units are rarely considered in this research tradition (Doggan and Rokkan, 1969; .... Dogan, Mattei and Stein Rokkan, eds. 1969. Quantitative Ecological ...
Class power, class mobilization, and class voting: The Canadian case* Robert J . Bryrn Michael W . Gillespie Rhonda L. Lenton Abstract. Neo-Mamist class categories, like conventional SES measures, generally predict only a
small proportion of the variation in Canadian class voting. This has led some social scientists to downplay the role of class factors in shaping voting behaviour, some political scientists have gone so far as to abandon sociological models of voting behaviour altogether. In this paper we argue, f i t , that emphasizing the proportion of explained variance is a methodological fetish. It thus makes little sense to abandon particular models of voting behaviour on the grounds that explained variance is low. Second, in extending research on the relationship between class and voting, we criticize previous research for ignoring the complex social processes by which objective class position may or may not be transformed into subjective class action. We contend that studying how classes become politically mobilized or demobilized draws attention to a set of structural variables that markedly increase our confidence in the existence of a relationship between class and voting and therefore our understanding of this relationship. We substantiate our argument by performing logistic regression analyses on data from the 1968 Canadian Federal Election Survey and other sources. RCsumC. La facon dont les classes sont conceptualis&s par les nto-marxistes n'est en gtntral
capable de pr6dire qu'une faible proportion de la variation du vote classiste canadien. En cela cette conceptualisation s'apparente aux variables conventionnelles de statut socio-Cconomique. Cela a entrain6 certains analystes b minimiser le rble de facteurs classistes dans la formation du comportement tlectoral. Cet article s'inscrit en faux contre une telle ligne de penste. Nous suggtrons d'abord que l'attention excessive accord& h la proportion de variance expliqute reltve d'un fttiche mtthodologique; en constquence, il est abusif d'abandomer certains modtles de comportement tlectoral simplement h cause de la faible proportion de variance expliqute. Ensuite, nous explorons plus avant la relation entre classe et vote; et nous critiquons l'absence, dans les recherches anttddentes, de considtration des processus sociaux complexes par l'intermtdiaire desquels la position de classe objective peut se transformer en action de classe subjective. Nous soutenons que l'ttude de la f a ~ o ndont les classes deviement politiquement mobilistes met en Cvidence un ensemble de variables structurelles qui renouvelle notre confiance en l'existence d'une relation entre classe et vote. Notre compdhension de ce phtnomtne en est ainsi amtlior&. Nous justifions cette interpdtation au moyen d'une sBrie d'analyses logistiques de rtgression, sur les domtes du Sondage des tlections fatrales canadiemes de 1968, ainsiqu'h l'aide d'autres sources.
* Most of the data utilized in this paper were made available by the Inter-University Consortium for Political and Social Research. The data for the 1968 Canadian Federal Election Survey were originally collected by John Meisel of Queen's University. Neither the original collector of the data nor the Consortium bear any responsibility for the analyses or interpretations presented here. The authors gratefully acknowledge the financial assistance of the Social Sciences and Humanities Research Council of Canada in the preparation of this paper. We also thank Doug Baer, John Fox, Ron Lambert, Lawrence LeDuc, John Myles, Mike Omstein, Jon Pammett, Michael Shalev, and the CJS anonymous reviewers for helpful comments on earlier drafts.Please address all correspondence and offprint requests to Professor Robert J. Brym, Department of Sociology, University of Toronto, Toronto, Ontario, M5S 1Al; Professor Michael W. Gillespie, Department of Sociology, University of Alberta, Edmonton, Alberta, T6G 2H4; or Professor Rhonda L. Lenton, Department of Sociology, McMaster University, Hamilton, Ontario, L8S 4L8. Canadian Journal of Sociology/Cahiers canadiens de sociologie 14(1) 1989
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The "failure" of class voting models The use of neo-Marxist class categories has become an increasingly popular alternative to SES measures. This is partly a result of changes in the intellectual climate of Westem sociology. It is also a consequence of the fact that the predictive power of neo-Marxist class categories has often proven to be superior to that of conventional SES measures. For example, Erik Olin Wright (1979) has developed a schema of six discrete class categories based on ownership versus non-ownership of the means of production and control versus non-control over the physical process of production and the labour power of others. This categorization has been used to explain larger proportions of variance in, for example, wage attainment (Omstein, 1983) and delinquency rates (Hagan et al., 1985) than can SES measures. In voting studies, however, the benefits of using neo-Marxist class categories have been much less evident. For example, in Canada (contrast Andersen, 1984) researchers report either slight gains in explained variance (Hunter, 1982; Zipp and Smith, 1982) or no gains at all (Ornstein et a1.,1980). Neo-Marxist class categories, like conventional SES measures (Brodie and Jenson, 1980; Clarke et al., 1979,1984; LeDuc, 1984; Pammett, 1987; Larnbertet al., 1987), continue to predict only a negligible proportion of the variation in Canadian voting; and Canada continues to have one of the lowest levels of class voting among the advanced capitalist countries (Myles, 1979). As Alfred Hunter (1982: 34) notes, class voting in Canada seems to be "a mixture of socio-economic and class effects which, when assessed simultaneously, generally amount to very little." Indeed, in Canada individual-level sociodemographic variables in general are very poor predictors of voting preference. In 1974,1979, and 1980, for instance, SES, region, religious affiliation, ethnicity, community size, age, and sex together explained only about 10-15 percent of the variance in the Liberal and Progressive Conservative vote and about 5 percent of the variance in the NDP vote (Clarke et al., 1979: 124-7; LeDuc, 1984). This state of affairs has led some Canadian political scientists to abandon sociological models of voting behaviour altogether. As one leading political scientist put it, "Canada remains a country in which socioeconomic or social cleavage models are basically unsatisfactory. . . . Better explanations of voting choice must be sought among psychological predictors or among the myriad of short-term forces that affect electoral behaviour and election outcomes" (LeDuc, 1984: 408). Theoretical considerations In this paper we take a different view. We argue that much of the disillusionment with sociological models of voting behaviour derives from amisplacedemphasis on the significance of R2as a summary measure of the adequacy of such models. Moreover, we show that the repeated finding of a trivial relationship between class position (however measured) and voting preference is the outcome of a dubious assumption -namely that there is a straightforward and structurally
