Abstract
This paper uses peer comparisons to analyze the impact of different types of inequalities (i.e., within-group, between-group and overall inequality) and reference income on subjective happiness. This study contributes to an emergent strand of research in which both relative levels of economic resources and the income distribution are regarded as determinants of individual happiness. The empirical findings show that overall and within-group inequality negatively affect individual happiness, inequality between reference groups does not affect happiness, and a higher average income of the reference group increases individual happiness. We examine whether people’s aversion to inequality is conditional on their income position within reference group and institutional differences across European countries. These tests indicate that an increase in inequality or a decrease in average income decreases the happiness of both the rich and the poor. Regarding the differences across countries, people who live in more mobile societies with better welfare systems (e.g. Social-Democratic countries) are less adversely affected by inequality than people living in countries with low social mobility and ineffective systems of social protection (e.g. the Mediterranean countries). The analysis is based on data from the European Quality of Life Survey.
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Kahneman (1999), Frey and Stutzer (2004) and Gruber and Mullainathan (2005) are several of the most explicit analyses of the relationship between happiness and utility. See Kimball and Willis (2006) for a discussion of the similarities and differences between utility and happiness. As is common in this literature (e.g., Frey and Stutzer 2000), this paper uses the term “happiness” interchangeably with the terms “satisfaction with life” and “subjective well-being” for the sake of simplicity.
See Alvarez-Cuadrado and Ngo (2012) for a theoretical model of the impact of positional concerns regarding the income distribution.
Corak (2013) provides possible support for the hypothesis that Mediterranean and Social-Democratic countries are on the opposite sides of a European ranking based on social mobility. In his classification of intergenerational earnings mobility, Scandinavian countries emerge as the most mobile societies, while a “Mediterranean” country (Italy) occupies the worst position among the European countries. Again, countries such as Germany and France, for which we obtain mixed results, also occupy intermediate positions in Corak’s ranking.
Falk and Knell (2004) present a theoretical model in which the reference group is endogenous.
Educational attainment is considered preferable to other traits (e.g., income or marital status) because, although education attainment may be affected by student happiness, and both of them can be affected by the same genetic traits, we believe that among the other potential endogenous criteria for clustering, educational attainment is both reasonably stable during adult life and also able to control for other factors that are not directly observable in our dataset (e.g., an individual’s skills, family background).
To ensure that the results are robust to outliers and to make reliable estimates within reference groups, we fix a threshold of the minimum reference group size at 50 individuals. Accordingly, the number of reference groups decreases from the potential (3 × 4 × 3 × 30 =) 1080 to 771, because 309 reference groups contain fewer than 50 individuals. These exclusions also decrease the total sample size by 8.97% (i.e., from 81,060 to 73,786). However, we find that the results are robust to other minimum thresholds (e.g., 20). We also define reference groups further in terms of gender. Although the results are qualitatively the same, the total sample size decreases by 39.77% (i.e., from 73,786 to 44,440) and generates sample bias because the excluded reference groups are concentrated in small countries. That is, we opt to omit gender as a criterion to define a reference group. These robustness checks are available on request from the corresponding author.
The EQLS happiness question is as follows: Q41: Taking all things together on a scale of 1–10, how happy would you say you are? Here, 1 means you are very unhappy and 10 means you are very happy.
It measures an individual’s real access to economic resources and is usually considered a key determinant of happiness because it is required to meet basic needs. The income used in this analysis is the monthly net household income converted through the OECD equivalence scale. It assigns a value of 1 to the first adult in the household, 0.5 to each remaining adult, and 0.3 to each child. The adjective “real” means that income is converted into euros as of the year 2000. This transformation makes comparable individual income given differences in household size and economies of scale, currencies and inflation rates across waves and countries.
The generalized entropy class of indices is a decomposable class of income inequality measures that are sensitive to a parameter a. If a is close to zero, the index is more sensitive to changes at the lower end of the distribution; it is equally sensitive to changes across the distribution for a=1 (Theil index), and it is sensitive to changes at the higher end of the distribution for higher values (Shorrocks 1984). Moreover, the three parameterizations of a are special cases of income inequality metrics; in particular, Ge(0) is the mean log deviation, Ge(1) is the Theil index, and Ge(2) is half the squared coefficient of variation.
