Abstract
We examine two explanations for peer effects in risk taking: relative payoff concerns and preferences that depend on peer choices. We vary experimentally whether individuals can condition a simple lottery choice on the lottery choice or the lottery allocation of a peer. We find that peer effects increase significantly, almost double, when peers make choices, relative to when they are allocated a lottery. In both situations, imitation is the most frequent form of peer effect. Hence, peer effects in our environment are explained by a combination of relative payoff concerns and preferences that depend on peer choices. Comparative statics analyses and structural estimation results suggest that a norm to conform to the peer may explain why peer choices matter. Our results suggest that peer choices are important in generating peer effects and hence have important implications for modeling as well as for policy.
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Notes
Peers might generally influence risk and other economic attitudes (Ahern et al. 2013). Peers also affect credit decisions (e.g., Banerjee et al. 2013; Georgarakos et al. 2012), savings decisions (e.g., Duflo and Saez 2002; Kast et al. 2012) as well as different teenager (risky) behaviors (for an overview, see Sacerdote 2011). Generally, peer effects are important in education (e.g., Sacerdote 2001; Duflo et al. 2011), in labor (e.g., Falk and Ichino 2006; Mas and Moretti 2009; Card et al. 2010), and pro-social behavior (e.g., Gächter et al. 2013).
To increase the salience of complete information in our experiment, instructions were read aloud, for both potential roles in the experiment, and roles were assigned randomly within the same session. Also, we designed the lotteries to have at most two outcomes to minimize complexity. For a given probability distribution over the good and bad outcome, there were always six pairs of choices, which featured the exact same risky lottery. In half of the situations the safe lottery had two outcomes, and only one in the other half. The number of outcomes of the safe lottery, which can be viewed as a measure of complexity, does not have a significant influence on peer effects.
There are a variety of studies examining social comparison effects in games such as public good games or coordination games (e.g., Falk and Fischbacher 2002; Falk et al. 2013). In social learning environments, Çelen and Kariv (2004) also study herding behavior, and identify substantial herding behavior.
In terms of risk preferences B cannot be labeled as safe since it does not necessarily yield a certain payoff. In comparison to A, we still label it as safe, for simplicity, as its variance is always smaller. But note that a risk averse individual does not necessarily prefer B over A.
Groups remain the same for the whole of Part II. All choices are made without any feedback until the end of the experiment. During Part I participants only know there will be a Part II in the experiment, but do not know anything about the decisions they will be asked to make. At the end of the experiment, individuals are informed about their payoff and, if Part II is drawn for payment, the choice and payoff of the other individual in the group. Throughout, we will refer to the peer as “she” and the decision maker as “he”.
An alternative definition of peer effect is to consider only imitation and deviation (conditional strategies) as peer effects, since revisions could be due to mistakes. In Section 3.2 we examine both types of definitions and find qualitatively similar results.
A similar control treatment was used by Cason and Mui (1998) to study social influence in dictator games. Also, we note that in the Coin treatment we can still examine four potential strategies of second movers in this treatment. However, in Coin the definition of imitation and deviation is arbitrary, as there is no direct link between the lottery choice of the decision maker and that of the peer.
More specifically, we conducted a Base treatment, in which choices were made twice, in Part I and Part II, without the strategy method and without social feedback. We also conducted an Anticipation treatment, without the strategy method, but where individuals were aware they would be given feedback about the peer’s choice at the end of the experiment. Consistent with the effects of our main treatments, we observe peer effects increase significantly with anticipated social feedback, from occurring in 6.7% of the decisions in Base to 17.5% in the Anticipation treatment (Mann-Whitney test, p-value=0.016).
Instructions for all treatments are available in the Electronic Supplementary Materials on the journal’s website. The raw data as well as the z-tree codes are included in the Electronic Supplementary Material as well.
To ensure credibility, one participant was randomly selected as assistant at the end of the experiment. The assistant drew one ball from an opaque bag containing balls corresponding to each part and from a second bag with balls corresponding to each decision problem. For each decision problem, the respective combination of black and white balls was put in an opaque bag and the assistant again drew one ball. Once all draws were done, payoffs were computed and subjects were paid out in cash.
This literature started with Veblen (1899) and Duesenberry (1949), who argued that conspicuous consumption choices can be explained by a desire to signal a superior status, prowess or strength. A game-theoretic literature has focused on the implications of status concerns on conspicuous consumption (see, e.g., Hopkins and Kornienko 2004) and conformity (see, e.g., Bernheim 1994). Here we focus on ex-post payoff differences between the decision maker and his peer, and measure strategy choices of decision makers, who make conditional choices for each of the two possible lotteries of the peer. Related studies on social preferences under risk (e.g., Trautmann 2009; Saito 2013) point out that individuals may exhibit ex-ante relative payoff concerns, i.e. dislike inequality in expected payoffs. In our setting, such concerns yield qualitatively the same predictions, since risks are perfectly correlated. By choosing the lottery of the peer, decision makers can equalize expected payoffs both in Random and Choice.
