Abstract
In this paper, we present a contribution to the analysis of the relationship between influence/power measurement and utility measurement, the two most popular social objective criteria used when evaluating voting mechanisms. For one particular probabilistic model describing the preferences of the electorate, the so-called impartial culture (IC) model used by Banzhaf, the Penrose formula shows that the two objectives coincide. The IC probabilistic model assumes that voter preferences are independent and neutral. In this article, we prove a general version of the Penrose formula, allowing for preference correlations and biases in the electorate. We use that formula to illustrate, for a spectrum of well-known probabilistic models, how the divergence between the two social objectives impacts the ranking and performances of the voting mechanisms.
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Notes
Since the word Sensitivity can receive many different interpretations, we prefer to use the term total influence, which is used in the mathematical literature on Boolean functions.
As pointed out by one of our referees, Schweizer (1990) should also be listed as an early forerunner.
This criterion typically is used in such works on voting design, which we view as the application of mechanism design to the normative analysis of political institutions.
As was mentioned earlier, utility measurement has received much less attention in the social choice literature than power/influence measurement, which has been investigated since the 1940s. We suspect that this unbalanced attention owns mainly to the Penrose formula, which establishes that the two notions coincide for the IC random electorate. It may also be explained by the quite intuitive relationship between the two concepts: if an individual is more likely to change the outcome, she will do so in a way which will increase her utility, generating a higher utility. As will be seen more clearly below, this intution turns out to be wrong in general.
We thank one of our referees as well as Federico Valenciano for calling our attention to these very relevant and important references.
Laruelle and Valenciano (2008, p. 67) write for instance that: “Considerable vagueness in the specification of the voting situation considered underlies most of the literature on voting power. On such vague grounds the notion of decisiveness has de facto been widely accepted as the right basis for the formalization of a measure of voting power. (...). In spite of this dominant view, some authors have raised doubts as to the relevance of this interpretation of “power” as decisiveness, suggesting as more relevant the notion of satisfaction or success. (...) Nevertheless, in general, the notion of success has usually been either overlooked or considered as just a secondary ingredient of decisiveness. Note that we use in this article a terminology different from theirs: what they call decisiveness, we call influence, and what they call satisfaction or success, we call utility.
The rest of their section, as in Laruelle and Valenciano (2005) paper, also contain, among other things, a clear distiction between ex ante, ex post and interim (conditional) versions of these two notions.
In particular, Laruelle and Valenciano (2005, p. 603) write: “We show how the optimization results on both the egalitarian and the egalitarian ideal can be connected to existing results in the voting power literature, drawing on the work by Felsenthal and Machover”.
This amounts to saying that all \(\left( {\begin{array}{c}n\\ k\end{array}}\right) \) vectors with k 1’s have the same probability.
The term exchangeable is due to de Finetti, while the term strongly exchangeable is ours. Finetti’s theorem asserts that an infinite sequence of \(\left\{ 0,1\right\} \) random variables is exchangeable iff it is strongly exchangeable (see Feller 1966). This equivalence does not hold in general when the sequence is finite. Contributions to voting theory using this model of a strongly exchangeable electorate include Chamberlain and Rothschild (1981) and Good and Mayer (1975).
Note that if \(\lambda \) is neutral and \({\mathfrak {C}}\) is neutral, then \(\Pr _{\lambda }\left[ X:{\mathfrak {C}}(X)=1\right] =\frac{1}{2} \).
See Definition 9 (ii) in Laruelle and Valenciano (2008, p. 54). What we call Influence they call Decisiveness.
Indeed, equation (3) yields:
$$\begin{aligned} Influence\,(i,\lambda ,{\mathfrak {C}})&=\lambda \left[ X:X_{i}=0\,{\text{ and }}\,{\mathfrak {C}}(X_{-i},X_{i})=X_{i}\quad{\text { and }}\quad{\mathfrak {C}}(X_{-i} ,1-X_{i})=1-X_{i}\right] \\&+\lambda \left[ X:X_{i}=1\quad{\text { and }}\quad{\mathfrak {C}}(X_{-i},1-X_{i} )=1-X_{i}\quad{\text { and }}\quad{\mathfrak {C}}(X_{-i},X_{i})=X_{i}\right] \\&=\lambda \left[ X:{\mathfrak {C}}(X_{-i},X_{i})=X_{i}\quad{\text { and }}\quad{\mathfrak {C}}(X_{-i},1-X_{i})=1-X_{i}\right] . \end{aligned}$$A proof for this formula will be provided in Sect. 3 when we will establish a generalized Penrose formula.
In this note, we focus on two criteria: total utility and total influence. Felsenthal and Machover (1998) add majority deficit to these two notions. It can be shown (See theorem 3.3.17 in Felsenthal and Machover 1998) that if the random electorate \(\lambda \) is IC , then the majority deficit is a negative affine transform of total influence. Therefore, \(\succ _{\lambda }^{TI}\)and \(\succ _{\lambda }^{TU}\)are also equivalent to the social ordering of mechanisms attached to the comparison of majority deficits when \(\lambda \) is IC.
Where \(\left\lfloor x\right\rfloor \) denotes the integer part of x.
The fact that even without correlations, the orderings \(\gtrsim _{\lambda }^{TI}\) and \(\gtrsim _{\lambda }^{TU}\) can be quite different, was already noted by Laruelle and Valenciano (2008; see their example 3.1 on p. 68), where they consider the case \(n=3\) and \(p=\frac{3}{4}\).
This term was called “luck” by Barry (1980, p. 338), in his informal equation in making explicit the relation between utility (what he called “success”) and influence (what he called “decisiveness”), that is, “success=luck +decisiveness”. Barry (1980, p. 184) indeed writes that for the member of a committee, “In addition to his power, [the likelihood of outcomes corresponding to his desire] depends on what the outcome would have been in the absence of his intervention. This is what I shall call luck.”
