Divided we stand - Unified we govern? Cohabitation and regime voting in the 2002 French elections

British Journal of Political Science

Volumn 35, Issue 4, 2005, Pages 691-712

  Gschwend, Thomas ^a   Leuffen, Dirk ^a  

^a UNIVERSITY OF MANNHEIM   (Germany)

Author keywords

[No Author keywords available]

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EID: 25444458826     PISSN: 00071234     EISSN: None     Source Type: Journal    
DOI: 10.1017/S0007123405000359     Document Type: Article

References (74)

1. We borrowed the term 'regime preferences' from Sartori, who described the French Fifth Republic as a system that oscillates between the two poles of presidential and parliamentary regime characteristics. See Giovanni Sartori, Comparative Constitutional Engineering: An Inquiry into Structures, Incentives and Outcomes (New York: New York University Press, 1994), p. 123. The 'presidential' phase here stands for unified government, the 'parliamentary' phase, i.e. cohabitation, for divided government.
2. In considering cohabitation as divided government, we follow Robert Elgie, 'What is Divided Government?';
3. "Cohabitation": Divided Government French Style', in Robert Elgie, ed., Divided Government in Comparative Perspective (Oxford: Oxford University Press, 2001), pp. 1-20 and 106-26;
4. Matthew Soberg Shugart, 'The Electoral Cycle and Institutional Sources of Divided Presidential Government', American Political Science Review, 89 (1995), 327-43;
5. Alberto Alesina and Howard Rosenthal, Partisan Politics, Divided Government and the Economy (Cambridge: Cambridge University Press, 1995), p. 244.
6. For a different view, see Roy Pierce, 'The Executive Divided against Itself: Cohabitation in France, 1986-1988', Governance, 4 (1991), 270-94.
7. Thomas Gschwend, Strategic Voting in Mixed Electoral Systems (Reutlingen: SFG-Elsevier, 2004).
8. What information does a regime voter require? First, a voter must be aware that unified or divided government might result from a particular election outcome. A voter thus has to understand the regime implications of her vote choice. Secondly, a voter must be aware of the outcome of the election for the first branch before voting for the second.
9. Exceptions are: Lee Sigelman, Paul J. Wahlbeck and Emmett H. Buell Jr, 'Vote Choice and the Preference for Divided Government: Lessons of 1992', American Journal of Political Science, 41 (1997), 879-94;
10. Dean Lacy, 'A Theory of Nonseparable Preferences in Survey Responses', American Journal of Political Science, 45 (2001), 239-58, p. 253.
11. Most other literature has focused on the explanation for divided government, probably because of its prominence in the United States since the Second World War. For a review of this literature, see Charles E. Smith Jr, Robert D. Brown, John M. Bruce and L. Marvin Overby, 'Party Balancing and Voting for Congress in the 1996 National Election', American Journal of Political Science, 43 (1999), 737-64, p. 738.
12. See, for example, Mark N. Franklin and Wolfgang P. Hirczy, 'Separated Powers, Divided Government, and Turnout in U.S. Presidential Elections', American Journal of Political Science, 42 (1998), 316-26, pp. 317 and 318.
13. G. Bingham Powell Jr, Elections as Instruments of Democracy: Majoritarian and Proportional Visions (New Haven, Conn.: Yale University Press, 2000), p. 11.
14. See again Franklin and Hirczy, 'Separated Powers, Divided Government, and Turnout in U.S. Presidential Elections', p. 318.
15. For France this is particularly true in the matter of nominations, but also in foreign, security and European policy making. See Dirk Leuffen, 'Europeanisation and the Probabilities of Power during Divided Government in France' (paper presented at ECPR Joint Sessions of Workshops, Turin, 2002).
16. Michael Laver and Norman Schofield, Multiparty Government: The Politics of Coalition in Europe (Oxford: Oxford University Press, 1990), p. 83, give an example of such reasoning.
17. If this micro-logic about the motivation of first forming regime preferences and then voting accordingly is correct, this should imply that these voters are neither less educated, less interested in politics nor less politically efficacious than the average voter. These observational implications are, indeed, supported by our data: simple tgg-tests show that regime voters are not systematically different from the average voter on our measures of education, political interest and political efficacy.
18. We thus assume that such voters hold 'non-separable' preferences, as introduced into the literature on electoral behaviour by Dean Lacy and Philip Paolino, 'Downsian Voting and the Separation of Powers', Amercian Journal of Political Science, 42 (1998), 1180-99.
19. See also Smith et al., 'Party Balancing and Voting for Congress in the 1996 National Election'
