974), the attention paid to legislative parties has surged in recent decades and party dominance has long been taken for granted in most legislatures outside the United States (Beer, 1965). Exactly how legislative parties exercise their powers, however, remains hotly contested. One perspective holds that the cohesion seen in the behaviour of legislative co-partisans is fundamentally based on shared policy preferences, often proxied by measures of ideology. In the US Congress, this case has been made most strongly by Keith Krehbiel (1993) who in 1993 famously defied the discipline to provide evidence of a significant partisan effect on members’ voting behaviour independent of shared preferences. In parliamentary analysis there is a long-standing assumption of ‘party cohesion’, perhaps best articulated and explicated by Ozbudun (1970; see also Bowler et al., 1999) and studies of free voting behaviour (about which more below) have generally uncovered considerable party consensus even on supposedly non-partisan votes (see, for example, Hibbing and Marsh, 1987; Marsh and Read, 1988; Moyser, 1979; Overby, 1996; Overby et al., 1998).

Numerous scholars took up Krehbiel’s challenge, producing a rich literature that has questioned the assumption of preference homogeneity within parties (see for example Kam, 2001) and examined the many ways that legislative party leaders can coerce disciplined behaviour from members even in the absence of policy agreement. This ‘parties-versus-preferences’ literature is most developed in studies of the US House of Representatives. Studies in the ‘cartel’ tradition of Cox and McCubbins (1993, 2005) argue that majority party leaders enjoy relatively invariant powers to control the legislative agenda, ensuring that the majority party will rarely if ever be rolled on the floor. Others, in the so-called conditional party government tradition of Aldrich and Rohde (2000; see also Aldrich et al., 2002; Rohde, 1991; Sinclair, 1992) contend that party leadership powers are highly contingent, expanding when party composition is internally cohesive and externally distinctive, contracting otherwise. For instance, in their study of members’ behaviour on discharge petitions, Miller and Overby (2014) found a significantly greater likelihood that House members would ‘waffle’ (that is, initially sponsor or co-sponsor a bill but then not sign a related petition to bring the bill to the floor for a vote) during periods of stronger conditional party government, when prospects of leadership punishment were more credible. Understandably, less empirical work has been undertaken on the ‘parties-versus-preferences’ debate in parliamentary settings due to the difficult of disentangling their effects in systems characterized by high levels of partisan voting cohesion. What limited research has been done, however, suggests – as Kam (2001: 115) puts it – that ‘party affiliation provides vastly more information about … behaviour than do … preferences’.

What are missing from most discussions of legislative parties are explanations based on social psychology (or, as some would have it, sociology). From the perspective of other corners of the discipline, this is odd, since at the mass level explanations of partisanship have long rested largely on psychological conceptions. Indeed, there is a distinguished literature in this tradition going back at least to The American Voter (Campbell et al., 1960; see also, notably, Lewis-Beck et al., 2008; Miller and Shanks, 1996), which holds that party affiliation in the electorate is based on psychological identification that goes well beyond simply agreeing with the party’s principal policy positions. In legislative studies, however, such psychological explanations are notably rare.² Indeed, they seem to be limited exclusively to accounting for behaviour in ‘unusual’ cases, where normal theories have less purchase, such as the US Senate (see Overby and Bell, 2004, who conclude that senators’ internalized norms account for the relative paucity of filibusters better than rational choice explanations) and the British House of Lords – see Norton (2003) and Russell (2014) – who argue that social, and social-psychological, factors help maintain party cohesion even in the absence of party discipline).

A social-psychological view of legislative parties

As Russell (2014) has noted, the scholarly inattention to social-psychological perspectives on legislative partisanship is both broad and surprising. Neither political scientists (Monroe et al., 2009) nor psychologists (Huddy, 2003) have employed psychological models to explain legislative party cohesion, even though individual–group interaction has been described as the ‘master problem’ of social psychology (Allport, 1962) and legislative parties would seem to be ideal candidates for the application of such theories. As Brown (2000: 64) has summarized, intra-group cohesion is to be expected in exactly the sort of environments in which legislative parties operate: physical proximity, frequent interactions, similarities among members, commitment to group goals and the presence of inter-group rivalries.

