Abstract
The growth of Euro-scepticism in the last decade and a half would seem to suggest a decline in public consensus on the European Union (EU) — that is, a decline in the congruence of individual's attitudes. Yet, Euro-scepticism may simply reflect movement of the distribution in the issue space, as opposed to a change in the shape of the distribution. The difference between these perspectives is important because while a shift in the distribution has implications for the EU, a change in the shape of the distribution has implications for both the EU and domestic politics. To address the issue, we examine cross-temporal variation in the mean, variance and kurtosis of public opinion distribution on the EU — that is, the shape of the distribution. Unsurprisingly, opinions are more dispersed and the distributions flatter in the 1990s, relative to the 1980s. Yet, when viewed over the long run the extent of the decline in consensus is far less marked. Indeed, we find cross-national convergence in opinion in the last three decades, and, evidence of greater attitudinal polarization in the 1970s than in the early years of the new millennium.
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Notes
Indeed, recent issues of both Acta Politica (Volume 42, Number 3, September 2007) and European Union Politics (Volume 8, Number 1, March 2007) have been entirely devoted to the subject.
The growth of Euro-sceptical parties and party factions may have driven the movement of the distribution, may be a response to that movement, or, as Steenbergen et al. (2007) argue, reflect reciprocal causation.
Kurtosis is defined as: K=Σ(X−μ)4/Nσ 4−3, where μ is the mean and σ is the standard deviation.
This assumes the standard ‘− 3’ correction included in the above formula. That is, a correction to allow a normal distribution to evince a 0 kurtosis.
A better indicator of attitudes towards the EU would be the question addressing whether respondents are for or against efforts to unify Western Europe. This question, however, ceased to be asked after 1995. In consequence, we lack significant post-Maastricht data. This is important because we seek to identify whether the pre- and post-Maastricht periods display similar trajectories in terms of changes in the distribution of attitudes.
The P-values are derived from an F-test of the hypothesis: [α+β 11991]−[α+β 11992+β 2+β 31992]=0.
The p-values are derived from an F-test of the hypothesis: [β 1+β 3]=0.
East Germany (sampled separately in our data) became a member upon unification, but there are not enough data prior to 1992 to justify including the full model with interaction terms. In East Germany, from 1990 to 2002, support for integration significantly declined.
As can be seen from Table 1, the slope coefficient on the Netherlands is significantly smaller than that of the other countries displaying a systematic pre-Maastricht increase in average support for the EU.
As noted earlier, given the relative lack of data for East Germany in the pre-Maastricht period (i.e. two data points) comparing this period with the post-Maastricht period and addressing the issue of a Maastricht effect seems neither sensible nor warranted. The sign and the significance of the coefficient for year are unaffected by whether or not the interaction terms are included.
Needless to say, the nature of inter-party competition and institutional features of the political system constitute critical intermediate variables that may serve to either advantage or disadvantage Euro-sceptical parties, quite apart from the distribution of public opinion.
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We thank Liesbet Hooghe, Gary Marks, and the anonymous reviewers for their helpful comments.
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Down, I., Wilson, C. From ‘Permissive Consensus’ to ‘Constraining Dissensus’: A Polarizing Union?. Acta Polit 43, 26–49 (2008). https://doi.org/10.1057/palgrave.ap.5500206
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DOI: https://doi.org/10.1057/palgrave.ap.5500206