Political hierarchies and political shirking

Journal of Economic Behavior & Organization Vol. 65 (2008) 334–356

Political hierarchies and political shirking Daniel Sutter a,∗ , Marc Poitras b a b

Department of Economics and Finance, University of Texas—Pan American, Edinburg, TX 78541-2999, United States Department of Economics, University of Dayton, Dayton, OH 45469-2251, United States

Received 5 March 2002; received in revised form 20 December 2002; accepted 22 February 2003 Available online 27 September 2006

Abstract We analyze how citizens can screen candidates for high political office in a political hierarchy. In our model politicians differ in willingness to misuse discretionary powers of office. Those politicians willing to abuse the powers of office (knaves) can be induced in a hierarchy to reveal their type while in low office, thus disqualifying themselves for advancement. Hierarchy’s capacity for screening knaves out of high office depends on the conditions for reappointment to low office. A seemingly perverse rule that reappoints shirking politicians can perform well. To demonstrate an empirical application, we analyze shirking in the House Bank check bouncing scandal. © 2006 Published by Elsevier B.V. JEL classification: D72; H1 Keywords: Political shirking; Screening; Hierarchy

1. Introduction One of the foremost tasks of representative government is to prevent misuse of political power. According to the modern literature on control of politicians, political shirking can be mitigated by repeated elections in two ways. First, the prospect of reelection can motivate politicians (agents) to behave as voters (principals) desire: an incentive effect. Second, politicians who better represent voters win reelection with greater frequency: a sorting effect.1 In some official positions, however, ∗

Corresponding author. Tel.: +1 956 381 3391. E-mail address: [email protected] (D. Sutter). 1 Incentive and sorting effects address moral hazard and adverse selection problems, respectively. See Barro (1973), Ferejohn (1986), Austen-Smith and Banks (1989), Lott and Reed (1989), Reed (1994), and Banks and Sundaram (1997). 0167-2681/$ – see front matter © 2006 Published by Elsevier B.V. doi:10.1016/j.jebo.2003.02.005

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Nomenclature κ L Sh S Th T Vh V

proportion of knaves in pool number of low offices payoff from shirking in high office payoff from shirking in low office cost to citizenry of shirking in high office cost to citizenry of shirking in low office payoff from good conduct in high office payoff from good conduct in low office

Greek symbols β payoff from good conduct relative to shirking, V /S πh probability high officeholder provides public goods probability low officeholder provides public goods π

repeated elections might not insure sufficiently good behavior, for two reasons. First, the process of periodic reelection or reappointment can undermine the public interest by injecting the interests of political factions. For this reason, scholars maintain that some public institutions such as the Supreme Court and the Federal Reserve must operate independently of the legislature, executive, and even voters. Second, the incentive and sorting effects of reelection require politicians to prove themselves on the job, but misuse of power in some positions can impose intolerable costs, so citizens might not want to allow politicians to prove themselves on the job. For example, an unprincipled national leader might plunge his country into a bloody war before citizens can inflict electoral punishment, or thwart electoral removal from office by subverting democracy and the rule of law. Politicians from Adolf Hitler to Ferdinand Marcos have used democratic institutions to obtain high office and then destroyed those institutions. For some political offices, citizens will therefore want to pre-screen candidates. The authors of The Federalist Papers clearly saw screening candidates for high office as a role for political institutions: This process of election affords a moral certainty that the office of President will seldom fall to the lot of any man who is not in an eminent degree endowed with the requisite qualifications. Talents for low intrigue, and the little arts of popularity, may alone suffice to elevate a man to the first honors in a single state, but it will require other talents, and a different kind of merit . . . to make him a successful candidate for the distinguished office of President of the United States. (Hamilton, Federalist #68) Citizens can pre-screen candidates for high office by observing conduct in prior offices. Hence we model a political hierarchy in which candidates must demonstrate satisfactory performance in low office in order to qualify for high office. We define two types of politician: angels, who always use the power of government to serve the citizenry, and knaves, who misuse power to benefit themselves.2 A politician’s type is private information. A knave serving in high office inevitably 2 For example, the great philosopher David Hume wrote that “in contriving any system of government, and fixing the several checks and controls of the constitution, every man ought to be supposed a knave, and to have no other end, in

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shirks, so voters wish to identify knaves and prevent them from taking high office. Shirking in low office unmistakably identifies a knave. But if low office shirking disqualifies knaves from advancing to high office, knaves may conceal their type by refraining from shirking until they reach high office. Hence, barring identified knaves from high office may not be incentive-compatible with revelation of knaves’s type. Knavish disguise can have disastrous consequences, as illustrated by the Nazi Party’s rise to power in Germany. The failure of the Munich putsch convinced Hitler that he could not pursue power outside of legal channels. After the Nazis secured power legally, Joseph Goebbels attested to the role played by deception: “Our domestic foes never saw where we were going or that our oath of legality was just a trick. We intended to come to power legally, but we did not intend to use power legally (Johnson, 1991)”. Politicians are strongly motivated by ambition, and political economy has long recognized the need to design institutions to mitigate the undesirable effects of political ambition. As James Madison put it, “Ambition must be made to counteract ambition” (Federalist #51).3 In models of retrospective voting and electoral control, ambition leads politicians to serve the interests of voters. But in our case ambition for high office can harm the public interest by driving knaves to disguise their type. In order to keep knaves out of high office, institutions must negate their ambition. In our model of political hierarchy, knaves choose to misbehave and reveal themselves, despite the disqualification for high office, if they can reap sufficient reward from shirking in low office. The gain to knaves from shirking depends on the payoffs associated with shirking and good behavior in high and low offices, as well as on the rule for reappointment to low office. In particular, knaves have more incentive to reveal themselves, and the hierarchy’s screening performance can improve, if the rule grants reappointment to shirkers. Paradoxically, the political system can benefit when corrupt politicians win reelection. Furthermore, the political hierarchy must give angels sufficient incentive to give up low office and run for high office. Angels can be forced “up-or-out” by a reappointment rule that limits the number of terms in low office. Term Limits also have the advantage of not discriminating against knaves who shirk, since the Term Limit precludes reappointment of both shirkers and non-shirkers. By compelling angels to run for high office, a Term Limit increases competition for high office, discouraging knaves from pursuing high office. Thus our model provides a unique perspective on the effects of Term Limits and other policies that influence tenure in political office. We also find that a perverse rule that reappoints shirkers but not non-shirkers can perform well; such a rule has the dual advantage of rewarding knaves who reveal themselves and forcing angels to run for high office. But no reappointment rule performs best unconditionally; relative performances depend on the payoffs from shirking and good behavior in high and low offices. The paper proceeds as follows. Section 2 presents the model, and Section 3 calculates the equilibrium screening performance of hierarchy with four different reappointment rules for low office. Section 4 compares the effects of the reappointment rules and considers implications for the organization of political systems. Section 5 demonstrates an empirical application of the model by examining the Congressional “check bouncing” affair. Our empirical results suggest that

all his actions, than private interest” (1964, p. 117–8). Papers employing similar types of politicians include Coate and Morris (1995), Brennan and Hamlin (1995), Le Grand (1997) and Sutter (1998). A number of papers (Harrington, 1993; Cowen et al., 2000; Sutter, 1998) incorporate screening as creating a (perhaps noisy) signal about a politician’s type. Frey (1997) considers the case where all politicians may behave knavishly and examines the incentive effects of political controls. 3 Modern political science treatments of ambition include Schlesinger (1966) and Black (1972).