From a social-psychological point of view, there are numerous reasons – both positive and negative – why legislators’ partisan identification should have an independent effect on their behaviour and produce party cohesion. On the positive side, it is natural to look to the behaviour of others in terms of making all sorts of decisions (Festinger, 1954). The actions of others are often ripe with useful information, and mimicking their behaviour allows for quick and efficient decision making. Conversely, shunning by other group members provides a negative incentive for cohesive behaviour, since the pain of ostracism has been shown to be similar to physical pain (MacDonald and Leary, 2005). The combination of such compelling positive and negative factors helps explain why generations of social-psychological research have found that individuals’ attachments to groups are quick to form (even when assignment is random) and deep (see for example Diehl, 1990; Sherif et al., 1961; Tajfel, 1970).

For our purposes, the social identity perspective – perhaps the most dominant viewpoint within social psychology – is most relevant. Social identity theory holds that personality has two components, individual identity and group identity, the latter influenced if not determined by group membership. As Russell (2014: 715) puts it, ‘we define ourselves, to a large extent, through the groups to which we belong’. As a result of the need to identify with a group, individuals will routinely minimize differences between themselves and their perceptions of the rest of their group (Tajfel, 1981; Tajfel and Turner, 1979), in some instances going so far as to alter their preferences to match their identities out of a need to maintain cognitive balance (Heider, 1958). In addition, to maintain the distinctiveness of their identities, individuals will also exaggerate the differences between themselves and their perceptions of out-groups (Vignoles and Moncaster, 2007). In the legislative arena, we speculate that due to biased perceptions of which groups’ positions are closest to their own, individuals may behave in ways that benefit the party and contradict their personal preferences (at least objectively).

This view of party identification as a social group identity is well-documented among voters in the mass electorate. Previous research has demonstrated that party identification is a social identity that produces worldviews pitting one’s in-group against out-groups (Abrams, 1994; Greene, 2004; Kelly, 1988). Moreover, party identification produces biased opinions that minimize the differences between themselves and their in-groups and exaggerate perceptions of out-group differences (Bartels, 2002; Greene, 1999). At the elite level, however, this perspective has been applied only rarely. Indeed, in the legislative literature, the only study of which we are aware that uses a social-psychology approach is Russell’s (2014) recent study of party cohesion in the British House of Lords. Attempting to account for the high levels of partisan solidity in a chamber devoid of the usual positive and negative incentives leaders use to encourage discipline, Russell employed a 2007 mail survey to query members on their personal partisan allegiances, their views of their colleagues’ loyalties, and the nature of the concerns they had about ‘voting against the whip’. The responses Russell reports (for example that 44% of respondents feared breaking ranks would ‘upset other members of the group’ and that 73% feared it would be ‘damaging to my party group’) and the results of her multivariate regression analysis (that is, the statistical significance of a ‘social identity’ factor motivating voting with party in the absence of policy agreement) both speak to the utility of a social-psychological lens for observing parliamentary partisanship.

Parliamentary free votes

In this paper, we seek to extend the nascent literature on a social identity basis for legislative partisanship by expanding it beyond the House of Lords to the more important – and more typical – House of Commons. While conceptually easy, in practice this requires overcoming several significant barriers. Specifically, in order to focus on the effects of identity we need to be able to control for discipline and preferences. Both can be tough nuts to crack. Party cohesion is storied in the Commons, with Rice index measures exceeding 99% (Sieberer, 2006). This lack of intra-party variation in voting behaviour renders it challenging to differentiate exactly what ‘party’ means and how ‘party’ works to structure voting choices. This difficulty is exacerbated by the relative lack of independent measures of policy preferences for MPs. Unlike in the US Congress, where scholars capitalize on the relatively low rates of party cohesion to profitably use interest group ratings and scaled measures such as NOMINATE scores as proxies for individual legislator ideology (if not preference), such measures are essentially unknown in Westminster-style parliaments.