Of course, this does not mean that they will, since as reminded when defining the expected utility of an individual, voting for one’s prefererred alternative is the best a strategic voter can do in this setting.
If one were to follow further the terminology used by Barry (1980), this correction term would be (half) the difference between “luck” (voter i gets her desired outcome even when voting against it) and “bad luck” (voter i does not get her desired outcome even when voting for it).
Recall that the popular IC random electorate is the strongly exchangeable electorate when F is the Dirac mass in \(\frac{1}{2}\); and the IAC random electorate is the strongly exchangeable electorate when F is the uniform distribution on [0, 1].
Our argument is a straightforward adjustment of the argument used by Chamberlain and Rothschild (1981) in proving their proposition 1. Note also that for the sake of simplification, we limit ourselves to \(\theta \) which are rational numbers and sequences of electorates where the use of the integer part can be avoided.
Using the identity:
$$\begin{aligned} \left( {\begin{array}{c}n\\ k\end{array}}\right) \int _{0}^{1}p^{k}(1-p)^{n-k}dp=\left( {\begin{array}{c}n\\ k\end{array}}\right) \frac{\left( k\right) !\left( n-k\right) !}{(n+1)!}=\frac{1}{n+1}. \end{aligned}$$Indeed as already pointed out, Influence \((i,IAC,{\mathfrak {C}} )\) is the Shapley imputation of player i in the simple game attached to \({\mathfrak {C}}\). If the game is symmetric, then all the players receive the same payoff in the Shapley solution (as it is a symmetric solution) and, since the total payoff is 1 (as the game is not constant), the claim follows.
In an earlier version of this article, we computed utility by evaluating separately the influence term and the correction term, and only in the IAC case. We owe this simpler and more general argument to compute Total Utility to one of our referees.
Our conjecture is that the relation holds for larger m, but we have not proven it.
If the conjecture stated in the previous footnote holds, the Stirling formula shows that the per capita social gain over the ex ante best alternative for a large electorate behaves when m gets large as \(\sqrt{\frac{1}{4\pi m}}\).
Recall that the IC electorate is the strongly exchangeable electorate when F is the Dirac mass at \(\frac{1}{2}\).
The stronger inequality \(\sqrt{2\pi k}\left( \frac{k}{e}\right) ^{k}\exp \frac{1}{12k+1}\le k!\le \sqrt{2\pi k}\left( \frac{k}{e}\right) ^{k}\exp \frac{1}{12k}\) is due to Robbins (1955).
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Acknowledgments
We would like to express our gratitude to our three anonymous referees and to the editor in chief for a very careful and constructive criticism of earlier versions of this manuscript. We would also like to thank Federico Valenciano for calling our attention to his joint work with Laruelle in which they also investigate the difference between what they call success (we call utility) and what they call decisiveness (we call influence) which is the starting point of our research. We thank Andy Eggers and Jonathan Klingler for helpful comments.
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Appendix
Appendix
Let us consider \(n=mr+1\) and \(q=lr+1\) with \(0<l<m\). We prove that if f is continuous at \(\theta \equiv \frac{l}{m}\), then:
Let \(A=\left\{ p\in \left] \theta -\epsilon ,\theta +\epsilon \right[ \right\} \) and \(B=\left\{ p\in \left[ 0,\theta -\epsilon \right] \cup \left[ \theta +\epsilon ,1\right] \right\} \) for some \(\epsilon >0\). By Formula (12):
Let us first consider \(\left( mr+1\right) \left( {\begin{array}{c}mr\\ lr\end{array}}\right) \int _{B} p^{lr}(1-p)^{(mr-lr)}f(p)dp\). By using two times the lower bound and one time the upper bound of the following general inequalityFootnote 32:
we get that
Therefore, rearranging terms, one gets that:
The logarithm of the expression \(\left[ \left( \frac{p}{\theta }\right) ^{\theta }\left( \frac{1-p}{1-\theta }\right) ^{(1-\theta )}\right] ^{mr}\) writes:
The function \(\theta \ln \left( \frac{p}{\theta }\right) +(1-\theta )\ln \left( \frac{1-p}{1-\theta }\right) \) is strictly concave (as a function of p) on \(\left] 0,1\right[ \). It is uniquely maximized when \(p=\theta \) and takes the value 0. Then there exists \(c>0\) such that \(\theta \ln \left( \frac{p}{\theta }\right) +(1-\theta )\ln \left( \frac{1-p}{1-\theta }\right) <-c\) for all \(p\in B\) and therefore:
yielding that
Let us now consider \(\left( mr+1\right) \left( {\begin{array}{c}mr\\ lr\end{array}}\right) \int _{A} p^{lr}(1-p)^{(mr-lr)}f(p)dp\). Since f is continuous at \(\theta \), for any \(\delta >0\), we can choose \(\epsilon \) small enough that \(-\delta \le f(p)-f(\theta )\le \delta \) for all \(p\in A\). Therefore:
Adding inequalities (17) and (18), we obtain:
Let us now show that:
From
and the identity in footnote 25, we get the following inequality:
From inequality (17) applied to the case where f is constant (i.e. the uniform distribution) we obtain:
and therefore
We conclude that:
Collecting inequalities (19) and (20), and taking the limit when r goes to infinity, we deduce that:
Hence:
and therefore:
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Le Breton, M., Van der Straeten, K. Influence versus utility in the evaluation of voting rules: a new look at the Penrose formula. Public Choice 165, 103–122 (2015). https://doi.org/10.1007/s11127-015-0296-1
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DOI: https://doi.org/10.1007/s11127-015-0296-1