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characteristics of state-specific political hierarchies can significantly influence politician behavior. Finally, Section 6 discusses the implications of our model for the design of political institutions. 2. The model Experience suggests that people differ in willingness to pursue self-interest at the expense of others. Our model therefore assumes two types of politician: angels and knaves. Angels never shirk or misuse the powers of public office. Knaves behave opportunistically, shirking when this raises their expected utility. We assume that post-political payoffs (Lott, 1990), such as a pension or criminal prosecution, are insufficient to deter knaves from shirking. A payoff from shirking (bribe, campaign contribution, etc.) that raises a knave’s utility enters negatively in an angel’s utility function, perhaps due to guilt. A politician has a two-period time horizon, and the political system is infinitely lived. We refer to a politician in his first (second) period of political life as young (old). A politician makes a discrete choice to shirk or behave conscientiously (provide public goods, for example). All citizens prefer that politicians not shirk, that is, we employ a representative citizen (common interest) framework and ignore issues of distribution among citizens. The common interest assumption is routinely employed by the literature on electoral control.4 We assume that citizens perfectly observe a politician’s behavior in office. Since shirking unmistakably identifies a knave, knaves can pursue high office only by disguising their type. Standard models of control of politicians consider repeated elections to the same office. In contrast, we assume two types of office: a total of L low offices and a single high office. The two types of office differ in the payoffs they offer to occupants and the services they provide to citizens. Each citizen resides in one low office district, and districts contain equal numbers of residents. 2.1. Politicians There exists an exogenous pool of political office candidates. Let κ be the proportion of knaves in the pool; κ does not vary as candidates are selected to fill the available posts. Random draws of young politicians from the pool fill low office vacancies. Politicians must serve satisfactorily in low office to qualify for high office.5 After serving in low office when young, eligible politicians decide whether to remain in low office or to run for high office. Politicians who choose to run for high office must give up low office. The promotion rule states that if more than one low office holder with a good record tries for high office, each has an equal probability of winning; that is, voters draw no distinctions among candidates with good first period records. Unsuccessful candidates for high office and politicians denied reappointment to low office, hold no public office when old.

4

Ferejohn (1986) and Dixit et al. (1997) explore the consequences of common agency for controls on politicians. Although in the United States prior public service is not a formal requirement for any office, a de facto political hierarchy nonetheless exists. Thirty-nine Senators in the 104th Congress had previously served in the House of Representatives, and 31 others had served as governors or state legislators. Among governors in 1995, 27 had prior service in Congress or state legislature, while nine more had served as lieutenant governor, attorney general, or mayor of a large city (Barone and Ujifusa, 1996). Running for higher office accounted for 42.9 percent of voluntary departures from the House between 1960 and 1990 (Schansberg, 1994). 5

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Let either type of politician receive single-period payoffs V and Vh for good conduct in low and high office. These payoffs include the official salary and any nonpecuniary benefits of holding office (ego rents). Knaves receive single-period payoffs of S and Sh , net of any ex post penalties, from shirking in low and high office. In a given period, knaves benefit most by shirking: Sh > Vh , S > V . A given behavioral choice yields a relatively greater payoff in high office, Vh /V > 1 and Sh /S > 1. Politicians out of office when old receive a payoff normalized to 0. For simplicity, we assume politicians do not discount second period payoffs.6 The payoffs are exogenous, and their impact on the effectiveness of screening within the hierarchy is discussed in Section 4. Politicians are risk neutral, and their advancement decisions and first period behaviors will depend on the probability of reappointment contingent on first period performance, σ(a), as defined below. 2.2. Citizens Citizen welfare depends on politician performance, which in turn depends on the reappointment rule for low office. We use the term reappointment rather than reelection because many political hierarchies appoint officials (Federal judges, colonial governors). The reappointment rule might be incorporated in the political constitution. Alternatively, under an elective hierarchy, the reappointment rule might represent a choice by citizens to commit to a voting strategy. Let ai signify low office occupant i’s action when young; ai ∈ {0, 1}, where 1 = behave conscientiously and 0 = shirk. For an official seeking a second term in low office, let σ(a) be the probability of reappointment as a function of first period job performance. Four deterministic rules exist. Retrospective Reappointment: Reappoint incumbents who have not shirked, but no incumbents who have shirked: σ(1) = 1, σ(0) = 0. Life Tenure: Reappoint all incumbents regardless of first period performance: σ(1) = σ(0) = 1. Life Tenure can also approximate circumstances where incumbents have a particularly strong advantage over challengers. Term Limits: Reappoint no incumbent: σ(1) = σ(0) = 0. Term Limits amount to an up-or-out rule for low office. Perverse Reappointment: Reappoint incumbents who have shirked, but no incumbents who have not shirked: σ(1) = 0, σ(0) = 1. Citizens select a reappointment rule based on each rule’s equilibrium screening performance. Let πh and π denote equilibrium probabilities of good performance in high and low office. Here π measures the rate of good performance across the several low offices. When the duties of high and low office are faithfully performed, the representative citizen receives a payoff normalized to 0. Let T and Th be the welfare losses from shirking in low and high office: T , Th < 0. Expected citizen welfare is given by W = (1 − πh )Th + (1 − π )T .

(1)

The values of Th and T depend on how the constitution allocates tasks to the two levels of government. We treat Th and T as exogenous, but discuss optimal allocation of tasks in Section 4. How will citizens benefit from employing hierarchy? Suppose the hierarchy raises πh by πh

6

Allowing politicians to discount future payments does not substantively alter any of our results.

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while lowering π by an absolute amount π . The change in welfare is W = −πh Th + π T , so the hierarchy raises welfare if Th π > . T πh

(2)

Reversing the inequality would imply that citizens prefer to prevent shirking in low office. The most important powers of state, however, usually reside in high political offices, despite prominent figurehead positions in some countries. Consequently, we assume a ratio Th /T sufficiently large to satisfy (2), which implies that hierarchy’s goal is to keep knaves out of high office. Citizens thus prefer whichever rule yields the largest value of πh ; that is, citizens have lexicographic preferences over relevant pairs of πh , π . In our model, citizens’ evaluation of various reappointment rules is complicated by multiple equilibria. Brennan and Buchanan (1983) argue that the appropriate criterion for evaluating constitutional rules is worst-case performance. We therefore restrict attention to the equilibria where knaves advance and angels remain in low office, or the equilibria with the lowest value of πh . 3. The effectiveness of screening in a hierarchy A reappointment rule σ and the promotion rule together define a Bayesian game with L players.7 Solution of the game is complicated by the large number of possible combinations of politician types in the L low offices. We focus here on the case where L = 2, which permits an explicit solution, albeit with a large number of cases. Section 4 allows L > 2. We solve the game backward in three stages. First, we easily determine politicians’ actions when old: angels perform dutifully and knaves shirk in whichever office they hold. Next, we examine the decision by eligible politicians to run for high office after completing the first term in low office. Finally, we consider knaves’ shirking decision when young. The decision to seek high office depends on the reappointment rule for low office. We begin with Term Limits. This up-or-out rule compels all qualified young politicians to try for high office. Knaves receive expected payoffs of S from shirking when young and V + Sh /2 from behaving themselves and trying for high office. Knaves thus wait until old to shirk if β≥

1 − Sh , 2S

(3)

where β denotes the payoff from good conduct relative to shirking: β = V /S < 1. If (3) holds, knaves behave and the probability of good conduct in high office equals the proportion of angels in the candidate population: 1 − κ. If we condition on the three possible cohorts of young officeholders (two knaves, one knave and one angel, two angels), we obtain conditional probabilities of good conduct in high office of 0, 1/2, and 1. We summarize the conditional probabilities for Term Limits in Table 1. If β < 1 − Sh /(2S ), knaves shirk and disqualify themselves for high office. Hierarchy thus induces a separating equilibrium under Term Limits if low office shirking provides sufficient reward, that is, if β is sufficiently small. In the separating equilibrium, an angel holds high office with probability one, not 1/2, if the cohort of young officeholders contains an angel and 7 In an appointive hierarchy, the reappointment rule is a choice variable, but in an elective hierarchy the rule must satisfy a rationality constraint for the relevant voting population.