Fortunately, unwhipped free votes provide rare, but useful, opportunities to peer inside the normally opaque black box of parliamentary party cohesion. For these votes, the whips are withdrawn and MPs are released, to vote according to their consciences and/or the preferences of their constituencies. Often utilized for divisive morality issues (for example abortion, gay rights, death penalty), these free votes are sometimes referred to as conscience votes. However, as noted above, studies in the British Commons (Cowley and Stuart, 1997, 2010; Hibbing and Marsh, 1987), the Canadian Parliament (Overby et al., 1998), the Ontario Legislative Assembly (Overby, 1996), the Australian Parliament (McKeown and Lundie, 2002), and the New Zealand Parliament (Plumb, 2013) have routinely found strong party cohesion, even in the absence of whips.³ In contrast, while party remains strong, research by Overby et al., (2011) on the Canadian Civil Marriage Act has demonstrated the utility of using free votes to tease out the effects of factors such as constituency preferences, which are generally overwhelmed by or hidden beneath the effects of party discipline. Similarly, research by Longley (1999), Mughan and Scully (1997), and Pattie et al., (1994) has found certain legislator characteristics to emerge as significant predictors of behaviour on free votes.

Differentiating fully among the effects of social identity, discipline and preferences requires not just a free vote, however, but also a vote on which we have a reasonably good take on MPs’ personal policy preferences. Understandably, such measures are unusual. Since backbenchers are assumed to be loyal followers, their individual policy predilections are rarely plumbed in any depth. This leaves researchers dependent on the confluence of two highly infrequent circumstances: a free vote taken on a matter that coincides with an appraisal of relevant parliamentary attitudes. Fortunately, the 1997 British Representation Study queried parliamentary candidates about their attitudes regarding several issues relevant to stem cell research in a way that permits us to control for such preferences in examining the 2000 free vote on the Human Fertilisation and Embryology (Research Purposes) Regulations (HFER).⁴ This in turn permits us to assess if there is an underlying social identity component to the partisan voting we observe. Our logic is simple. If, after controlling for party discipline via the use of a free vote and for policy preferences via survey data, our analysis still uncovers evidence of a party effect, it is reasonable to conclude that parliamentary partisanship has a significant social identity component. Before undertaking that analysis, however, we turn to a discussion of the HFER.

The human fertilisation and embryology (research purposes) regulations 2001

When it was adopted on 24 January 2001, the Human Fertilisation and Embryology (Research Purposes) Regulations 2001 amended the Human Fertilisation and Embryology Act 1990, which regulates the use of all embryos for purposes both research-based and reproductive (for example in-vitro fertilisation), in order to allow for research using embryonic stem cells for the purpose of increasing knowledge regarding embryonic development, as well as knowledge about and treatment of serious diseases (Janson-Smith, 2002). The draft of the bill, tabled on 12 December 2000, was debated and a free vote on the third reading of the bill in the House of Commons was held on 19 December (see the debates in Hansard on that date starting at column 211). The House of Lords followed suit on 22 January 2001.