unmediated connection between class and voting. The bulk of our theoretical discussion deals with this assumption. We return to the question of RZlater. The assumption of a structurally mediated connection between class and voting is problematic because it ignores the complex social process by which objective class position may or may not be turned into subjective class action. In contrast, we contend that by studying how classes become politically mobilized or demobilized (Griffin et al., 1986), attention is drawn to a set of variables that markedly increase our understanding of class voting. Specifically, political mobilization -the process by which a Klasse an sich is transformed into a Klasse fur sich - and political demobilization - the opposite transformation -depend heavily on the relative power of the class in question (Brym, 1979, 1986a, 1986b, 1986c, 1986d; Bryrn and Neis, 1978; Korpi, 1974,1983; Korpi and Shalev, 1980; Shalev and Korpi, 1980). Power is a function of resource control, and in capitalist societies a class's access to resources is contingent above all else on the relationship to means of production common to its members. In a given time and place, firm limits on a class's access to resources are set by (a) property ownership versus non-ownership and (b) control versus non-control over the physical process of production and over the labour power of others. There are, however, wide longitudinal and crosssectional variations in class relations that result in different balances of power. For example, some forms of capital investment help create large, highly concentrated, and socially solidary working classes within which lines of communication are strong and organizational efforts may proceed with comparative ease. Here, unionization and involvement in independent workers' parties tend to be more widespread than in circumstances where capital investment has failed to increase workers' access to politically valuable resources. In general terms, the resources that constitute class power include, in ascending order of importance: (a) the size of the class and its allies; (b) the density of social ties among its members; and (c) normative, material, and coercive resources (Bierstedt, 1974). For purposes of illustration, consider a simplified two-class case. Assume, furthermore, that it is possible to measure the ratio of superordinate to subordinate class power. One would predict that the likelihood of members of the subordinate class supporting a party that represents their class interests would vary inversely with the magnitude of this ratio. That is, relatively powerful superordinate classes are able to allocate abundant resources to fashioning ideological and political systems within which independent, subordinate class parties have great difficulty flourishing. Conversely, relatively powerful subordinate classes possess sufficient resources to forge ideologies and parties that better represent their interests than do the ideologies and parties of the superordinate class.' One deviation from the ideal case sketched above is especially important in Canada. Colonial or quasi-colonial experiences can colour the political ideologies and practices of subordinate and superordinate classes with nationalist
overtones. The two outstanding examples of this in Canada are, of course, Quebec and Alberta. Separatist and autonomist movements and parties have been exceedingly important in the political histories of both provinces. Until very recently, such movements and parties have tended to draw the attention of subordinate classes in those provinces away from democratic socialism. It therefore seems reasonable to expect that, in a given province, the association between (a) the balance of power between subordinate and superordinate classes, and (b) the propensity of subordinate classes to vote NDP, should be attenuated to the degree that separatist and autonomist movements and parties have shaped the history of that province. Our general argument, however, is that class voting is not a simple function of class position. Between class structure and voting preference lie a set of power relations that determine the degree to which voters are able to translate their class interests into class voting. It follows that variations in these power relations should be systematically related to the magnitude of the class vote. Previous research on contextual effects Our study uses measures of the relative power of subordinate classes to predict NDP vote. A few Canadian researchers have recently employed contextual variables in voting analyses. Their findings may be interpreted in a way that is entirely consonant with the theoreticalargument of this paper, and they are worth reviewing here. Linda Gerber (1986) analyzed constituency effects in the electoral districts of Ontario. She discovered that the class composition of tidings was a powerful predictor of party strength. Net of urbanity, ethnicity, and religion, the socioeconomic composition of constituencies in 1968 accounted for over 61 percent of the variation in the Progressive Conservative vote, 41 percent of the variation in the Liberal vote and more than 72 percent of the variation in the NDP vote. Similarly, Donald Blake (1978) has shown how the predominance of nonmanual workers in a riding weakens and deflects the non-Liberal voting predispositions of manual workers (see also Brodie and Jenson, 1980: 3 10-11). These findings support our view that the relative size of the working class in a given context has a bearing on the likelihood that people will support the NDP. In such circumstances, we surmise, working-class institutions and culture help legitimize democratic socialism so that workers (and, to a lesser degree, some non-workers) are more likely to vote NDP than is the case where workers comprise only a small proportion of the population. 1. Two important questions are not dealt with here. First, we ignore the manner in which purely "conjunctural" circumstances -that is, causes historically unique to our time frame - affect electoral contests. Second, we pay no attcntion to the way the "relative autonomy of the state" -which we understand to mean the institutional rcsidues of past class and other group conflict, as embodied in electoral and industrial relations systems, for example -helps shape current electoral contests.