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Table 1 Determination of equilibrium with Term Limits Cohort

πh

π

Pooling equilibria (β ≥ 1 − Sh /2S ) Two knaves One angel, one knave Two angels

0 1/2 1

1 1 1

Separating equilibria (β < 1 − Sh /2S ) Two knaves One angel, one knave Two angels

1−κ 1 1

0 1/2 1

a knave since the knave disqualifies himself. If the young cohort contains two knaves, they both shirk, and high office in period 2 must be filled by a draw from the pool; the probability an angel holds high office is now 1 − κ (instead of 0). Using the cohort probabilities of κ2 , 2κ(1 − κ) and (1 − κ)2 gives an overall probability of an angel holding high office equal to κ2 (1 − κ) + 2κ(1 − κ) + (1 − κ)2 = (1 − κ)(1 + κ + κ2 ). We can evaluate the screening effectiveness of hierarchy by comparing hierarchy’s performance to a non-hierarchical system that dispenses with the low office prerequisite for high office, thus making high and low offices independent. Under independent offices, politicians selected directly from the pool fill high office and serve one term. This yields πh = 1 − κ, which we take as our benchmark. Comparison with the above expression reveals that if hierarchy with Term Limits induces a separating equilibrium, high office performance improves relative to the independent offices benchmark. With Retrospective Reappointment, the advancement decision is not trivial. Since high office is relatively more rewarding (Sh > S and Vh > V ), both knaves and angels try for high office if eligible and unopposed. When opposed, a knave compares Sh /2 with S and an angel Vh /2 with V . Consequently, the cases in the advancement stage turn on whether payoff ratios Vh /V and Sh /S are greater or less than 2. If Vh /V , Sh /S ≥ 2, both angels and knaves pursue high office even if opposed. If Vh /V , Sh /S < 2, only one politician, angel or knave, seeks high office, and this implies two equilibria if low office is occupied by one angel and one knave. Table 2 displays the advancement decisions of each cohort of young officeholders. The final column indicates the number of low offices open to young politicians from the pool in period 2. The Term Limits rule forces low office holders up-or-out, but if Sh /S or Vh /V < 2, Retrospective Reappointment does not open all low offices for the next cohort.8 Since politicians in our model do not discount future payoffs, Retrospective Reappointment fails to produce a separating equilibrium in any case. Shirking yields knaves a payoff of only S , while behaving and retaining low office pays V + S . We have then a situation where Retrospective Reappointment works well as an incentive device for a single office, yet performs poorly as a screening device. Also, when Sh /S or Vh /V < 2, Retrospective Reappointment leaves one low office occupied in period 2. Consequently, there begins in period 2 a low office cohort of one young

8 Table 2 also presents the expected payoffs to old knaves as a function of cohort and the four relevant inequalities among Vh /V , Sh /S , and 2. When two knaves hold office and Sh /S < 2, ex ante we assume each knave has an equal chance of moving up, so the payoff when old is (Sh /S )/2.

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Table 2 Advancement stage outcomes Condition on payoffs

Cohort

Decision on advancement Angels

Knaves

πh

Knaves’ expected payoff when old

No. of low offices open in period 2

Vh /V , Sh /S ≥ 2

Two knaves One angel, one knave Two angels

– Yes Yes, yes

Yes, yes Yes –

0 1/2 1

Sh /2 Sh /2 –

2 2 2

Vh /V ≥ 2 > Sh /S

Two knaves One angel, one knave Two angels

– Yes Yes, yes

Yes, no No –

0 1 1

(Sh + S )/2 S –

1 1 2

Sh /S ≥ 2 > Vh /V

Two knaves One angel, one knave Two angels

– No Yes, no

Yes, yes Yes –

0 0 1

Sh /2 Sh –

2 1 1

Vh /V , Sh /S < 2

Two knaves One angel, one knave Two angels

– No Yes, no

Yes, no Yes –

0 0 1

(Sh + S )/2 Sh –

1 1 1

politician. A knave in a cohort of one definitely behaves and takes high office since V + Sh > S .9 Since the lone officeholder from the second generation inevitably advances, the second cycle πh equals 1 − κ, the proportion of angels in the pool. The second cycle π equals 1 or 1/2 depending on whether the holdover politician is angel or knave. Both low offices open up for young politicians in period 3, so we need contemplate only a two-period cycle. Table 3 presents the two-period averages of πh and π for Retrospective Reappointment as well as the other three rules. Finally, we derive the conditions for separating equilibria under Perverse Reappointment and Life Tenure. A young knave in low office assigns probability κ to the prospect that the other occupant of low office is a knave. The calculations are tedious; Table 3 presents the resulting cutoff values for β necessary for a separating equilibrium. As in the case of Retrospective Reappointment, the outcomes for Life Tenure involve the four advancement stage cases defined by Vh /V and Sh /S . Unlike Retrospective Reappointment, Life Tenure and Perverse Reappointment can induce knaves to shirk by permitting a second period in low office, thus yielding a shirking knave an overall payoff of 2S . Otherwise the analysis of Life Tenure mirrors that of Retrospective Reappointment. Under the Perverse Reappointment rule, knaves behave and deny themselves reappointment to low office only if they intend to try for high office; knaves shirk if they prefer two periods of low office shirking. 4. Comparison of hierarchies Table 3 reveals a tradeoff between πh and π . Hierarchy cannot improve on the benchmark without also reducing π . Citizens can use hierarchy to improve either high or low office performance, but not both. Citizens can thus choose at the constitutional level to assign πh = 1 − κ

9 The starkness of the knave’s position in this instance, a certainty of reaching high office if desired, results from L = 2. Realistically, an entering cohort will have more than one politician. Nonetheless, a small cohort increases the probability of reaching high office, creating a margin on which knaves behave.

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Table 3 Equilibrium outcomes with different reappointment rules Condition on Vh /V , Sh /S

Condition on β

πh

π

Pooling equilibria Life Tenure Vh /V , Sh /S ≥ 2 Vh /V ≥ 2 > Sh /S Sh /S ≥ 2 > Vh /V Sh /S , Vh /V < 2

β ≥ 2 − Sh /2S β ≥ 1 − κ(Sh /S − 1)/2 β ≥ 2 − (Sh /S )(1 − κ/2) β ≥ 1 + (1 − κ/2)(1 − Sh /S )

1−κ (1 − κ)(1 + κ − κ2 /2) (1 − κ)2 (1 + κ/2) (1 − κ)(1 − κ/2)

1 1 − κ/2 − κ2 /4 1 1 − κ2 /4

1−κ (1 − κ)(1 + κ − κ2 /2) (1 − κ)2 (1 + κ/2) (1 − κ)(1 − κ/2)

1 1 − κ/2 − κ2 /4 1 1 − κ2 /4

β ≥ 1 − Sh /2S β ≥ 2 − Sh /2S

1−κ 1−κ

1 1

β < 2 − Sh /2S β < 1 − ␬(Sh /S − 1)/2 β < 2 − (Sh /S )(1 − κ/2) β < 1 + (1 − κ/2)(1 − Sh /S )

(1 − κ)(1 + κ) (1 − κ)(1 + κ) (1 − κ)(1 + κ) (1 − κ)(1 + κ/2 + κ2 /2)

1−κ 1−κ 1−κ 1−κ

β < 1 − Sh /2S β < 2 − Sh /2S

(1 − κ)(1 + κ + κ2 ) (1 − κ)(1 + κ)