Opposition to the HFER in both the public debates and the debates in Parliament came primarily from those opposed to abortion and those citing moral concerns rooted in their Christian faiths. During the debates in the House of Commons regarding passage of the HFER, a number of MPs voiced their opposition to the bill, raising concerns about the compatibility of the HFER with religious values (with multiple references to ‘playing God’ in the debates in Hansard). Other MPs voiced concerns that passage of the HFER would allow for human reproductive cloning (which is different from the therapeutic cloning practiced in stem cell research). Similar to previous research showing that opposition to embryonic stem cell research among the public is rooted in large part in religious values and concerns about abortion (see for example Nisbet, 2005), the public debate over this issue was no different. The groups opposed to the HFER included pro-life groups and religious organizations opposed to abortion. These groups oppose stem cell research in part because non-embryonic stem cell research often involves the use of aborted foetuses, while embryonic stem cell research involves the use of fertilised embryos (which, for those who assume life begins at conception, is very much akin to the use of aborted foetuses). Like many MPs, these groups also made the connection between abortion and embryonic stem cell research (and many also made the case that embryonic stem cell research involves human reproductive cloning).

Despite the contentious nature of the debate in the public and in Parliament, the HFER ultimately passed easily, with 366 MPs voting to allow embryonic stem cell research and 174 voting against (119 MPs either abstained or were absent for the vote). As seen in Table 1, majorities of the Labour and Liberal Democrat caucuses voted in favour of the HFER, a large (65%) majority in the case of Labour. Conversely, but consistent with previous findings (Cowley and Stuart, 2010; Plumb, 2013), the Conservatives were more split. While a plurality (48%) of the more religious and socially traditional Conservative caucus voted against the HFER (including a majority – 55% – of those who cast a vote), a sizable percentage (38%) of Conservative MPs voted in favour. The figures summarized in Table 1 suggest that while there was some considerable disunity within party ranks (especially among Tories), in general it was the case that partisans continued to vote like partisans. More formally, the Cramérs V value of 0.27 establishes a strong statistical relationship between partisanship and voting behaviour among MPs on the HFER. Whether this relationship between party and vote choice is determined more by shared preferences among MPs or by party identification qua identification remains to be seen.

Table 1.

Vote choices on the human fertilisation and embryology (research purposes) regulations free vote by party.

	Vote choice
Party	Noes	Ayes	Abstaining and absent MPs	Total
Conservative	78 (47.56)	63 (38.41)	23 (14.02)	164 (100.00)
Labour	76 (18.14)	273 (65.16)	70 (16.71)	419 (100.00)
Liberal Democrat	12 (25.53)	27 (57.45)	8 (17.02)	47 (100.00)
Other	8 (27.59)	3 (10.34)	18 (62.07)	29 (100.00)
Total	174 (26.40)	366 (55.54)	119 (18.06)	659 (100.00)

Entries are the number of MPs in each vote choice category on the Human Fertilisation and Embryology (Research Purposes) Regulations free vote (with percentages in parentheses). Percentages may not sum to 100 due to rounding error. Likelihood ratio chi-square: 88.10 (DF = 6), significant at < 0.001. Cramér’s V: 0.27.

Data and methods

To determine whether party identification has an effect on MPs’ voting behaviour independent of shared policy preferences, we match voting decisions on the HFER with survey data regarding MPs’ preferences. These data are taken from the British Representation Study 1997 (see Norris, et al., 1997), a postal survey sent to all major-party parliamentary candidates prior to the 1997 general election (the election preceding the HFER). The BRS received a response rate of 61.4% (including 277 incumbent MPs, or roughly 42%).⁵ While some have used statements made during parliamentary debates to measure MPs’ preferences (see for example Balint and Moir, 2013), we prefer the BRS measures, for two reasons. For one, not all members express opinions during debates, in part because not all members have equal opportunity to speak. Second, speeches made during debates often reflect strategic motivations (for example, MPs demonstrating commitment to the party or voicing constituency preferences), while survey data from the BRS collected in confidence are more likely to reflect MPs’ sincere, personal preferences.

