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Political Generations and Electoral Change in Canada

Published online by Cambridge University Press:  27 January 2009

Richard Johnston
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Abstract

This article lays out the elementary logic of age structures in party preference data and proposes a simple estimation model with demographic and historical elements. As voters age their preferences intensify. But they do not intensify much and generational differences in the direction of party preferences are correspondingly weak. The Canadian electorate does not seem all that strongly anchored by the accumulated experience of the individuals that make it up. The major source of long-term electoral change, therefore, is conversion in the existing electorate. Consideration is given to how distinctive the Canadian pattern is.

Type
Articles
Information
British Journal of Political Science , Volume 22 , Issue 1 , January 1992 , pp. 93 - 115
Copyright
Copyright © Cambridge University Press 1992

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References

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16 The image is from LeDuc, , ‘Canada: Politics of Stable Dealignment’.Google Scholar

17 An interpretation like this has been placed on American data by Crittenden, J., ‘Aging and Party Affiliation’, Public Opinion Quarterly, 26 (1962), 648–57CrossRefGoogle Scholar; and by Knoke, D. and Hout, M., ‘Social and Demographic Factors in American Party Affiliation’, American Sociological Review, 39 (1974), 700–13CrossRefGoogle Scholar. It is also conceivable that Liberal identification is subject to a life-cycle effect. Clarke, H. D., Jenson, J., LeDuc, L., and Pammett, J. H., in Political Choice in Canada (Toronto: McGraw Hill-Ryerson, 1979)Google Scholar and again in Absent Mandate: The Politics of Discontent in Canada (Toronto: Gage, 1984)Google Scholar found that newly-minted voters were typically more Liberal than the pre-existing electorate. Sometimes this pattern contributed to whole-electorate shifts; sometimes, it dampened those shifts. But the fact that the Liberal share in the whole electorate was not displaced over the period Clarke et al. examined suggests that many of these young Liberals – if they were reporting their behaviour truthfully to begin with – eventually moved on to other parties.

18 Särlvik, B. and Crewe, I., Decade of Dealignment (Cambridge: Cambridge University Press, 1983), pp. 91–3, 110Google Scholar. The Särlvik–Crewe verdict is mixed: abiding age differences are weak, so much so that the variable was dropped from their tree analysis, but younger voters exhibited more swing between 1974 and 1979 than older voters.

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22 Most pointedly by Erikson, R. S. and Tedin, K. L., ‘The 1928–1936 Partisan Realignment: The Case for the Conversion Hypothesis’, American Political Science Review, 75 (1981), 951–62.CrossRefGoogle Scholar

23 ‘Political Immunization’, p. 159. Emphasis in original.

24 This strategy is associated with Mason, K. O., Mason, W. M., Winsborough, H. H. and Poole, W. K., ‘Some Methodological Issues in Cohort Analysis of Archival Data’, American Sociological Review, 37 (1973), 242–58.CrossRefGoogle Scholar

25 See, in particular, Rodgers, W. L., ‘Estimable Functions of Age, Period, and Cohort Effects’, American Sociological Review, 47 (1982), 774–87.CrossRefGoogle Scholar

26 The categories are, in increasing partisan intensity: no identification with a party in response to the basic item; ‘not very strong’ identification; ‘fairly strong’ identification; and ‘very strong’ identification. These are slightly different categories from the American ones. ‘Leaners’ in the American sense are few and seemed best left in the no-identification category.

27 For the natural-age estimation, β3 itself represents the slope. For the logarithmic transformation the estimated slope is β3/Age; as age increases, the slope must decrease. In the logarithmic estimation gives the age at which partisan intensity is zero.

28 The 1988 survey is not included here. The 1988 study employed a different sampling frame: where earlier samples were clustered 1988 approximated to a simple random sample. In addition, the wording of the 1988 identification item differed from the earlier studies. Until the impact of the changes is better known it is unwise to merge 1988 with its predecessors. For more detail on the 1965–84 studies, see the Appendix.

29 Multicollinearity remains a problem for demographic variables, even with the restrictions we have placed on coefficients. Of the ten demographic variables (age plus nine dummies), the cohort dummies are the most collinear with the rest of the bloc and the period variables are the least collinear. This may help to explain the relatively large standard errors on cohort coefficients and relatively small standard errors on the period ones. But the collinearity cannot explain differences in coefficients themselves. The power of period effects may reflect peculiarities in each study as much as anything intrinsic to the elections. For instance the large coefficient on E1980 probably reflects the fact that that sample is a proper subset of the 1979 sample, purged of many apolitical respondents by panel mortality (see Appendix). Note, however, that the statistical robustness of age coefficients survives in the face of high collinearity. In a re-estimation (not reported in a table) with all other demographic variables dropped, the age coefficients remained exactly as presented in Table 1, although with much smaller standard errors.

30 Evidence from official returns indicate that only in British Columbia was the Liberal party worse off than it had been before the advent of Mr Trudeau.

31 Converse, , Dynamics, pp. 51, 57Google Scholar; Abramson, , ‘Developing Party Identification’, pp. 89–90n.Google Scholar