20. Lacy, 'A Theory of Nonseparable Preferences in Survey Responses'. Whereas these authors similarly recognize that voting behaviour can be better understood in a two-dimensional framework, their notion of separable and non-separable preferences remains more general than our term of regime preference. When using regime preferences as a variable (that embraces the two categories of divided and unified government preference), we highlight the micro-mechanisms of vote choice. In contrast to the work cited above, our research design allows us to get around counterfactual reasoning, since in our case the voters already know who the president is.
21. Sartori, Comparative Constitutional Engineering.
22. The French presidential and parliamentary elections were held on 21 April, 5 May, 9 June and 16 June. The same structure, presidential elections closely preceding legislative elections, brought about a unified government in 1981 and a minority government in 1988. The relevance of a particular sequence of elections again points out the impact of institutions on vote choice, as for example highlighted by Robert Elgie, 'Institutions and Voters: Structuring Electoral Choice', in Michael Lewis-Beck, ed., The French Voter: Before and After the 2002 Elections (Basingstoke, Hants.: Macmillan, 2003), pp. 110-25.
23. The UMP was at first an election alliance combining the RPR (Rassemblement pour la République), parts of the UDF (Union pour la Démocratie Française) and DL (Démocratie Libérale). Not until November 2002 was the alliance transformed into a party under the new label of 'Union pour un mouvement populaire'.
24. David S. Bell and Byron Criddle, 'Presidentialism Restored: The French Elections of April-May and June 2002', Parliamentary Affairs, 55 (2002), 643-63, p. 663.
25. John A. Ferejohn and Randall L. Calvert, 'Presidential Coattails in Historical Perspective', American Journal of Political Science, 28 (1984), 127-46;
26. Robert S. Erikson, "The Puzzle of Midterm Loss', Journal of Politics, 50 (1988), 1011-29, p. 1023;
27. Warren E. Miller, 'Presidential Coattails: A Study in Political Myth and Methodology', Public Opinion Quarterly, 19 (1955), 353-68, p. 353. Usually, one would expect that on-year legislative elections would fall into a president's 'honeymoon' term since the two sets of elections are only separated by about a month. This period could be considered as too short to expect an occurrence of 'negative voting'.
28. See Samuel Kernell, 'Presidential Popularity and Negative Voting: An Alternative Explanation of the Midterm Congressional Decline of the President's Party', American Political Science Review, 71 (1977), 44-66.
29. This assessment is confirmed by Shugart, who claimed that 'elections held early after a presidential election are likely to produce a surge in support for the new president's party'; see Shugart, 'The Electoral Cycle and Institutional Sources of Divided Presidential Government', p. 337. However, in our case Chirac had already been in power since 1995 and he therefore cannot really be considered to be a new president.
30. Olivier Schrameck, Matignon Rive Gauche 1997-2001 (Paris: Seuil, 2001), p. 23.
31. Cited in Cole, 'A Strange Affair: The 2002 Presidential and Parliamentary Elections in France', p. 335.
32. Libération, 4 June 2002.
33. We used the French Inter-election Survey 2002 - The French CSES II Study, administered by CSA-TMO, Paris. As is usual in France, the survey (CATI) takes a large national sample of respondents randomly selected from national telephone numbers until quotas based on gender, age, occupation, size of city and region are fulfilled. Up to five contacts were made with every selected household. The response rate is 42 per cent. The main part of the survey is comprised of the common CSES II module of public opinion questions, which can be downloaded from the CSES homepage. We also analysed additional items about the electoral system and the nature of party competition that we built into this survey. These items will also be included in a replication dataset. The use of quota sampling is unfortunately a common practice among French polling organizations, see Michael Lewis-Beck, 'Editorial: An Introduction', French Politics, 1 (2003), 135-6.
34. Using the standard panoply of significance tests on data generated by any sampling method demands among other things that the achieved sample behaves as if it were generated by a simple random selection process, as emphasized by Michael Oakes, Statistical Inference: A Commentary for the Social and Behavioural Sciences (New York: John Wiley & Sons, 1986), p. 156.
35. So we have to evaluate this assumption empirically. In order to see whether our sample represents the population, following a strategy suggested in Thomas Gschwend, 'Analyzing Quota Sample Data and the Peer-Review Process', French Politics, 3 (2005), 88-91, we relied on additional data sources (i.e., other surveys and administrative statistics). We found that the marginal distributions of our sample in terms of attitudinal variables correspond rather well to the ones of the most prominent French election study conducted around the same time, the Panel électoral français, 2002. Moreover, our sample is representative (by design) in terms of occupation and, furthermore, the distribution of the remaining socio-demographic characteristics (religion and trade-union membership) recorded in our sample is consistent with administrative data sources. Thus we are confident that our sample does not systematically differ from the population of eligible French voters.