To establish that the sample of MPs who returned the BRS was broadly representative of the Commons, we compared the percentages for each vote choice between the actual observed outcome on the HFER and the sample of MPs produced by the BRS. These comparisons are presented in Table 2. Generally, they show that the sample produced by the BRS is quite representative of the final vote on the HFER, particularly for Labour MPs. While the differences between the BRS sample and the actual outcome produce values for Conservative and Liberal Democrat MPs that are slightly distorted – Conservative MPs voting ‘no’ and Liberal Democrats voting ‘aye’ are over-represented, while Conservatives abstaining and Liberal Democrats voting ‘no’ are under-represented – the greatest difference between observed and actual percentages exceeds six percentage points. Additionally, while the percentages for MPs of ‘Other’ parties voting ‘aye’ and abstaining differ much more than for the principal parties, this is due simply to the very small number (n = 29) of such MPs and the even smaller sample of them produced in the BRS (n = 5). In general, we are confident that our analyses are not unduly affected by sampling bias in the BRS.

Table 2.

Comparison of the BRS sample and the observed percentages on the human fertilisation and embryology (research purposes) regulations free vote.

	Vote
	Noes		Ayes		Abstaining and absent MPs
Party	Actual	BRS	Actual	BRS	Actual	BRS
Conservative	47.56	52.83	38.41	39.62	14.02	7.55
Labour	18.14	21.02	65.16	64.20	16.71	14.77
Liberal Democrat	25.53	21.88	57.45	62.50	17.02	15.63
Other	27.59	20.00	10.34	20.00	62.07	60.00

Entries are the percentages in each party according to their vote choices on the Human Fertilisation and Embryology (Research Purposes) Regulations free vote in 2000. ‘Actual’ refers to the percentages observed for all party members. ‘BRS’ refers to the smaller sample of MPs obtained in the British Representation Study. Percentages may not sum to 100 due to rounding error.

We use the BRS to generate several variables to control for MPs’ policy preferences. The most important of these is attitudes toward abortion. As noted above, for those who believe that life begins at conception, the use and/or destruction of human embryonic stem cells implicates concerns with the larger issue of abortion: those who oppose abortion usually oppose the use of stem cells in research, while those in favour of abortion also tend to support the use of stem cells.⁶ We use an item from the BRS measuring MPs’ perceptions regarding the availability of abortion, creating a variable coded from zero (for those who believe that availability of abortion on the National Health Service has ‘[n]ot gone nearly far enough’) to four (for those who believe that availability of abortion on the National Health Service has ‘[g]one much too far’). We also control for MPs’ degree of religious commitment since it is reasonable to expect that MPs professing to hold particular, strong religious beliefs may be less inclined to support the HFER (Bryant and Gudgin, 2008). Based on BRS responses, we created a variable measuring MPs’ frequency attending religious services. This variable is coded four for MPs who attend once a week, three for those attending once a month, two for attending once a year, one for those who attend less often than once a year, and zero for those who never attend.

Since some previous research has found that constituency concerns can assert themselves on free votes (Overby et al., 2011), we include five variables to control for such effects. The first two of these deal with perceived and actual electoral security. Previous research in the United States (Abramowitz, 1995; Koch, 2001) suggested that abortion issues are more motivating to conservative/Republican identifiers than to liberal/Democratic voters. As such, it is reasonable to suspect that less secure MPs might be less willing to vote in favour of the HFER, since their support could trigger an asymmetrically negative response among constituents. Our measure of perceived electoral insecurity is coded one for MPs who expressed concern that the 1997 election would be close in their constituency and zero otherwise.⁷ A second variable measures MPs’ levels of electoral security more objectively using their majorities in the 1997 election, calculated as the difference between MPs’ vote percentage and that of the runner-up.

A third variable measures the ideological congruence between MPs and the constituency wing of their parties. We combine six ideological variables – left–right placements, economic priorities, taxes and spending, nationalisation of industry, the European Union, and women’s rights – that are measured on a scale from zero to nine (with the leftist response coded zero) into a single index for both MPs’ self-placements and their perceptions of the constituency party.⁸ We then subtract perceptions of the constituency party from MPs’ self-placements. Higher values reflect MPs who are farther to the right of their constituency party, while smaller values reflect MPs who are farther to the left.