1−␬ 1−␬

Retrospective Reappointment Vh /V , Sh /S ≥ 2 Vh /V ≥ 2 > Sh /S Sh /S ≥ 2 > Vh /V Sh /S , Vh /V < 2 Term Limits Perverse Reappointment Separating equilibria Life Tenure Vh /V , Sh /S ≥ 2 Vh /V ≥ 2 > Sh /S Sh /S ≥ 2 > Vh /V Sh /S , Vh /V < 2 Term Limits Perverse Reappointment

tasks and responsibilities to high office, which implies setting Th high relative to T . Otherwise, vesting powers of the state that are most subject to misuse in entry positions defeats hierarchy as a screening device. Allocating more tasks to high office, however, probably tends to attract knaves by increasing Sh /S , thus making a separating equilibrium less likely and lowering screening performance. For individual tasks, citizens can evaluate the ratio of private benefits from abuse to social cost of abuse, S /T . Allocating tasks with high S /T ratios to low office provides knaves an incentive to remain in low office while limiting the social cost of their shirking. Figs. 1 and 2 depict for various parameters the rule that produces the highest equilibrium value of πh . The figures apply to separate cases defined by Vh /V , the ratio of high to low office payoffs from good behavior. Fig. 1 illustrates the case where high office is attractive, Vh /V ≥ 2, and Fig. 2 portrays relatively attractive low office, Vh /V < 2. The figures graph Sh /S , the ratio of high to low office payoffs from shirking, on the horizontal axis, and β, the low office payoff from good behavior relative to shirking, on the vertical axis. Inspection of Figs. 1 and 2 and Table 3 yields several results: • If Term Limits can induce a separating equilibrium, it weakly dominates the other three rules in high office performance; otherwise, Perverse Reappointment weakly dominates in high office performance. • Perverse Reappointment dominates Life Tenure in high office performance. • In all but one subcase, Term Limits weakly dominates Retrospective Reappointment in high office performance. Term Limits can in some cases improve on Retrospective Reappointment’s performance in one office, high or low, while maintaining or improving performance in the

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Fig. 1. Top performing reappointment rules with Vh /V ≥ 2.

other office, but Retrospective Reappointment cannot improve on Term Limits in one office without worsening performance in the other. • The peak high office performance, πh = (1 − κ)(1 + κ + κ2 ), occurs under Term Limits. Performance under Life Tenure and Perverse Reappointment never exceeds πh = (1 − κ)(1 + κ). No rule unconditionally performs best, but Term Limits and Perverse Reappointment perform remarkably well. These two rules prohibit reappointment of angels to low office, compelling them to run for high office. In contrast, Retrospective Reappointment and Life Tenure fail to induce angels to pursue high office if the reward is too small, Vh /V < 2. By declining to run, an angel concedes to a knave colleague an uncontested opportunity to secure high office.10 Angels who remain in low office also reduce the size of the subsequent cohort, creating opportunities for future knaves. If angels remain in low office, Retrospective Reappointment and Life Tenure can perform even worse than the independent offices benchmark. Retrospective Reappointment overall performs poorly, failing to induce a separating equilibrium. In only one case in Figs. 1 and 2 does Retrospective Reappointment provide the best high office performance, for Vh /V ≥ 2 when all rules fail to improve on the benchmark. Thus we see a tension between incentives for performance in repeated elections for a single office and screening in a hierarchy. This point is further

10 A particularly angelic politician could promote citizen welfare by selflessly running for high office just to keep the office out of knavish hands. A selfless angel would have to keep in mind, however, that a knave might take the vacated low office.

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Fig. 2. Top performing reappointment rules with Vh /V < 2.

underscored by Perverse Reappointment, which is nonsensical for a single office but performs well in a screening function. Angels can be lured to run for high office by increasing the official salary for high office so that Vh /V ≥ 2, and in this case Retrospective Reappointment and Life Tenure perform no worse than the benchmark. Increasing Vh /V and prohibiting reappointment to low office are thus substitute means of overcoming the desire of angels to remain in safe low office positions.11 But increasing the salary for high office may also increase Sh , the payoff to knaves in high office. Raising Sh encourages knaves to pursue high office, which can offset the effect on angels. In this case, it can be preferable to eject angels from low office with Term Limits or Perverse Reappointment.12 11 Another possible solution involves staggering elections, such as for the U.S. Senate, which allows angels to run for high office without forfeiting low office. In our model, all angels and eligible knaves try for high office if they retain the option of remaining in low office. In this case, equivalent to Vh /V , Sh /S ≥ 2, the pooling equilibria yield πh = 1 − κ, so hierarchy does not improve upon the benchmark. In the separating equilibria, however, allowing unsuccessful candidates to keep low office can improve performance by attracting angels to high office candidacy. Citizens can thus avoid resorting to a perverse or an up-or-out rule that imposes costs by removing angels from public service. But retaining officeholders reduces the size of the subsequent cohort, which at least partly offsets the gain from attracting angel candidates. 12 The great post-war German chancellor, Konrad Adenauer, provides an example of how the political system can benefit when good officials are ejected from low office. In October 1945, Adenauer was sacked as mayor of Cologne by the British Military Government, although Adenauer had been anti-Nazi and there existed no evidence of dereliction of duty. According to his biographer, had Adenauer not been sacked, he would have chosen to remain as mayor and probably would never have entered German national politics. Adenauer’s sacking was unwittingly “the most momentous step taken by any of the Western Occupying Powers in the field of German domestic politics,” for it had the “vastly important indirect result of launching Adenauer into the party-political arena” (Prittie, 1971).

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If politicians discount future payments, then a separating equilibrium becomes possible with Retrospective Reappointment. A separating equilibrium, however, is still less likely under Retrospective Reappointment or Term Limits than under Perverse Reappointment or Life Tenure, because the latter two rules offer shirking knaves a second period in low office. As Table 3 and Figs. 1 and 2 indicate, the cutoff values of β are larger for Perverse Reappointment and Life Tenure. Political scientists typically regard reelection of corrupt politicians and “entrenchment” of incumbents as signs of political infirmity.13 Similarly, McChesney (1997, p. 49) deplores the Federal Election Commission’s failure to punish politicians for violating the prohibition on personal use of campaign funds. But toleration of low office shirking can induce knaves to reveal themselves, which facilitates high office screening. Reappointing shirkers with Perverse Reappointment or Life Tenure is counterproductive, however, if low office shirking for one period suffices to induce a separating equilibrium. Reappointing a shirker reduces the next cohort of young officeholders to one, thus giving a knave appointed in period 2 a free shot at high office in period 3. Needless reappointment of shirkers explains why Perverse Reappointment and Life Tenure have lower peak performance than Term Limits. Reed finds that for reelection to a single office, higher pay unambiguously improves politician performance. In contrast, our hierarchy model reveals that an increase in high office salary improves screening only if the effect on Sh does not excessively attract knaves. Further raising the official salary for low office, V , has two adverse consequences; it reduces the patience penalty (S − V ) knaves must bear to run for high office, and it lowers the payoff ratio Vh /V , making angels less willing to seek advancement. We now analyze the effect of increasing the number of low offices, L. Explicit calculations become impractical for L > 2. Instead we consider only an advancement decision based on the expected number of angels and knaves in a cohort. With all low offices open, the expected cohort of young politicians contains κL knaves and (1 − κ)L angels. If we grant to politicians the option of reappointment to low office, the advancement decision depends only on the payoff ratios for shirking and behaving, Sh /S and Vh /V . With L = 2, a politician succeeds in a run for high office with minimum probability .5; as L increases, this minimum falls. Knaves (angels) try for high office as long as the probability of moving up exceeds V /Vh (S /Sh ). If Sh /S > Vh /V , knaves are more willing than are angels to try for high office, so the set of contenders for high office consists first of knaves, with angels joining the set if the expected number of knaves is sufficiently small. Reversing the inequality yields the opposite conclusion—angels try for high office with knaves joining them only if the expected cohort of angels is sufficiently small.14 With Vh /V > Sh /S , raising L can increase the expected cohort of angels, (1 − κ)L, enough to make the prospect of attaining high office too remote to attract knaves. In this case the set of candidates consists entirely of angels, and increasing L thus improves the screening performance of hierarchy. Even if Sh /S > Vh /V , the candidates can consist mostly or entirely of angels if knaves are attracted by an opportunity to shirk in low office for one or possibly two terms.15 Increasing 13 A notable case involves James Michael Curley, who in 1946 was reelected mayor of Boston after being convicted of fraud and paying an enormous fine. Perhaps significantly, he was never subsequently elected to any office higher than mayor, despite having previously served as governor and U.S. Senator. As if to underscore the attraction of second round shirking in low office, Curley in 1947 continued to hold the office of mayor while serving sentence in a federal prison for conviction on additional fraud charges. 14 Ignoring integer constraints, the equilibrium number of contenders equals either L, κL, (1 − κ)L, V /V , or S /S , h  h  depending on the specific parameter values. 15 Reappointment, however, reduces the next young cohort, giving knaves a better chance to attain high office. For sufficiently large L, the size of the second cohort is max (Vh /V , Sh /S ). Also, the welfare function in (1) does not include