In the context of risk taking in the presence of others, whether individuals exhibit a desire to be ahead or not may depend on the situation (see Maccheroni et al. 2012, for a discussion). In our context, in which payoff differences are relatively small and the situation allows for a simple comparison with the peer, we would rather expect individuals dislike falling behind others, but enjoy being ahead. In Appendix A.1 we propose such a model in which decision makers are loss averse with respect to the peer’s outcome, and derive conditions under which peer effects are expected to occur. Note that assuming a dislike to being ahead of the peer would even strengthen the incentive to imitate the peer.
One may also consider intention-based social preferences as an alternative explanation for why choices matter for relative payoff concerns (see, e.g., Blount 1995; Bolton et al. 2005; Falk and Fischbacher 2006). In our experiment there is no scope for reciprocity and it hence is unlikely that intention-based preferences are a driver of behavior. However, if intention-based social preferences would play a role, we would predict these to increase the weight on (negatively valued) payoff differences in the decision maker’s utility when moving from Random to Choice, which in expectation crucially depends on how A and B relate in terms of their expected values. Hence, moving from Random to Choice, not only would this theory predict imitation increases, but also that the increase in imitation depends on f. We do not find evidence for this in our data.
According to Festinger, “an opinion, a belief, an attitude is ‘correct’, ‘valid’, and ‘proper’ to the extent that it is anchored in a group of people with similar beliefs, opinions and attitudes”; Festinger (1950), p. 272.
See Appendix A.2 for a straightforward model based on social comparison theory.
A detailed overview of choices in Part I is provided in Table 1 in Online Appendix C.1
We also controlled for consistency of decisions in Part I. If we assume that subjects have CRRA preferences and given the design of our lotteries, we can classify second movers as consistent or inconsistent decision makers. We find across different probability panels, controlling for certainty, that at most 15.4% of decision patterns are inconsistent. If we exclude inconsistent decision makers from our sample our results remain qualitatively the same.
At the individual level, the distribution of switching rates also differs across treatments. It is significantly different in Choice, compared to Random and Coin (Kolmogorov-Smirnov test, p-value=0.02 compared to Coin, p-value=0.09 compared to Random). But it does not differ significantly across Random and Coin (Kolmogorov-Smirnov test, p-value=0.96). Figure 1 in Online Appendix C.1 displays the distribution of switching rates by treatment.
Table 2 in Online Appendix C.1 displays the frequency of each strategy choice for each decision, by treatment.
In Random, the switching rate is close to 50%, since lotteries are randomly assigned to the peer with probability 0.5.
We thank a referee for this suggestion.
One might also argue that an increase in imitation from treatment Random to Choice might depend on the expected value of A relative to B. Intuitively, if agents exhibit relative payoff concerns and lottery B yields a higher expected payoff ( f < 1), the marginal increase in utility from imitation is stronger in magnitude in case the decision maker chooses B. This implies a stronger incentive to imitate B if f < 1 compared to f ≥ 1. We ran additional regressions, similar to the one presented in Table 4, distinguishing between lottery panels with f < 1, f > 1 and f = 1, and find that the interaction between Choice and imitation of B is only significant, and negative, when f = 1. That is, we do not find any systematic relationship between expected values and the increase in imitation. Details can be obtained from the authors upon request.
Another approach could be to simultaneously estimate parameters defining relative payoff concerns and an additional utility from conforming to the social anchor. However, imitation (or deviation) can very generally be explained by a positive (or negative) estimate of γ as well as by λ > 1 (or λ < 1). Identifying both parameters, for both treatments simultaneously, is not possible with our data, but would be an interesting task for future work. Another approach might be to estimate mixture models, a procedure that we applied in a previous version of this paper. Mixture models have been used to estimate risk preferences in heterogeneous populations, amongst others by Conte et al. (2011) and Harrison and Rutström (2009). However, in our setting, assuming heterogeneity with respect to whether decision makers derive a social utility or not causes the following concern. The probability to be of a certain type enters into the log-likelihood function as a multiplicative weight of the social utility, and in this way scales the estimates of λ and γ. Moreover, it leaves one additional degree of freedom.