We also include two variables measuring MPs’ degree of commitment to representing their constituencies. This controls for the possibility that any residual party effects might be driven by (principally Conservative) MPs who were highly concerned with representing their constituents’ opposition to stem cell research. One of these variables measures the importance MPs attach to living in the constituencies they represent (coded one for MPs who think it is important to live in the constituencies they represent and zero otherwise).⁹ The other measures MPs’ professed interest in representing their constituencies. This variable is coded one for those MPs who listed constituency representation as their most important reason for running and zero for all others.¹⁰

We also include a variable in order to account for any residual effects of party discipline. While the 2000 HFER free vote appears to have been a genuinely discipline-free event, with all parties releasing their members to vote as they saw fit,¹¹ it is possible that MPs who fear their national party leaders might wish to demonstrate discipline even on a free vote. To account for this, we include a variable coded one for MPs who feel that national party leaders are too influential in the nominating process and zero otherwise.¹²

Guided by previous research on free votes, we also control for several personal characteristics of MPs that might be expected to influence voting behaviour. Because better- educated MPs may have a greater understanding of stem cell research than less well-educated MPs, and therefore may be more favourable to such research, we include a variable coded one for MPs with a university education and zero otherwise. To account for any differences in support for the HFER related to gender, we include a variable coded one for female MPs and zero for males.¹³ Finally, because older MPs may be more socially conservative (and therefore less supportive of stem cell research) than younger MPs, we include a variable controlling for MPs’ ages.¹⁴

Finally, we also include a measure of party identification, which all previous studies have shown to retain overwhelming importance even on free votes. Of course, in this case, since we have controls in place to account for policy preferences, constituency effects, personal characteristics and even the residue of party discipline on a free vote, we have greater confidence that our measure of party identification really does tap identification per se, in the psychological, social identity sense of that word. In our analysis, because Conservatives appear the most interesting and most cross-pressured MPs, we code them as one and all others as zero. Since our dependent variable (vote choice on the HFER) is categorical, we use logistic regression (where the dependent variable is coded one for those voting ‘aye’ and zero for those voting ‘no’, ignoring abstainers) and multinomial logistic regression (where abstainers are coded two) regression models. Summary descriptive statistics for our independent variables may be found in Table 3.

Table 3.

Summary descriptive statistics for independent variables related to the 2000 free vote on the human fertilisation and embryology (research purposes) regulations.

Variables	Mean	Standard deviation	Range
Attitude toward Abortion	1.91	0.77	0 – 4
Church attendance	2.10	1.18	0 – 4
Perceived closeness of race	0.49	0.50	0 – 1
MP’s electoral margin	0.17	0.11	0.00003 – 0.51
Ideological congruence	–0.65	4.20	–13 – 10
Importance of representing constituency	0.06	0.24	0 – 1
Importance of living in constituency	0.07	0.26	0 – 1
Influence of national party leaders	0.04	0.20	0 – 1
University education	0.75	0.43	0 – 1
Gender	0.19	0.39	0 – 1
Age	49.80	8.62	26 – 72

Results

The results of our logistic regression model are presented in the left column in Table 4. As the coefficient for the Conservative Party variable shows, party identification had a significant effect on vote choices, with Conservative MPs demonstrating significantly greater opposition to the HFER, ceteris paribus. This finding is noteworthy not just because it is (further) evidence that party matters on free votes, but because for the first time we show a partisan effect even after controlling for a raft of personal, constituency, and – especially – preference characteristics. If our logic is correct, the partisan effect that we show net of personal and constituency factors and policy preferences must be based upon a social identity shared among co-partisans.

Table 4.

The determinants of vote choices on the human fertilisation and embryology (research purposes) regulations.