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the number of low offices reduces the expected payoff from pursuing high office, (V + Sh /L), lowering knaves’ opportunity cost of shirking. Increasing L thus gives knaves more incentive to reveal themselves and tends to increase the rate of shirking in low office. Considering L > 2 yields another substantive result: if knaves have little incentive to pursue high office, then in expectational terms Retrospective Reappointment can provide the uniquely best performance. When L > 2, Retrospective Reappointment’s advantage emerges when the prospect of reappointment to low office leads knaves to behave themselves and causes the subsequent small cohort to do the same. In this case, the other rules deny the reappointment incentive for knaves in low office, reducing low office performance (π ) compared to Retrospective Reappointment.16 Retrospective Reappointment is thus best suited to circumstances where incentive effects are relatively more important than screening effects, such as a non-hierarchical (i.e., single-office) system. 5. An application: congressional “Check Bouncing” There’s no job in the world better than being a United States Senator. (Paul Douglas, economist and U.S. Senator, 1949–67) We now demonstrate an empirical application of the hierarchy model by analyzing political shirking in the Congressional “check bouncing” affair. In September 1991, a published report revealed that numerous members of the U.S. House of Representatives had massively overdrawn their checking accounts at the House Bank, an institution that provided accounts exclusively for the 435 members of the House. The report touched off several months of national uproar, as much of the public viewed the episode as a wanton abuse of privilege. Consequently, the House closed the bank and in April 1992 divulged names of overdrawn representatives and their corresponding numbers of overdrawn checks.17 We use the theoretical model of political hierarchy to specify an econometric model of shirking as reflected in overdrafts. We treat the House as low office and the Senate and governorships as high offices. The available data consist of numbers of overdrafts by each of 520 House members who served during the interval July 1, 1988 to October 3, 1991.18 Table 4 displays summary statistics. Notice that 205 representatives made no overdrafts at all. The abundance of zero observations suggests that the stochastic process is subject to some form of censoring. Consequently, we employ

the total cost of legislator salaries, but increasing L while holding constant low office pay (V ) causes this cost to escalate. In addition, expanding the number of low offices may increase the average cost of providing local public goods. These effects, however, are unlikely to offset significantly the gain from improved screening. Legislator salaries represent only a tiny fraction of government budgets, and additional legislators probably do not raise transaction costs in legislatures until L becomes quite large. 16 Note that, with L > 2, Perverse Reappointment continues to weakly dominate Life Tenure. 17 Offending representatives appear to have paid a price. Only 6 of the 17 “worst offenders” serving in 1992 were returned to Congress in 1993. Groseclose and Krehbeil (1994) estimate that 6 (of 53 total) voluntary retirements can be attributed to check bouncing. Similarly, Jacobson and Dimock (1994) find that a representative’s number of bounced checks significantly influenced the probability of retirement and also the probability of electoral defeat. Thorbecke and Matzelevich (1995) find that representatives who bounced more checks also voted for more spending, suggesting that check bouncing correlates with the legislative incentives of individual representatives. 18 We use the list of overdrafts published in Congressional Quarterly (1992). The source for the independent variables is the Barone and Ujifusa, except for gubernatorial term limits and salary, which are obtained from Council of State Governments.

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Table 4 House bank overdrafts by congresspersons: summary statistics

All representatives Overdrawn representatives

Reps.

Mean no. of overdrafts

Standard deviation

Minimum

Maximum

520 315

45.6 75.2

135.0 166.9

0 1

996 996

10–49 70

50–149 55

150–499 21

≥500 14

Frequency distribution of overdrafts 0 1–9 205 155

a familiar two-equation generalization of the tobit model (Heckman, 1976). The empirical model implies a two-stage decision process with one set of determinants of the decision to overdraw an account, and another set of determinants of the number of overdrafts to make. The first step employs a probit model to estimate the probability that a representative overdraws his or her account. Let ODRAWN* be a latent index such that ODRAWN∗i > 0 if and only if representative i makes at least one overdraft. Also, let W be the matrix of variables that influence the probability a representative makes at least one overdraft, with wi a row vector of W. We assume ODRAWN∗i = wi ␥ + νi ,

(4)

where νi is a random disturbance with standard normal distribution and ␥ is a vector of coefficients. We estimate (4) by maximum likelihood. The second step specifies an equation to estimate expected numbers of overdrafts by overdrawn representatives. Let ODRAFTSi represent the number of overdrafts by congressperson i, conditional on i making at least one overdraft. Suppose X represents the matrix of variables that influence the number of overdrafts, and xi is a row vector of X. We have log ODRAFTSi = xi ␪1 + θ2 IMRi + i ,

(5)

where vector ␪1 and scalar θ 2 represent parameters, and i is a random disturbance that shares with νi a bivariate normal distribution with correlation ρ. We estimate (5) by least squares using only the subsample of 315 overdrawn representatives. To account for non-random selection of the subsample used for estimation, the specification includes an explanatory variable known as the inverse Mills ratio (IMR), which is computed from the estimates of the probit model (4).19 The explanatory variables in W and X fall into two categories: state-specific and representativespecific. The representative-specific variables reflect incentives for shirking specific to individual representatives or their districts. These variables are typically used in single-office models of control of politicians and ideological shirking.20 In our hierarchy model, these variables can also have interpretation as measures of variation in κ, the rate of selection of knaves, across House seats. In contrast, the state-specific variables measure the expected value of attaining statewide office. Various state-specific factors can influence the probability of reaching the Senate or governorship, 19 The inverse Mills ratio equals the ratio of the density and distribution functions of the standard normal: ␾(w ␥)/(w ␥), i i where wi ␥ is the fitted value from the probit model (see Greene). The two-equation model collapses to a single equation tobit model in the special case of ρ = 1 and ␥ = ␪1 . 20 In addition to ethical misbehavior, House members can serve constituents poorly by engaging in ideological shirking on roll call votes. We focus on ethical misbehavior since bounced checks offer an unmistakable and easily quantifiable measure of political shirking. Furthermore, studies that attempt to measure ideological shirking in roll call votes are flawed because multiple dimension voting models generally have no Condorcet point (Goff and Grier, 1993).