Other research on contextual effects emphasizes the importance of organizational resources in determining level of support for the NDP. It is well known that union membership increases the propensity of workers to vote NDP (Keddie, 1980). But Keith Archer (1985) demonstrates an interesting contextual effect of union membership: if union locals are NDP affiliates the probability of workers supporting the NDP triples. Moreover, it has been shown that the proportion of workers and farmers2 organized in unions and cooperatives in each province influences the size of the NDP vote (Brym, 1986b, 1986c; see also Lambert et al., 1987). Similarly, the constituency strength of the NDP has an independent effect on support for that party. Where the NDP is traditionally strong, workers are more likely to support the party than where it is traditionally weak (Zipp and Smith, 1982; see also Irvine, 1976; Ogmundson, 1976; Pinard, 1973; Shiry, 1976). In short, if workers have access to organizational resources - notably trade unions and a viable democratic-socialist political party - they are considerably more likely to vote NDP than where these resources are lacking. Finally, it has been demonstrated that a very crude measure of the relative access of workers in each province to material resources is associated with the size of the NDP vote (Brym, 1986b, 1986~).This finding, too, lends credibility to the argument presented above.
Data Most of the studies that employ contextual variables to predict votingpattems are deficient in that they rely only on data aggregated at the constituency or provincial level. Thus, the patterns of individual-level behaviour that produce aggregate-level patterns are unclear. In order to overcome this problem we reanalyze data from the 1968 Federal Election Survey in combination with aggregate (provincial-level) data. The 1968 survey was selectedbecause it is the first Canadian voting survey that contains questions on organizational affiliations. In future work we intend to reanalyze more recent surveys. Provinciallevel (rather than, say, constituency-level) aggregate data were employedfor two reasons. First, for some of the important contextual variables that interest us, data aggregated in units smaller than provinces are not available. Second, it is evident that provinces are exceedingly important socio-political milieux in Canada. Even at the federal level, provinces are a major basis of Canadian political organization; the structure of the federal government and of federal political parties clearly reflect this (Stevenson, 1982 [1978]). Moreover, survey data show that there are strong inter-provincial differences in many political attitudes (Omstein et al., 1980). This is not to suggest that constituency-level effects are negligible -in due course we intend to examine these as well, especially for the 1984 Canadian Federal Election Survey -but only that provincial-level effects are likely to be significant. 2. Farmers are included here because they have traditionally played a very important role in Canadian democratic-socialist politics, at least in some paxts of the country.
We shall restrict ourselves to predicting whether respondents voted for the NDP in the June 1968federal election. We examine only voters. After a decade of debate, a measure of consensus appears to have emerged that the NDP is the only Canadian federal political party that can unambiguously be considered moderately left-wing (Brym, 1986a: 57-8; Castles and Mair, 1984). Accordingly, we predict that voting NDP will tend to be an expression of subordinate class interest. Class will be measured in the only ways afforded by the 1968 Canadian FederalElection Survey: according to a crude occupational scale and by income. We would have preferred measuring class using Wright's (1979) neo-Marxist schema as well. However, the limitations of the data precluded this possibility. All other individual-level sociodemographic variables are measured in an obvious manner, as will be evident from our discussion. We include two individual-level measures of access to resources in our analysis. The first is trade union membership, which taps access to organizational resource^.^ The second measure taps access to material resources. It is the product of income times union membership. We predict that the effect of union membership on voting NDP will increase with the union member's income. (We also include the main effect of income both on statistical grounds and on the grounds that income will reduce the propensity to vote NDP among people who do not belong to unions.) Our reasoning for specifying an interaction between income and union membership is twofold. First, the higher the income of its members, the more material resources the union has at its disposal for political action. Under this interpretation, the income of union members provides a base for working-class power. Second, the income of union members also partly reflects the unions' power in the sense that powerful unions are better able to negotiate favorable contracts for its members. Sincethese interpretations are not mutually exclusive, we suspect that both processes are at work. Our final set of variables consists of contextual variables in which we take the province of residence as the context. Following Brym (1986b), we use two variables to measure working-class resources at the provincial level: the proportion of the provincial population who belong to unions and the proportion of the provincial population who belong to cooperatives. We predict that propensity to vote NDP will increase as either proportion increases, even after controlling for the individual effects of union membership. Moreover, we predict that the variation of these two variables will capture most of the inter-provincial differences in the propensity to vote NDP. Also following Brym (1986b), we introduce two dummy variables for residence in Quebec and residence in Alberta, respectively. The separatist and autonomist movements in these two provinces reduce the proportion who vote .