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Acknowledgments
We would like to thank Kenneth Ahern, Jim Andreoni, Daniel Clarke, Dirk Engelmann, Florian Englmaier, Fabian Herweg, Alex Imas, Martin Kocher, Johannes Maier, Jan Potters, Justin Sydnor, Stefan Trautmann, Lise Vesterlund and Joachim Winter for their useful comments as well as seminar participants at Royal Holloway, University of Bamberg, University of Gothenburg, University of Innsbruck, University of Munich, University of Pittsburgh, at the 2014 AEA Meetings, 2012 CESifo Area Conference on Behavioural Economics, 2012 European and North-American ESA Meetings, 7th Nordic Conference in Behavioral and Experimental Economics in Bergen, ESI Workshop on Experimental Economics II, the CEAR/MRIC Behavioral Insurance Conference 2012, the Risk Preferences and Decisions under Uncertainty SFB 649 Workshop, the 2013 theem meeting in Kreutzlingen, and the 2013 Workshop on Behavioral and Experimental Economics in Florence. We gratefully acknowledge funding from the Fritz Thyssen Foundation (Project AZ.10.12.2.097).
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Appendix A: Theoretical framework
Appendix A: Theoretical framework
1.1 A.1 A model of relative payoff concerns
Assume the utility in state j ( j ∈ {g, b}) of having chosen lottery i ( i ∈ {A, B}) and earning \({m^{j}_{i}}\), to be given by the sum of two terms: a consumption utility, which is solely determined by individual risk preferences, plus a social utility term, which depends on payoff differences. This implies \(v^{j}_{i,k}=u\left ({m^{j}_{i}}\right ) + R\left ({m^{j}_{i}}-{m^{j}_{k}}\right ),\) where k ∈ {A, B} is the lottery of the peer, and R(⋅) is a function of payoff differences and defined as follows:
The parameter λ captures how large losses with respect to the peer loom relative to gains. An individual’s expected utility from choosing lottery i is
where U i is the expected consumption utility of lottery i. If the peer holds a lottery that yields a lower consumption utility, the individual may nevertheless choose it, if he experiences a strong disutility from falling behind the peer, i.e. if λ is large enough. Let us define an individual’s strategy space as \(\mathcal {S}=\{\text {imitate} = (i; AA,BB),\, \text {deviate} = (i; BA,AB),\, \text {stay} = (i; iA,iB), \,\text {change} = (i; -iA,-iB); \,\text {for } i\in \{A,B\}\}\). Here i ( −i) denotes his (opposite) choice in Part I, and the tuple ik describes the choice of lottery i in Part II given that his peer has lottery k. Then, the cutoffs are given by the following proposition.
Proposition 1
Define \({\Delta }\equiv \frac {U_{B}-U_{A}}{p\delta }+\frac {(1-p)(c-\delta )}{p\delta }\) and \({\Theta }\equiv \frac {U_{A}-U_{B}}{(1-p)(c-\delta )}+\frac {p\delta }{(1-p)(c-\delta )}\) . An individual imitates if λ > max{Δ, Θ}. An individual deviates if λ< min{Θ, Δ}. An individual stays with his Part I choice otherwise.
Note that whether Δ is smaller or greater than Θ is determined by the individual’s choice in Part I, i.e. by his expected consumption utility U A and U B .
Proof
An individual imitates if V A, A > V B, A and V B, B > V A, B . V B, B > V A, B is equivalent to
V A, A > V B, A is equivalent to
Hence, for an individual to imitate it must hold that λ > max{Δ, Θ}.
Similarly, an individual deviates if V A, A < V B, A and V B, B < V A, B . It follows directly from above that this is satisfied if λ < min{Δ, Θ}. □
1.2 A.2 A model based on social comparison theory
Consider a model in which the closer the individual risky choice is to the social anchor, the more utility the individual derives. In a setting with only two options, this can be captured by an additional utility γ when the option chosen coincides with the social anchor. In particular, the expected utility of lottery i given the anchor k is
where 1(⋅) is the indicator function. (Cooper and Rege, 2011, also assume this form of utility when examining conformity.) Based on the argument above, we would expect γ to differ across treatments and γ C , in Choice, to be larger than γ R , in Random. This would generate an increase in imitation in Choice. Further, since the effect of γ is independent of lottery characteristics, we would expect the change in imitation across treatments to be symmetric with respect to the two available options, A or B.
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Lahno, A.M., Serra-Garcia, M. Peer effects in risk taking: Envy or conformity?. J Risk Uncertain 50, 73–95 (2015). https://doi.org/10.1007/s11166-015-9209-4
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DOI: https://doi.org/10.1007/s11166-015-9209-4