36. Considering a possible future cohabitation, which of these opinions do you feel closer to?
37. (a) Cohabitation would be a good thing for France;
38. (b) Cohabitation would be a bad thing for France.
39. In fact, only 5 per cent of all respondents of our survey were unable to provide an answer to this question.
40. John Zaller, The Nature and Origins of Mass Opinion (New York: Cambridge University Press, 1992);
41. John Zaller and Stanley Feldman, 'A Simple Theory of the Survey Response: Answering Questions Versus Revealing Preferences', American Journal of Political Science, 36 (1992), 579-616.
42. For the cohabitation item we purposely did not offer an 'I-do-not-really-have-an-opinion' category in order to get respondents to take a stand on this issue. This does not, however, bias our results in any way since we control for partisan encodings of this issue.
43. Comparing the distribution of vote intentions in our sample with the actual results of the first round that are published by the Assemblée Nationale (http://www.assemblee-nationale.fr/elections/resultats.asp), we get the following: 36 per cent report an intention to vote for a candidate of the right (actual: 39 per cent), 49 per cent report an intention to vote for a candidate of the left (actual: 37 per cent) and 15 per cent report an intention to vote for an extremist candidate (actual: 24 per cent). Thus, based on our sample we get an over-report of the intention to vote for a candidate of the left while we get an under-report of the intention to vote for an extremist candidate. This is consistent with prior research and, unfortunately, seems to be a tradition of election studies in France. Since we include non-voters in our analysis, the distribution of the dependent variable in our sample is as follows: 47 per cent intend to vote for a candidate of the left, 32 per cent report an intention to vote for a candidate of the right and 19 per cent report no vote or a vote for an extremist candidate. Since we are interested in estimating causal effects on the individual-level a slightly skewed distribution of the dependent variable in the aggregate is not problematic.
44. When the survey was designed the development of the UMP was not evident and we expected that voters would still use well-known party labels as their political referents.
45. Even if a respondent most prefers a moderate (left or right) and an extreme party at the same time, she is coded as an extremist. Based on respondents' placement of parties on the 10-point likes/dislikes scale, we divide up the electorate into 50 per cent supporters of the left, 32 per cent supporters of the right, 7 per cent extremists and 11 per cent respondents with no clear, i.e., unanchored, preferences.
46. We will not engage in the debate about whether left-right ideological attachment or party identification weigh more heavily with the French voter, see Christopher J. Fleury and Michael S. Lewis-Beck, 'Anchoring the French Voter: Ideology versus Party', Journal of Politics, 55 (1993), 1100-9;
47. Christopher J. Fleury and Michael S. Lewis-Beck, 'Déjà Vu All Over Again: A Comment on the Comment of Converse and Pierce', Journal of Politics, 55 (1993), 1118-26;
48. Philip E. Converse and Roy Pierce, 'Comment on Fleury and Lewis-Beck: "Anchoring the French Voter: Ideology versus Party" ', Journal of Politics, 55 (1993), 1110-17;
49. Philip E. Converse and Roy Pierce, Political Representation in France (Cambridge, Mass.: The Belknap Press of Harvard University Press, 1986). Instead, we use party preferences as well as ideological self-placement to determine an anchoring preference.
50. We find that 16 per cent of all supporters of the right have conflicting regime and partisan preferences, i.e., 5 per cent of all respondents despite most preferring a party on the political right favour cohabitation. Moreover, we find that 37 per cent of all supporters of the left have conflicting regime and partisan preferences, i.e., 19 per cent of all respondents despite most preferring a party on the political left oppose cohabitation.
51. This is the case for 61 per cent of all unanchored respondents, i.e., 7 per cent of all respondents are both politically unanchored and oppose cohabitation.
52. The literature on economic voting clearly suggests, though, that the attribution of credit or blame for the state of the economy gets diluted in times of cohabitation; see Michael Lewis-Beck, 'Who's the Chief? Economic Voting Under a Dual Executive', European Journal of Political Research, 31 (1997), 315-25;
53. Michael Lewis-Beck and Richard Nadeau, 'French Electoral Institutions and the Economic Vote', Electoral Studies, 19 (2000), 171-82.