.-
.. -
-
-
.
..
-
3. We had intended to study the effects of cooperative membership. Unfortunately, the number of respondents who reported membership in a cooperative was too small to permit estimation of its effect.
30
NDP from the proportion expected given the levels of union and cooperative membership in these two provinces. To capture the greater strength of the separatistJautonomist movement in Quebec, we specify interaction terms between residence in Quebec and the individual-level variables income and union membership. Before turning to our regression analysis, let us consider the problems that result when voting research relies too heavily on R2.
T h efetish of R2
As noted above, many researchers dismiss the existence of a significant class vote in Canada on the grounds that their models yield a low R2.We argue against this reasoning on two grounds. First, there are technical problems associated with measuring explained variance in the case of a dichotomous variable. Second, the proportion of explained variance is sometimes a poor criterion for judging the adequacy of structural equationmodels regardless of how the dependent variable is measured. Let us discuss the second issue here. We defer discussion of the technical problems to the section on data analysis. Duncan (1970,1975) has long been a vociferous critic of the overuse of R2,4 and King (1986) is even more critical in a recent article. Both argue that in order to judge the effect of one variable on another, researchers should look to the "effect" coefficient, the unstandardized slope. The importance of a variable depends on the magnitude of change in the dependent variable produced by a unit change in the independent variable. The reason that R2 can mislead is because it is affected by the degree of variation in the independent variable. Even if a set of variables have strong effects, the R2produced by the regression equation will be relatively low if the variation in these variables is low.5 This point is particularly relevant for our study since, as we shall see, the odds onvoting NDP is fairly high for high-income union members who live in a province other than Quebec. Since most respondents fall outside this particular combination of values, the value of the R2 analog that we use is smaller than it would be if the composition of the population (and, therefore, our sample) were, say, half in this particular combination and half distributed among the remaining combinations. Because R2 depends on both the size of the slope and the variation in the independent variable, it is more anchored to the particular context in which the research was done. From an historical point of view, the value of R2 lies in its indication of what was (or was not) responsible for the variation in a dependent variable in a particular context. In contrast, focusing on unstandardized slopes provides researchers with the opportunity for generalizing beyond the specifics 4. Although we couch this discussion in terms of R2, the arguments extend to the overuse of correlations, partial correlations, and standardized slopes (beta weights). 5. For example, assume that proportion of respondents voting NDP for a five-category independent variableis 0.0,0.3,0.5,0.7, and 1.0.If the sampleisevenly distributedamong the five categories (SOthat 20 percent of cases fall in each category), the R2willbe 0.75. The only way that R2could attain unity is if the intermediate values of the independent variable were empty.
of the situation in which the research was done. Assuming that the slopes reasonably portray the causal forces at work (a strong assumption), the researcher can foresee the change in the dependent variable that will be produced by a given change in the independent variable. Since we propose to generalize beyond the particular circumstances of the 1968 federal election, we choose to emphasize unstandardized slopes rather than R2.
Results We restrict our analysis to respondents who voted in the 1968 election. This reduces the sample from 2,767 to 2,366 cases.6In order to preserve the relatively large number of respondents for whom income information is unavailable (n = 161),we adopt the strategy recommended by Cohen andCohen (1983: 275-300) of assigning an arbitrary income value to these respondents and creating a dummy variable that classifies all respondents on the basis of whether income information was available. We include both variables in the regression analysis, along with other independent variables. Although estimates of the effect of income are limited to respondents with valid income information, this strategy allows us to make use of the entire sample in estimating the effects of the other independent variables in the equation. We analyze the data using the logistic regression option of the PROBIT subroutine of SPSSX (SPSS Inc., 1986: 602-20): The statistical superiority of logistic regression for dichotomous dependent variables is well known (Fox, 1984: 302-4). The logistic regression slopes are more difficult to interpret than the slopes from ordinary regression, but we deal with this problem by presenting expected values of the dependent variable, the probability of voting NDP, for different values of the independent variables. Since the SPSSX version of logistic regression divides the logit by 2, we double the regression coefficients given in the output to obtain those reported in Tables 1 and 2. In addition, the subroutine also adds five to the intercept, so we subtract five from the intercept -
6. A somewhat different sample used in earlier analyses yielded essentially the same results as those reported below. In these earlier analyses we dealt with individuals who had not voted by using information on how they would have voted, Limited our sample to people with full-time work, and eliminated respondents for whom income information was unavailable. 7. In logistic regression, the dummy dependent variable is transformed as follows. Let p represent the proportion of cases that fall in a given category of the dependent variable - for example, the proportion who vote NDP. The following quantity, termed the "logit," is then regressed on the independent variables: log[p/(l-p)], where "log" refers to the natural logarithm. In the case of categorical independent variables, the values of p can be calculated from the data. In the case of continuous independent variables, initial values of p are estimated using the results of OLS dummy dependent variable regression. 8. The slope is statistically "significant" at the .05 level when it is at least 1.645 times greater than its standard error. We justify the use of 1.645 on the grounds that 1.645 is the critical value of t for a one-tailed test of the null hypothesis at the .05 level of significance. Since the ratio of the logistic regression coefficient to its standard error only approximates a t-distribution, we use this value only as a rough rule of thumb.