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as well as the benefits attached to those offices. We can therefore consider each of the fifty states to comprise a unique political hierarchy. Unlike alternative models of politician behavior, the model of hierarchy enables us to derive determinants of House shirking that are state-specific. The state-specific variables in W and X form the basis of our empirical test. The first state-specific hierarchy variable, Delegation Size, equals the number of House seats allocated to each state in reapportionment following the 1980 Census. An increase in the number of low offices reduces the likelihood of promotion and encourages shirking in low office. Hence we predict a positive coefficient for Delegation Size. A Representative has a prime opportunity to advance to the Senate when an open seat election occurs. The probability that retirement opens up a Senate seat increases with the age of the senator. We therefore include in our model the variables Younger Senator Age and Older Senator Age and expect them to affect overdrafts negatively. Since defeating an entrenched Senator is difficult, we also use variables representing the current tenures of the junior and senior senators for each representative’s state. We predict positive coefficients for Junior Senator Tenure and Senior Senator Tenure. We also include two variables relating to the governorship. Term limits make the governorship more attainable by insuring a periodic governor’s race with no entrenched incumbent. We therefore predict a negative coefficient for Governor Term Limit, a dummy variable that equals one if a representative’s state has a Term Limit and zero otherwise.21 To gauge the attractiveness of the governorship we use the official salary and predict Governor’s Salary to have a negative coefficient. We include among the representative-specific variables two measures of the strength of a House member’s political opposition. Opposition Vote equals the percentage of votes cast against a representative in the most recent House race (1988 or 1990). A second measure, 1984 Opposition Presidential Vote, equals the most recent percentage of votes from a representative’s district cast in opposition to the presidential candidate of the representative’s party. Because greater political opposition implies a smaller margin for misbehavior, we predict both of these variables to have negative coefficients. Tenure is the length of the representative’s service in the House at the beginning of the period of bounced checks. We predict a positive sign for Tenure, as representatives who have won more elections have more security in office and can afford more shirking.22 Age is a representative’s age in years at the beginning of the period of bounced checks. We expect a positive coefficient for Age since an older representative approaching the end of his political career sacrifices less (including the opportunity to run for higher office) by shirking. We use two representative-specific explanatory variables to control explicitly for differences across House districts in κ, the probability a knave occupies low office. The value of κ depends on screening accomplished prior to a representative’s initial election to the House. Our two measures rely on the incentive of political parties to maintain their brand-name capital by screening candidates. Party is a dummy variable that equals one for Democratic members of the House and zero for Republican members. We have no a priori expectation concerning this variable.23

21 Term limits also create a positive effect on shirking by abbreviating the flow of benefits from holding the governorship. We presume this effect only partially offsets the negative effect of opening up the office since a first term in the governor’s office tempts a gubernatorial aspirant with a more likely and immediate reward than does a third term. 22 Also, the House’s de facto seniority system for committee and subcommittee chairs makes representatives with longer tenure more valuable to voters, so voters will tolerate more shirking. 23 As Groseclose and Krehbeil point out, Party could affect the value of holding office and thus shirking. Members of the minority party (Republicans throughout the check bouncing period) might attach a lower value to holding office, a possible difference in V across members. On this interpretation Party would be expected to have a negative sign, since members of the Democratic majority have greater incentive to behave.

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The second variable, State Party Strength, equals the percentage of seats in the state legislature controlled by the opposition party at the time of the representative’s initial election.24 Vigorous two party competition should provide better screening of politicians as they rise through the ranks to the House of Representatives. Consequently, we expect a negative sign for State Party Strength. Single-office models of politician behavior generally mirror the predictions of hierarchy regarding the roles of the aforementioned representative-specific variables. For a number of these variables, however, hierarchy can generate additional effects that conceivably alter the predicted signs. For example, Opposition Vote could reflect a politician’s potential to succeed in elections for higher office, which tends to reverse the prediction that a politician facing weaker opposition will shirk more. In contrast, the state-specific variables derive exclusively from the hierarchy model. The empirical relevance of hierarchy therefore hinges on the explanatory power of the state-specific variables. Table 5 presents estimates of probit models (4) and mean equation (5). The fully specified models include in W or X all the hierarchy variables, plus a constant term. Since some representatives did not hold office throughout the entire 39-month observation period, the models also feature a variable equaling the number of months served by each representative. All strictly positive variables are in either log form, or percentage form. Column (a) displays maximum likelihood estimates of the full probit model. The estimated effects of two hierarchy variables, Governor Term Limit and Age, differ significantly from zero at conventional levels. A Wald test reveals the thirteen hierarchy variables to jointly possess considerable statistical significance; Table 6 reports chi-square statistics and p-values. The p-value for falsely rejecting the null hypothesis of zero coefficients equals only about 1/1600. Moreover, the variables of primary interest also prove jointly significant; Table 6 shows that a Wald test on the seven state-specific variables yields a p-value of less than 1/400. The estimates reveal numerous individually insignificant variables, suggesting a high degree of multicollinearity. Hence column (b) of Table 5 presents a more parsimonious probit specification that retains only variables individually or jointly significant at the .10 level. Reducing the multicollinearity leads to additional significant coefficients for Party and two state-specific variables, Governor’s Salary and Older Senator Age. For each of the significant variables except Age, the estimated coefficient signs agree with the a priori signs.25 The estimated coefficient of Governor Term Limit, in addition to being significant statistically, implies an effect of considerable magnitude. The (parsimonious) probit model, evaluated at the median values of the explanatory variables, predicts that imposing a Term Limit on the governorship reduces the probability of overdraft by 19 percentage points: from 81 to 62 percent. The estimated marginal effects of the other state-specific variables are comparatively modest. Increasing Governor’s Salary by $10,000 relative to its median value ($85,000) lowers the predicted probability of overdraft by 1.6 percentage points; adding 10 years to the median (61) of Older Senator Age reduces the predicted probability by 3.3 percentage points.

24 State Party Strength has no available observations on the five Nebraska representatives due to the state’s nonpartisan legislature. Deleting these five from the data reduces the available observations from 525 to the 520 we report in the text and summarize in Table 4. 25 Stewart (1994) formulates an empirical model of check bouncing that features several of the same representativespecific explanatory variables including Age, Tenure, and Party. Stewart’s estimated coefficient signs concur with ours, including the unexpected negative coefficient for Age. The lower incidence of check bouncing by older representatives might reflect the moral scruples of an earlier generation, or possibly personal wealth, which correlates positively with age.

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Table 5 Determinants of house bank overdrafts A priori sign

Dependent variable/(estimation technique) ODRAWN/(ML probit) (a)

Delegation Size Governor Term Limit Governor’s Salary Older Senator Age Younger Senator Age Senior Senator Tenure Junior Senator Tenure State Party Strength Opposition Vote 1984 Opposition Presidential Vote Age of Representative Tenure of Representative Party Months Inverse Mills Ratio

+ − − − − + + − − − + + ? + +

.0539 (.0937) −.583*** (.131) −.601 (.387) −.759 (.767) −.480 (.566) −.0022 (.0126) .0191 (.0159) −.072 (.393) −.205 (.471) .403 (.610) −.619* (.370) −.00059 (.00951) .243 (.188) .505*** (.100) –

ODRAFTS/(OLS) (b) – −.568*** (.128) −.516* (.314) −.767* (.465) – – – – – – −.657** (.288) – .340*** (.118) .528*** (.097) –

(c) .401***

(.151) −.35 (1.24) −1.93 (1.24) −2.02 (2.05) 1.28 (.98) .0297 (.0218) .0225 (.0271) .056 (.663) −.388 (.765) −2.47*** (.92) −3.29** (1.56) .0569*** (.0170) .938 (.797) .87 (1.30) .36 (4.02)

(d) .358*** (.134) – −1.62*** (.55) – – – – – – −2.69*** (.85) −3.11**** (.70) .0644*** (.0164) .801*** (.289) .603** (.291) −.599 (.690)

Observations on individual congresspersons, 7/88 to 10/91. Notes: ODRAWN equals 1 for congresspersons making at least one overdraft, and zero otherwise (520 observations). For overdrawn congresspersons, ODRAFTS equals the log of the number of overdrafts perpetrated (315 observations). All strictly positive variables are in log form or in percentage form. Models (b) and (d) are derived by deleting variables from (a) and (c) not individually or jointly significant at the .10 level. Parentheses contain asymptotic standard errors. In columns (c) and (d), the standard errors are corrected for heteroscedasticity and also for the presence of a generated regressor (IMR); see Greene (1981). The Inverse Mills Ratio is computed from the probit estimates in (b). Party takes the value one for a Democrat and zero for a Republican. *,**,*** indicates significance at .10, .05, .01 levels.