54. See also Nonna Mayer and Vincent Tiberj, 'Do Issues Matter? Law and Order in the 2002 French Presidential Elections', in Michael Lewis-Beck, ed., The French Voter: Before and After the 2002 Elections (Basingstoke, Hants.: Macmillan, 2003), pp. 33-46.
55. Only twenty-three respondents were lost from subsequent analysis in that they could not provide an answer to the cohabitation question and at the same time had no missing data on all the other variables of our analysis. Thus the cohabitation issue was not a 'hard' one in the sense that respondents could not provide an answer to it.
56. Ideology and partisan preferences generally do not simply measure the same thing, particularly for partisans of the political left. Less than two out of three supporters of a left party place themselves on the left ideological spectrum. Furthermore, one out of five respondents most preferring a party of the political right do not actually place themselves on the ideological right.
57. Morris Fiorina, 'An Era of Divided Government', Political Science Quarterly, 107 (1992), 387-410.
58. Hausman tests show that independence of irrelevant alternatives (IIA) is not a problem in our data.
59. Our main interest is to model the cross-pressure mechanisms for voters with conflicting anchoring and regime preferences. With our design it is not possible to estimate directly the number of voters who would have voted differently were it not for their regime preference because for many respondents anchoring preference and regime preference are observationally equivalent. Comparing the model predictions with the ones of a model where all variables based on the cohabitation items are excluded is a way to estimate at least a lower bound for this number. Applying this technique, as a lower bound we predict that at least 7.4 per cent of all extremist or non-voters would have voted differently, mainly for a party of the left, while only minor differences (at most 1.2 per cent as a lower bound) can be observed for the total vote share of the right and the left. Here we nevertheless detect a clear tendency that the likelihood of voting for one's own camp is weakened if one holds a conflicting regime preference.
60. We use CLARIFY, a set of Stata ado-files, to carry out these simulations; see Michael Tomz, Jason Wittenberg and Gary King, CLARIFY: Software for Interpreting and Presenting Statistical Results, Version 2.1 (Cambridge, Mass.: Harvard University, 5 January 2003 (http://gking.harvard.edu/)).
61. Gary King, Michael Tomz and Jason Wittenberg, 'Making the Most of Statistical Analyses: Improving Interpretation and Presentation', American Journal of Political Science, 44 (2000), 347-61.
62. William L. Miller, Electoral Dynamics in Britain since 1918 (London: Macmillan, 1977);
63. Jonathan Katz and Gary King, 'A Statistical Model for Multiparty Electoral Data', American Political Science Review, 93 (1999), 15-32;
64. King et al., 'Making the Most of Statistical Analyses'.
65. Nicholas J. Cox of the University of Durham has written a very helpful Stata module (-triplot.ado-) to draw such ternary plots.
66. Even if left supporters were brought to abstain from the election because they had been persuaded that cohabitation was a bad idea, this had an indirect effect on the election outcome because it would have weakened the political left. See Thomas Gschwend and Dirk Leuffen, 'Stuck between a Rock and a Hard Place: Electoral Dilemmas and Turnout in the 2002 French Legislative Elections', in Michael Lewis-Beck, ed., The French Voter: Before and After the 2002 Elections (Basingstoke, Hants.: Palgrave, 2003), pp. 155-77.
67. That, indeed, can become politically relevant. In 2002, public opinion towards cohabitation strongly declined in just a few weeks. Compare the Louis Harris polling results about cohabitation published in Libération on 29 March 2002 with the ones published on 7 March 2002.
68. Morris Fiorina, Divided Government (New York: Macmillan, 1992).
69. Fiorina, Divided Government, pp. 72-81.
70. Fiorina, Divided Government, p. 76. Contrary to the case of France, in the United States electors cast their vote for president and Congress at the same time, at least in presidential election years. Scholars of divided government in the United States, therefore, often focus on ticket-splitting behaviour as an indication of divided government voting.
71. This difference is significant given its 95 per cent confidence interval.
72. Fiorina, 'An Era of Divided Government'.
73. Compare Gérard Grunberg, 'Du cohabitationnisme de l'opinion', Pouvoirs: Revue française d'études constitutionelles et politiques, 91 (1999), 83-95, with the survey data published on 11 June 2002 in Libération. There, 26 per cent of the voters declared to have voted to give a coherent majority to the president. However, the validity of this information should be handled with care, since the respondents were able to state three reasons, and right and left supporters were not differentiated.
74. Michael S. Lewis-Beck and Richard Nadeau, 'Split-Ticket Voting: The Effects of Cognitive Madisonianism', Journal of Politics, 66 (2004), 97-112.