Because Gini is not a decomposable measure of income inequality, we do not report this index.
The countries are grouped by extending the classification of welfare systems, as proposed by Esping-Andersen (1990), Ferrera (1996) and Fenger (2007), as follows: Mediterranean (Cyprus, Greece, Italy, Malta, Portugal, Spain, Turkey); Conservative (Austria, Belgium, France, Germany, Luxembourg, the Netherlands); Social-Democratic (Denmark, Finland, Sweden); Liberal (Ireland, the UK); CEE (Bulgaria, Croatia, Czech Rep., Estonia, Hungary, Latvia, Lithuania, Macedonia, Poland, Romania, Slovakia, Slovenia).
However as a robustness check, we also estimated all the regressions by Ordered Probit, and no relevant differences in coefficients or significance were observed.
These broad definitions of “rich” and “poor” allow us to maintain the sample size.
Note that in a benchmark regression with only controls (see Table 8 in Appendix 2), education levels positively affect individual happiness; however, these effects became statistically insignificant when we include in the model specification indexes of income inequality or a measure of relative income.
We address this caveat by clustering standard errors at the level of reference groups.
An alternative rationale for this result is also possible: in countries with inadequate social mobility (e.g., Mediterranean countries), higher inequality means that economic disparities will remain wide in the future; thus, given that inequality is a form of income concentration, the poor are more numerous than the rich, and thus, the estimated overall effect is guided by the disadvantageous prospects for poor peoples’ incomes, namely, they are likely to continue to be poor in the future \( (\beta_{1}^{Med} < 0) \). Otherwise, in more mobile societies or those with generous welfare systems, income inequality does not affect individual happiness because the overall effect, which is always led by the poor, indicates a greater likelihood that their position in the economic ranking could change in the future.
With exclusion of model IV.6, where \( \alpha_{1}^{{\left( {P - R} \right)}} > 0 \).
This result does not hold for \( P_{i}^{j} \sigma_{j}^{whitin} \), which exhibits statistical significance at the 5% level. However, when we apply the Probit estimator, this coefficient is no longer statistically significant (p value 0.196).
This result does not support the findings of Satya and Guilbert (2013), which are based on Australian data. In their research, an increase in peer group income harms the poor more than the rich.
Alternatively, we could assume that these two opposite effects occur but cancel one another out.
Ferrer-i-Carbonell (2005) appears to indirectly validate this supposition, since her estimates on Germany change sign and statistical significance when the full sample is split between East and West Germany.
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We would like to thank anonymous reviewers for providing insightful comments and suggestions. Any remaining errors or inaccuracies are, of course, our own.
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Appendices
Appendix 1: Variable definitions
The econometric analysis is based on three cross-sections (waves 2007, 2011 and 2016) of the Integrated European Quality of Life Survey (EQLS) dataset.
European Foundation for the Improvement of Living and Working Conditions (2018). European Quality of Life Survey Integrated Data File, 2003–2016 [data collection], 3rd Edition. UK Data Service, SN: 7348. http://doi.org/10.5255/UKDA-SN-7348-3
The original dataset collects variables at the individual level for 31 countries (27 EU Member States, Croatia, FYR Macedonia, Turkey and Norway) in the 2007 wave, 34 countries (27 EU Member States and Croatia, Iceland, FYR Macedonia, Montenegro, Serbia, Turkey and Kosovo) in the 2011 wave; and 33 countries (28 EU Member States and Albania, FYR Macedonia, Montenegro, Serbia, Turkey) in the 2016 wave. We include only the 30 countries surveyed in all the three selected waves: the EU 28, FYR Macedonia and Turkey. The survey is publicly available, and all the information can be found on the EQLS website (http://www.eurofound.europa.eu/surveys/european-quality-of-life-surveys-eqls).
Appendix 2: Control variables
In this appendix we show the estimates of control variables for Tables 2, 3, 4 and 5 in Tables 8, 9, 10 and 11, respectively.
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Amendola, A., Dell’Anno, R. & Parisi, L. Happiness and inequality in European countries: is it a matter of peer group comparisons?. Econ Polit 36, 473–508 (2019). https://doi.org/10.1007/s40888-018-0130-6
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DOI: https://doi.org/10.1007/s40888-018-0130-6