given by the output before doubling it. These tables also indicate the significant slopes with an asterisk (*).' Tables 1 and 2 contain the slopes, intercepts, and R2 analogs for equations that link the logit of voting NDP to a set of individual-level and contextual independent variables. As pointed out above, there are technical problems with using R2 in connection with a dichotomous dependent variable. These stem from the fact that R2 is a product of ordinary regression which, as we also point out above, should not be used in the case of dichotomous dependent variables. With logistic regression, however, we find no agreement on an appropriate R2 analog (Aldrich and Nelson, 1984). We choose a coefficient that shifts the focus from the observed dichotomy to the latent continuous variable (the propensity to vote NDP) that underlies the dichotomy (whether or not the respondent actually voted NDP) (McKelvey and Zavoina, 1976). The latent variable is a linear function of the independent variables and we obtain the expected values of the latent variable from a weighted sum of the independent variables, where the weights are the slopes from the logistic regression. The sum of squares for this variable represents the explained sums of squares for the latent variable. The unexplained sums of squares is the number of cases (n) multiplied by 3.29. This expression follows, first, from the fact that the probit expression norms the variance of the disturbance at 1and, second, from the fact that logistic regression norms the variance of the disturbance at 1.814 in order to minimize the difference between the probit and logistic curve^.^ Table 1 contains two equations for the individual-level independent variables. The individual-level variables are union membership, occupation (represented as a set of dummy variables), income, and valid income (a dummy variable coded 1 for respondents with valid income information and 0 for respondents with missing values on this variable). Equation 2 adds interaction terms represented by the product of union membership times income and valid income, respectively. We label these variables "union x income" and "union x valid income," respectively. 9. If the variance of the disturbance for the probit explanation is 1, the sums of squares for the probit specification will be n. Since the latent variable under the logistic specification is 1.814 times the latent variable under the probit specification, the sums of squares for the disturbance under the logistic specification will be n times the square of 1.814, or n times 3.29. 10. The values of the union variable were recoded in order to make them ordinal with respect to nonmembership since respondents who reported " very active," "active," and "inactive" membership were coded 2, 3, 4, respectively. Even after recoding, results for this variable are difficult to interpret since the four categories are ordinal rather than interval. However, adjusting the values of the union variable according to the category differences following methods recommended by Clogg (1982) does not appreciably change the results. Occupation is represented as a set of dummy variables. The reference category contains mainly housepeople plus a small number of students, service people, and people who refused to provide occupation information. Even though not all the occupation categories differ from the reference category, we keep these dummy variables in all five equations.
Union membership is divided into four categories: union nonmembers (n = 1,949: the bulk of the sample), inactive, fairly active, and very active mernbers.I0 Income is coded as eleven ordinal categories exclusive of missing values. The first category contains individuals whose annual income is less than Table 1. Effects of individual-level variables on voting NDP. Slope Variable Union member Professional Owner Sales Clerk Skilled labour Unskilled labour Farmer Retired Service Unemploy ed Income Valid income Union x income Union x valid income Intercept R2
Equation I .373* ,013 -.630* -1.513, ,044 .406* ,254 -.083 ,287 ,067 ,963 ,013 ,006 -
-2.729* ,074
Equation 2
-349' -3.556* ,077
*Slope significant u the .05 level (1 > 1.645). - - -
$1,000, while the last category contains individuals who are paid $15,000 or more. In order to reflect dollar amounts, we recode these categories in $1,000 units." When we look at the first equation in Table 1,we see that the tendency to vote NDP varies "significantly" only with union membership and three occupational categories (owners, salespeople, skilled labor). First, voting NDP increases with tradeunion involvement. Second, respondents who report occupations classified as skilled labour are more likely to vote NDP than respondents in the reference category.12In contrast, owners and salespeople are less likely to vote NDP than respondents in the reference category. Note that neither reporting income nor the amount reported appears to affect voting NDP. 11. Specifically, we set each category equal to the midpoint in thousand dollar units: .5, 1.5, 2.5, 3.5, 4.5, 5.5, 6.5, 7.5, 9, 12.5, and 17.5. Because the eleventh category is open-ended, the assignment of 17.5 as a midpoint of $17,500 is somewhat arbitrary. Respondents who did not report income are given a value of 6 on the income variable. This corresponds to $6,000 and is close to the income average for those who did report income. 12. Unskilled labourers also are more likely to vote NDP than housepeople (as well as respondents in most other occupational categories), but this difference never surpasses our critical value of 1.645 in either Table 1 or Table 2.