	Logistic Regression	Multinomial Logistic Regression
Variables	Aye vs. No	Aye vs. No	Abstain vs. No
Conservative Party identification	–1.60** (0.79)	–1.54** (0.75)	–2.50** (0.92)
Opposition to abortion	–1.40*** (0.44)	–1.34*** (0.40)	–0.68 (0.44)
Church attendance	–0.57** (0.21)	–0.58** (0.21)	–0.25 (0.25)
Perceived close race	–0.90 (0.62)	–0.75 (0.60)	–1.73** (0.71)
MP’s vote majority	–1.04 (2.93)	–1.31 (2.94)	–1.88 (3.44)
Ideological congruence	–0.13** (0.05)	–0.11** (0.05)	–0.17*** (0.06)
Importance of representing constituency	0.89 (0.64)	0.75 (0.61)	–0.47 (0.84)
Importance of living in the constituency	–0.64 (0.53)	–0.72 (0.52)	–0.07 (0.60)
Influence of national party leaders	–0.51 (0.94)	0.20 (0.88)	1.78** (0.92)
University education	0.71 (0.52)	0.77 (0.49)	1.25** (0.63)
Gender	0.47 (0.64)	0.37 (0.61)	0.62 (0.73)
Age	–0.03 (0.03)	–0.03 (0.03)	–0.03 (0.03)
Constant	6.68*** (1.94)	6.84*** (1.88)	3.26 (2.22)
Likelihood Ratio Chi-Squared	69.00***	90.28***
Per cent Correctly Classified	81.77%	N/A
McFadden’s R²	0.30	0.22
n	181	217
	Predicted Probabilities
Party Identification	Aye	Aye	Abstain
Conservative	0.35	0.38	0.03
Baseline	0.73	0.66	0.12

**p < 0.05, ***p < 0.01, one-tailed tests. Entries are log-odds ratios with standard errors in parentheses.

It is worth noting that our findings are robust and not contingent upon estimator choices. We find substantively similar results when bootstrapping our standard errors and estimating the model with other techniques, such as probit and OLS. In addition, our results are also not contingent upon how we treat abstentions. MPs may have a variety of reasons for not voting, especially on a free vote. Failure to vote may be due to unavoidable absence from the chamber or it might reflect cross pressures faced by the MPs. In short, abstentions may be more or less randomly distributed among members and betray little useful information; or they might relay quite important insights into voting behaviour. While the abstentions among major party MPs reported in Table 2 suggest little of partisan pattern, we nevertheless estimated a multinomial logistic regression model,¹⁵ which included abstentions as a separate category. As can be seen in the results summarized in the rightmost columns of Table 4, this makes little difference, because our substantive results remain virtually unchanged: Conservative MPs continue to display significantly more opposition to the HFER (and are significantly less likely to abstain as well).

The predicted probabilities presented at the bottom of Table 4 help demonstrate the substantive effects of party identification more clearly. Specifically, we present the predicted probabilities of Conservative MPs versus other MPs voting in favour of the HFER, holding all other variables at their median values. Using the results from the logistic regression model, the probability of voting ‘aye’ on the HFER for the average MP was predicted to be 73% (66% using the results from the multinomial logistic regression model), while the predicted probability of voting ‘aye’ among Conservative MPs was only 35% (38% using the results from the multinomial logistic regression model). In other words, Conservative Party identification led to a sizable (38%) decrease in the probability of voting in favour of the HFER among the average Conservative MP.