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Independent variables

Null hypothesis

Test

Chi-square, test statistics, degrees of freedom, p-values Model (a)

Coefficients equal zero, all hierarchy variables Coefficients equal zero, state-specific variables Normality Additional statistics Proportion of correct predictions/mean of dependent variable R2 or pseudo-R2 Estimated correlation of disturbances with those of selection model (b)

Model (b) (3.7 × 10−6 )

Model (c) 47.2 [13]

(9.1 × 10−6 )

Model (d) 41.7 [6] (2.1 × 10−7 )

Wald

35.9 [13] (.0006)

33.7 [5]

Wald

22.5 [7] (.0021)

20.5 [3] (.0001)

14.7 [7] (.040)

10.1 [2] (.0063)

LM

1.29 [2] (.525)

.091 [2] (.956)

1.48 [3] (.687)

1.55 [3] (.671)

.662/.606

.663/.606

–

–

.163 –

.157 –

.202 .213

.190 −.349

Notes: Models (a)–(d) correspond to those specified in Table 5. Parentheses contain p-values; brackets contain degrees of freedom [d.f.]. The reported R-squareds for the probit models, (a) and (b), are pseudo R-squareds as formulated by Cragg and Uhler (1970); the R-squareds reported for regression models (c) and (d) are ordinary, unadjusted, R-squareds.

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Table 6 Test statistics

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Column (c) of Table 5 displays estimates of the fully specified mean equation for the log of overdrafts, and column (d) displays estimates of a parsimonious version that excludes all individually and jointly insignificant variables. The same inverse Mills ratio computed from the reduced probit estimates (b) appears in both the full and reduced mean equations.26 Table 5 indicates that four hierarchy variables obtain significant coefficients in the full model: Delegation Size, 1984 Opposition Presidential Vote, Age, and Tenure. Reducing the model to remove some of the multicollinearity reveals two additional hierarchy variables to have significant coefficients: Governor’s Salary and Party. Again, with the sole exception of the coefficient for Age, the significant coefficients all have the a priori signs. Table 5 also shows the estimated mean and probit equations to have in common only three significant variables: Governor’s Salary, Age, and Party. Consequently, the implied process determining the number of overdrafts differs considerably from the implied process for the probability of an overdrawn account. This result justifies our estimation of two stochastic processes as specified by the generalized tobit model, rather than imposing the single stochastic process specified by the standard tobit model. The hierarchy variables possess significant explanatory power for overdrafts, as indicated by the chi-square statistics reported in Table 6. A Wald test for the joint significance of the thirteen hierarchy variables yields a chi-square statistic of 47.2, with only one chance in about 110,000 of obtaining a statistic so large under the null hypothesis. A test on the seven state-specific variables yields a more modest statistic of 14.7, but are still significant at about the .04 level. The generalized tobit model crucially assumes the probit and mean equation disturbances, νi and i , to follow a bivariate normal distribution. To check the validity of this assumption, we perform a test devised by Pagan and Vella (1989). The Pagan–Vella test forms a kind of probit or tobit analogue to the RESET test for linear regression. Although the test specifies the null hypothesis as normality, for the probit model the test can also detect the effects of misspecified mean or heteroscedasticity. Checking the statistical properties of the probit models is particularly important since inconsistent probit estimates can also corrupt the estimates of the mean equation (5) via the Mills ratio (Olsen, 1982). Table 6 presents the resulting Lagrange Multiplier test statistics, which have chi-square distributions. The LM test statistics yield p-values of .53 for the full probit model and .96 for the reduced probit model. Applying to the mean equations a Pagan–Vella test for normality yields p-values of .69 and .67. The Pagan–Vella tests thus fail to reject normality and suggest that our econometric models have valid statistical properties.27 In sum, the generalized tobit estimates demonstrate considerable support for the explanatory power of political hierarchy, manifested in the significance of several state-specific variables. Particularly, Delegation Size significantly influences overdrafts at better than the .01 level; Governor Term Limit significantly affects the probability of an overdrawn account at the .01 level, and Governor’s Salary has explanatory power for both the probability and number of overdrafts.

26 We find that computing the inverse Mills ratio from the full probit model (a) yields invalid estimates of the full mean equation because the resulting estimate of ρ, the correlation coefficient for νi and i , falls outside the interval [−1, 1]. This nonsensical result can occur because the estimate of ρ is not a sample correlation (Greene). The inverse Mills ratio from the full probit does produce valid estimates in the reduced form of the mean equation, but these results do not substantively differ from those of our reported models. 27 We also performed direct tests for heteroscedasticity of the probit models. Like the Pagan–Vella test, the heteroscedasticity test serves as a general specification test, since misspecification of the conditional mean of a probit model also affects the higher order moments. Our LM tests for both probit models define the alternative hypothesis as σi2 = exp[zi b], where z includes the full set of explanatory variables and b is a vector of coefficients. The resulting p-values equal .65 for the full model and .75 for the reduced model; hence we cannot reject the null hypothesis of homoscedasticity.

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Table 7 House bank overdrafts by 34 future senators and governors Observed

Overdrawn reps.

Total overdrafts

17

449

Expected value Random sample of 34 Probit model (a) Probit model (b) Mean equation (c) Mean equation (d)

20.5 19.6 19.5 – –

p-Value for observed .138 .211 .220 – –

Expected value 1526 – – 1293 1284

p-Value for observed .038 – – .101 .097

Notes: p-values are one-tailed and computed at the observed values. For a random sample of 34, the expected values and p-values are estimated by simulating 2500 random draws of size 34, without replacement, from the overall sample of 520 House Reps. For econometric models (a)–(d), see Table 5; p-values for the stochastic processes implied by the estimated models are based on 2500 simulated random draws from each of 34 bivariate normal distributions.