When we turn to equation 2, however, we see that both the amount of income reported and whether the respondent reported income interact significantly withunion membership in their effects on voting NDP.I3 This combination of effects indicates that among those who report income, the effect of union membership increases as the respondent's income increases. The effect of union membership, however, is strongest among those respondents who did not report income.I4 Table 2 adds the contextual variables to the individual-level variables contained in Table 1. In equation 3, the contextual variables consist of dummy variables for seven provinces with residence in Newfoundland, Prince Edward Island, or Nova Scotia as the reference category. In equation 4, we model interprovincial differences in voting NDP with the percentage of the provincial population that belongs to a trade union (percent union members) and the percentage of the provincial population that belongs to a cooperative (percent coop members) (Minister of Industry, Trade and Commerce, 1971; Statistics Canada, 1978; Sullivan, 1970).15We interpret these variables as indicators of the strength of the resources available to subordinate classes. Since these two variables do not completely capture the interprovincial differences, we retain the dummy variables for residence in Quebec and Alberta. Finally, equation 5 adds two interaction terms to equation 4 in order to capture the different effects of union membership and income for residents of Quebec: the product of union times the dummy variable for residence in Quebec (union x Quebec) and the product of income times residence in Quebec (income x Quebec). Looking at the R2analogs, we see that adding the contextual variables to the equation substantially increases our ability to predict voting NDP. The "explained variance" increases from around 7 percent in Table 1 to around 25 percent in Table 2. We alsonotice that the effects of the individual-levelvariables are either unaffected by the addition of the contextual variables or else increase. The exception to this trend is the effect of ownership. Although owners are still less likely to vote NDP than the reference category (and most other occupational categories), the ratio of this effect to its standard error is not statistically significant. The most dramatic increase in regression coefficients occurs in the case of unemployment. However, this probably reflects the instability of the coefficient that results from the small number of unemployed respondents (n = 16). The other major increase is in the negative main effects of income. The absolute value of this coefficient triples from -.041 to -.121. Although the magnitude of this coefficient seems small, the range of the income variable (from 1.5 to 17.5 in thousand dollar units) is much greater than the range of the other (individual-level) variables. 13. We retain the main-effect coefficients, whether significant or not, when they are part of significant higher-order interactions. 14. Since the numbcr of active union members for whom income information is unavailable (n = 12), we make no attempt to interpret this result. 15. These data are available from Robert Brym on request.
Table 2. Effects of individual-level and contextual variables on voting NDP.
Individual-level variables Union member Professional Owner Sales Clerk Skilled labour Unskilled labour Farmer Retired Service Unemployed Income Valid income Union x income Union x valid income Contextual variables New Brunswick Quebec Ontario Manitoba Saskatchewan Alberta British Columbia 70union members % cooperative members Union x Quebec Income x Quebec Intercept R-square
Equation I 1.025* ,054 -.519 -1.759* ,027 .437* ,365 -.338 ,050 ,006 .365* -.086* 1.556* .055* - 1.MO*
Slope Equation 2 .989* ,052 -.505 -1.704* ,049 .483* .331 -.3M ,007 -.017 1.164* -.OM* 1.536* .059* -1.023*
-
-.