While Conservative party identification significantly reduces the likelihood of voting ‘aye’, several control variables reaching statistical significance help to explain why so many Conservatives voted in favour of the HFER. Opposition to abortion, church attendance, and ideological congruence significantly reduce the likelihood of voting in favour of the HFER. Table 5 illustrates the effects of these three variables, presenting predicted probabilities at different levels of each of these three control variables. The results show that while the average Conservative MP is likely to vote against the HFER, a sizable number – roughly 35% – still do. The percentage deviating from the rest of the party increases further for those Conservative MPs less opposed to abortion, who attend church less frequently and perceive themselves to be to the left of their constituents, with a majority or near-majority voting in favour of the HFER for those Conservative MPs one unit or standard deviation below the median/mean of these three variables. By comparison, among MPs comprising the baseline category (primarily Labour and Liberal Democrat MPs), higher values of each of these three variables produced less disunity (that is, MPs were more likely to vote in favour of the HFER). While the predicted probabilities of voting ‘aye’ are lower at values below the median/mean of each of the three variables for all MPs, the predicted probabilities for the baseline are considerably higher than the probabilities among Conservative MPs. Not only does this further underscore the effect of Conservative party identification, it also shows that Labour and Liberal Democrat party identification increased support for the HFER.¹⁶

Table 5.

The effects of several control variables on predicted probabilities.

	Party identification
Opposition to abortion	Conservatives	Baseline
One unit below median	0.69	0.92
Median	0.35	0.73
One unit above median	0.12	0.40
	Party identification
Church attendance	Conservatives	Baseline
One unit below median	0.49	0.82
Median	0.35	0.73
One unit above median	0.23	0.60
	Party identification
Ideological congruence	Conservatives	Baseline
One SD below mean	0.50	0.83
Mean	0.37	0.74
One SD above mean	0.26	0.63

Entries are predicted probabilities of voting ‘aye’ on the HFER when holding each control variable at different values. All other control variables are set at their median values.

Conclusion

In this paper we have capitalized on an unusual confluence of circumstances surrounding the 2000 HFER to establish the presence of a significant, underlying social identity dimension to parliamentary partisanship. Delving into the party cohesion observed even in the absence of whips, we used elite survey data from the 1997 British Representation Study to control also for other suspected foundations of party solidity, such as shared policy preferences. Our finding that party still matters net of these other bases strongly suggests a fundamental social-psychological root to partisan identification. Indeed, it is quite possible that, if anything, we underestimate the psychological basis to partisanship in the House of Commons. Since psychologists have long recognized that expectations of conformity are lower on matters involving moral issues (Hornsey et al., 2003), our use of a conscience vote on stem cell research is inherently conservative in nature.

Our findings have implications both scholarly and practical. On the academic level, our study contributes to a small but growing literature attempting to integrate what we know of partisanship at the mass level with our understanding of elite-level partisanship in policy-making institutions such as legislatures. Furthermore, our study is one of the first to test for a social identity basis to partisanship outside of rather unusual environments such as the US Senate or the British House of Lords. In the future, when we ask questions like ‘where’s the party?’ we would be well advised to take into account the psychologically binding nature of party identifications.

At the practical level, our findings suggest that we may need to re-evaluate how we conceptualize legislative bargaining. With party identification independently producing party cohesion, legislative bargaining becomes more complex than trying to ‘buy’ the votes of individuals; instead, bargaining must focus (at least to some degree) on bargains among groups of legislators. This can have severely negative consequences if legislators share similar preferences on the issues, yet maintain distinctive party identities; in such situations, no amount of issue agreement will overcome gridlock unless bargains can be reached between the parties as groups. While there are potential downsides to group-based behaviour in legislatures, there are also potential upsides, because fostering bargains between groups of legislators may be sufficient to help overcome differences in issue positions: if bargains can be reached across parties as groups, such bargains may help identifiers to support their party by voting in favour of bargains even if their preferences are not totally satisfied by the terms of the agreement. While such speculation is tentative, the findings presented here open up such possibilities for future research to explore.

Footnotes

Acknowledgement

We thank Pippa Norris for generously providing us with the data from the British Representation Survey. All responsibility for any errors, however, belongs to the authors.

Declaration of Conflicting Interests

The author(s) declared no potential conflicts of interest with respect to the research, authorship, and/or publication of this article.

Funding

The author(s) received no financial support for the research, authorship, and/or publication of this article.

Notes

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