If voters use evidence of misbehavior to screen Senate and gubernatorial candidates, then House members who win promotion to these offices must generally have cleaner records than their unsuccessful opponents. Moreover, if the Senate and governorships offer sufficient reward to attract good candidates, then promoted House members should tend to have cleaner records than their House colleagues. To test these hypotheses, we examine overdraft data for the 34 House members in our sample who served as governor or U.S. Senator at some time before the spring of 1999. Table 7 reports that 17 of the 34 future senators and governors overdrew their House bank accounts and made a combined total of 449 overdrafts. To compare this behavior to that of the whole House, we simulate 2500 random draws of size 34, without replacement, from the entire sample of 520 House members. Table 7 shows the mean number of overdrawn accounts to equal 20.5 and a number equal to 17 or fewer to occur less than 14 percent of the time. The expected number of overdrafts is 1526, and the p-value for 449 or fewer overdrafts equals only .038. On average, the future senators and governors did behave significantly better than their House colleagues. Instead of analyzing only unconditional means, we can also compare the overdrafts of the 34 promoted representatives to the conditional means of our econometric models. We now ask whether the future senators and governors comported themselves well relative to expectations based on their particular representative-specific and state-specific characteristics. Posing this question allows for the possibility that voters evaluate House members relative to circumstances by comparing them to representatives of similar age, from the same or comparable states, etc., rather than comparing them to the average of their House colleagues. To address this question, we simulate the stochastic processes implied by the estimates of Eqs. (4) and (5). First, we simulate realizations of νi and i by generating random draws from the bivariate normal distribution. Next, we obtain 2500 sets of simulated overdrafts by adding our generated νi s and i s to the fitted values of (4) and (5) for the 34 promoted representatives.28 Table 7 displays the results. The probit models predict about 19.5 overdrawn accounts out of 34, with a p-value somewhat larger than .20 for the observed value of 17. The expected number of combined overdrafts falls just short of

28 The fitted values alone do not suffice to compute expected values because of the interaction between Eqs. (4) and (5). We observe the dependent variable in (5) if and only if the latent dependent variable in (4) assumes a positive value. Hence the need for simulation.

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1300, with a p-value of approximately .10 for the observed value of 449. Relative to expectations, the promoted representatives were not particularly likely to avoid overdrawing their accounts, but they did significantly limit their total overdrafts. The evidence suggests that voters did use check bouncing as a relevant criterion for promotion of House members, or at the very least that check bouncing correlates statistically with the relevant criteria for promotion. This result lends support to our choice of the check bouncing episode as an appropriate test case for the predictions of the hierarchy model. 6. Implications for the design of political institutions Corporations also employ hierarchies, and models of corporate hierarchy assume that all workers shirk in the absence of incentives (Lazear and Rosen, 1981; Rosen, 1986). In contrast, our model of political hierarchy assumes heterogeneity among agents in their willingness to shirk. The corporate hierarchy operates as an incentive device by rewarding effort in low office with a payoff in high office that exceeds marginal product. At the top of the corporate hierarchy, executives can earn vastly more than other employees, but political hierarchies exhibit much less gradation in pay. Our model can explain the difference in terms of the tasks that the hierarchies perform. Employing hierarchy as an incentive device requires generous remuneration for the top positions (Rosen, 1986), but high political offices require modest salaries to contain the ambitions of undesirable politicians. Political theorists such as Hume maintain that all politicians are knaves. In our model, if all politicians are knaves, κ = 1 implies that performance in high office is always bad, πh = 0. In this case, political hierarchy is best employed as an incentive device, like corporate hierarchy. Additionally citizens prefer to place constitutional limits on power in high office and to shift powers and responsibilities to low office. Our hierarchies model also provides a unique perspective on term limits and the professionalization of politics. Proponents of term limits are critical of professional politicians who spend their entire adult lives running for office (Will, 1993), claiming that long tenure in Washington produces a Congress out-of-touch with average Americans (Payne, 1991). Opponents of term limits stress the importance of on-the-job learning by politicians, and argue that members of a term-limited Congress would be forced to retire as soon as they learn how to legislate effectively. Under a strict hierarchy, only career politicians hold high office, but the rationale does not involve apprenticeship or learning. Hierarchy makes low office a prerequisite for high office in order to give knaves a chance to take themselves out of the running for high office. In order for hierarchy to screen knaves, citizens must forego professional diversity in high office. In low office, however, our hierarchy model does reveal some novel virtues of term limits: compelling angels to try for high office, opening entry positions for a full cohort of next generation politicians, and convincing knaves they face long odds of ascending to high office. The political system could counteract some adverse effects of professionalization by allowing inexperienced candidates to bypass the hierarchy and run immediately for high office, but such outsiders carry greater risk. How readily can other forms of public service substitute for political apprenticeship? The answer depends on the relative importance of character and political skills in determining performance in high office. Military leaders often acquire records of public service that provide evidence of their characters. A military career, however, might offer little opportunity to exhibit or develop the political skills necessary to govern effectively. The surprising performance of the Perverse Reappointment rule points out the value under some circumstances of reappointing shirkers to a second term in low office. The benefits from reappointing shirkers, however, are mostly external: all citizens benefit from keeping knaves

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out of high office, but local district residents bear the cost of low office shirking. The Perverse Reappointment rule further burdens local voters by removing satisfactory politicians. Since local voters have an interest in enhancing low office performance, they might find Retrospective Reappointment more appealing than Perverse Reappointment. If Perverse Reappointment proves to be infeasible in a democratic hierarchy, local democracy might not be desirable. To attain the optimal reappointment rule, society might have to make the hierarchy appointive rather than elective. For instance, local magistrates in the republic of Lycia were appointed by a representative body called the Common Council, an arrangement praised by Montesquieu (1966 [1748], Book IX) and by Hamilton (Federalist #9). Our hierarchies model suggests that newly democratic nations face an unavoidable dilemma: the first generation of leaders could not have risen through the ranks. The breakdown of an authoritarian regime or decolonization might enable knaves to disguise their type long enough to reach the top. If knaves are relatively likely to acquire the first generation leadership posts, this might help to explain why democracy often does not survive the first leaders, that is, why so many nations have experienced “one man, one vote, one time.” If the first cohort of leaders should happen to be angels, democracy survives and subsequent generations of politicians have the opportunity to reveal their types before assuming power. Political scientists have in fact observed that democracies are typically most fragile when new, but once consolidated democracy is reasonably stable. For instance, following World War II, several nations of Africa and Asia achieved independence, and among them 12 new democracies were established. Eight of these democracies did not survive past 1975. Similarly, the global turmoil of the 1930s extinguished 13 of 17 new democracies that had originated after 1910, but of the 12 older democracies that dated to before 1910, only one perished (Huntington, 1991, pp. 15–20). To the extent that getting the wrong people at the top can endanger democracy, the screening effects of hierarchy potentially assume a crucial role in preserving democracy’s viability. Acknowledgements We thank Stephen Coate, Tyler Cowen, Kevin Grier, Robin Grier, Joe McGarrity, and Eric Schansberg for helpful comments. Earlier versions of this paper were presented at the University of Oklahoma and at meetings of the Public Choice Society and the Southern Economics Association. References Austen-Smith, D., Banks, J.S., 1989. Electoral accountability and incumbency. In: Ordeshook, P. (Ed.), Models of Strategic Choice in Politics. University of Michigan Press, Ann Arbor. Banks, J.S., Sundaram, R.K., 1997. Electoral accountability and selection effects. In: Paper Presented at 1997 Public Choice Society Meetings. Barone, M., Ujifusa, G., 1996. The almanac of American politics. National Journal, Washington DC. Barro, R.J., 1973. The control of politicians: an economic model. Public Choice 14, 19–42. Black, G.S., 1972. A theory of political ambition: career choices and the role of structural incentives. American Political Science Review 66, 144–159. Brennan, G., Buchanan, J.M., 1983. Predictive power and the choice among regimes. Economic Journal 93, 89–105. Brennan, G., Hamlin, A., 1995. Economizing on virtue. Constitutional Political Economy 6, 35–56. Coate, S., Morris, S., 1995. On the form of transfers to special interests. Journal of Political Economy 103, 1210–1235. Congressional Quarterly, 1992. Congressional Quarterly Almanac, 1992. Congressional Quarterly, Washington, DC. Council of State Governments, various years. The Book of the States. Council of State Governments, Lexington, KY. Cowen, T., Glazer, A., Zajc, K., 2000. Credibility may require discretion, not rules. Journal of Public Economics 76, 295–306.

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