Equatron 3 1.068* ,084 -.522 -1.737* ,054 .424* ,341 -.291 ,031 -.017 1.344* -.121* 1.616* .072* -1.106*
* Slope sigmficant at the .05 level (t > 1 645)
In addition to the effects of the individual-level variables, equation 3 shows the interprovincial differences in the likelihood of voting NDP. In equation 4, we represent the bulk of this difference in terms of two contextual variables: percent union members and percent coop members. Both contextual variables have significant effects on voting NDP. However, both Quebec and Alberta are outliers in the sense that the percent voting NDP is significantly less than would be predicted from the scores of Alberta and Quebec on the two variables. Still, the fit of the model, as indicated by the R2analogs, drops only marginally when the two contextual variables are used to capture the differences among the remaining eight provinces. Of the two contextual variables, percent union members has the stronger effect on voting NDP. For example, increasing the percentage of union members
in Ontario by 1 percent would, according to the model represented by equation 5, increase the NDP vote by 10 percent in the case of skilled workers who are active union members and who earned $6,600 annually. In contrast, increasing the percentage of coop members by 1percent would increase the NDP vote for this group by only 1percent. The effect of union membership seems particularly strong and may be overestimated. We arrive at these percentages by raising the constant e (2.72) to the power that is given by a combination of the coefficients in equation 5 and selected values of thevariables. Toestimate the effect of changing the percentage ofunion members for Ontario residents, we set the value of the union variable to 3 (active union members), the value of income to 6.6 (the average income for respondents who reported income), and the value of the dummy for skilled labour to 1(and, therefore, the value for the other occupation dummies to zero).I6The percentage of Ontario residents who are union members is 10.74; the percentage whobelong to a coop is 2.67. With an intercept of -8.309, the expected logit for this group is -.736. Raising 2.72 to this power gives us .478, which represents the odds on voting NDP. Converting the odds to a proportion, we get .323.17 If we increase the percentage of union members from 10.74 to 11.74, the expected odds of voting NDP is .752, and the corresponding proportion who vote NDP is .429. If we increase the percentage of coop members from 2.67 to 3.67 (and leave the percentage of union members at 10.74), the expected odds of voting NDP is .5 10, and the corresponding proportion is .338. Because the regression equation is nonlinear, these differences differ slightly for other values of the independent
variable^.'^ We use this method to construct Table 3, which reports the expected proportion voting NDP for selected values of union membership and provincial residence. The values of the union variable are 1 and 3, nonmember and active member, respectively. The values of the income variable are 4 and 10, which represent $4,000 and $10,000, and lie approximately one standard deviation below and above the mean income of $6,600. We also report the expected proportions for respondents who provided no income information. Finally, we report the expected proportions for Ontario and Quebec residents, respectively. The purpose of this table is to aid the interpretation of the interaction of union membership with the income variables and the interactions of union membership and income with residence in Quebec. Note that the proportion differences for 16. These choices also constrain the other values of the individual-level independent variables: valid income = 1; union member x income = 3 x 6.6 = 19.8; and union member x valid income = I x3=3. 17.The odds onvoting NDP is the ratio of the frequency who votedNDP to the frequency whovoted for another party. To obtain the proportion who voted NDP, we divide the odds by the 1 plus the odds. 18. Since the logits are linearly related to the independent variables, the proportions, which are a nonlinear transformation of the logits, will naturally relate nonlinearly to the independent variables.
Ontario residents apply approximately to the respondents of the other provinces, with, of course, the exception of Quebec.19 For respondents who reported income, we see that the expected proportion who vote NDP is greater for active union members than for non-members. This difference is most pronounced in the case of relatively well-to-do active union members. Active union members who eam $4,000 annually are approximately 9 percent more likely to vote NDP than non-members who earn $4,000annually. In the case of respondents who earn $10,000 annually, this difference increases to 27 percent. This result is interestingsince it suggeststhat organizational Table 3. The expected probability of voting NDP by selected values of union membership, income, and residence. Union membership and income non-member active member $10,000 missing $4,000 $10,000 Residence missing $4,000 Ontario ,139 ,148 ,122 ,585 ,239 .392 Quebec ,023 ,056 ,098 ,044 ,023 ,102
resources (union membership) and material resources (income) combine to influence voting NDP. For Quebec residents, we notice, first, that the proportion voting NDP is generally less than for all categories of Ontario residents. Second, we see that this difference is most pronounced in the case of active unionmembers. For residents outside Quebec, union membership increases the likelihood of voting NDP; in the case of Quebec residents, this difference is limited to respondents who provide no income data. For respondents who earn $4,000, the relationship is n e g a t i ~ eThe . ~ other interesting feature of the second line of Table 3 is that it shows a positive effect of income on voting NDP for both members and nonmembers of unions. If we accept the interpretation that the low NDP vote in Quebec represents the channeling of left-wing political activity into Que'be'cois nationalism, this finding suggests that this channeling occurs more among the low-income segments of the Quebec population. In general, our analysis lends credibility to the power resource theory of voting. For the most part, the properties of provinces earlier observed to affect voting NDP (Brym, 1986b, 1986c) persist at the individual level even when we control for the effects of the corresponding individual-level variables. For all provinces apart from Quebec and Alberta, measures of subordinate class and subordinate class power are associated -significantly and in the predicted 19. Again, the differences are only approximate because the model is linear in the logits and not in the proportions, Note also that interactions such as those found for Quebec might have been found for Alberta if the number of Albertans in the sample were not so small. 20. Although the difference between the proportions is small, this difference is misleading as a measure of the relationship since all the proportions are small. A more appropriate indicator of the relationship is the ratio of the two proportions which, in effect, gives the odds ratio, the measure of the relationship used in log-linear analysis.
direction-with NDPvoting. Aspredicted, the provinces of Quebec and Alberta persist as outliers with respect to our model of class voting. And, again as predicted, Quebec, with its much more pronounced nationalist tradition, is more of an outlier than Alberta: the effects of the individual-level variables are the same for Alberta as they are for the other provinces, while the effects of some important individual-level variables (income and union membership) differ for Quebec residents.
Discussion Since its founding in 1961, support for the NDP has risen slowly and unevenly to just over 20 percent of the popular vote in the 1988 election. Meanwhile, a 1987 Gallup poll revealed that fully 28 percent of Canadians felt that the country should be led "in the direction of socialism."This was up from 17 percent in 1983 and 20 percent in 1984 (Toronto Star, 1987). Increased support for the NDP since the beginning of this decade is particularly noteworthy in Quebec and Alberta. Based on the results of public opinion polls, by-elections, the last provincial election in Alberta, and the last federal election, it seems safe to conclude that popular support for the NDP has approached the national average in both provinces. Quebec and Alberta have thus fallen into line with respect to our model of voting; support for the NDP seems to have become in part a function of the ratio of superordinate to subordinate class power in these two provinces, as disillusionment with separatism and autonomism has spread. While the NDP was clearly not a national party twenty years ago (Meisel, 1975 [1972]: 50-l ), recent developments suggest that it is now a national party as class forces have become more pronounced determinants of voting behaviour. Of course, this conjecture is possible only if one simultaneously takes into account individual-level and contextual determinants of class voting; failure to do so can easily lead one to conclude that class voting, even in recent elections, is virtually non-existent and declining. An atomized view of politics is inherent in the latter, predominant interpretation. This view is widespread because Canadian voting studies have been strongly influenced by the University of Michigan tradition of voting research. Contextual effects were an important focus of the voting studies directed by Bernard Berelson and Paul Lazarsfeld at Columbia University in the 1940s and 1950s (Berelson et al., 1954; Lazarsfeld et al., 1948). But in the Michigan tradition, which soon overshadowed the work at Columbia, individuals are assumed to be bundles of sociodemographic attributes; and these attributes are usually assumed to be the sole sociological determinants of electoral choice. The influence on voting of the attributes of personal networks, neighbourhoods, communities, class-based voluntary organizations such as unions and cooperatives, constituencies, and larger (state- or provincial-level) political units are rarely considered in this research tradition (Doggan and Rokkan, 1969;
Sheingold, 1973). In contrast, we contend that individuals are embedded in a whole panoply of social organizations. The structure of these organizations helps shape electoral choice to some degree independently of the individual's sociodemographic attributes. Especially important in this regard is the manner in which these organizations structure the distribution of power among classes. Rather than suggesting the need to abandon sociological models of voting, the results of our research illustrate well the benefits of extending sociological models of voting by combining the analysis of individual-level and contextual effects that tap this class power dimension. This emphasis on the class power dimension also distinguishes our research from the small body of Canadian literature dealing with contextual effects on voting. Some of this research is atheoretical. We hope to have placed it in auseful analytical framework. The signal theoretical contribution of Canadian research on contextualeffects is the argument that class voting is a strategic choice based in part on left party strength (Ogmundson, 1976; Pinard, 1973; Shiry, 1976; Zipp and Smith, 1982). In this view, there is a latent class vote on the left that is often not translated into an actual left vote due to strategic considerations on the part of individual voters. Specifically, the left class vote is not fully realized when and where an appropriate vehicle -a strong left party -is lacking. It follows that, all else the same, class voting on the left willbe stronger when and where the NDP is a more viable alternative. While in general terms we agree that voting NDP is as much a strategic as a left choice for many voters, we are disturbed by the circularity of this argument: its proponents hold that the left party is weak because left class voting is weak, and left class voting is weak because the left party is weak. What seems to us to be missing here is some sense of why the left party is weak in the first place. We have sought to overcome this problem by arguing that left party strength is related to the distribution of power resources among classes. Our research also extends some of the insights of the largely Swedish or Swedish-inspired work on left party growth focusing exclusively on structural variables (e.g., union strength) as determinants of subordinate class mobilization (Korpi, 1983; Korpi andshalev, 1980; Shalev and Korpi, 1980). We have sought to relate structural variables to the rational calculus of individual voter^.^' In effect, we have made a three-stage argument. 1. As subordinate classes become resource-rich relative to superordinate classes there develops a social base for recruitment to the NDP. 2. The expansion of this recruitment base raises individuals' exposure to the NDP and its milieu, and also makes the party's chances of electoral success look better to voters. (We surmise there is a threshold effect here, not a linear relation~hip.)~~ 21. Note that Korpi and Shalev regard union and party growth as constituting power resources, while we regard union and party growth as the outcome of the distribution of power resources.
3. The effects of 1 and 2 on NDP voting are mitigated by cross-cutting (nonclass) loyalties, such as extreme forms of regional identity (autonomism and separatism) that predate the large-scale mobilization of subordinate classes. The quality of our measures do not permit us to claim that our results strongly support this argument. On the other hand, our results do lend credibility to our theory. As such, they should help dissuade students of voting from giving up the search for class effects.
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