Rationality of executive compensation schemes and real accounting … · executive compensation...

Rationality ofexecutive compensation schemes

and real accounting changes*

A. RASHAD ABDEL-KHALIK University of FloridaCHARLES CHI University of Florida

DIMITRIOS GHICAS Baruch College - CUNY

Abstract. Managerial preference for accounting methods is examined with respect to theeffects ofthe chosen methods on executives' annual compensation (salary and bonus). Twopropositions are considered: (a) the bonus-hypothesis, and (b) the rationality of compensa-tion schemes. While the former asserts a tie between compensation and accounting income,the latter stipulates a connection between the real (cash flow) consequences ofthe choice ofaccounting methods and executives' compensation.

Two accounting method changes were used: (a) the change in pension costing andfunding (which increases both income and operating cash flows), and (b) the change ininventory valuation method to LIFO (which decreases income and increases cash flows).Unexpected compensations in the switch-year were correlated with the effects of eachaccounting change on income. The results suggest that top executives' salary and bonuspayments increased in the switch-year beyond levels predicted by the pre-switch compensa-tion parameters. Several validation checks were also performed using comparison samples.Furthermore, the effects of these two accounting changes on income were found to correlatewith unexpected compensation in the directions predicted by the hypothesis of rationality ofcompensation schemes.

Resume. Les pr6f6rences de la direction quant aux methodes comptables sont examineesrelativement aux effets des methodes retenues sur la r6mun6ration annuelle des dirigeants(salaires et gratifications). Deux propositions sont envisagees: (a) l'hypothese de gratifica-tion, et (b) la rationality des systfemes de r6muneration. Alors que la premiere propositionaffirme l'existence d'un lien entre la remuneration et le b^n6fice comptable, la seconde est al'effet qu'il existe une relation entre les consequences reelles (flux de tresorerie) de lamethcxle comptable choisie et la remuneration des dirigeants.

Deux types de changement de m^thode comptable furent consideres: (a) le changementdans la comptabilisation des co(its et la capitalisation des regimes de retraite (augmentant ala fois le beneflce comptable et le flux mon6taire d'exploitation), et (b) le changement demdthode d'^valuation des stcxrks pour DEPS (diminuant le bdn^fice comptable et augmen-tant les flux de tresorerie). Les remunerations imprevues versees dans l'annê oil il y eutchangement ont 6te corr616es aux effets de chaque modiflcation comptable sur le bdneflce.

* The authors would like to thank Gary Biddle, Steve Kachelmeier, Sundaraman Thiagarajan,an anonymous reviewer and Haim Falk for commenting on earlier drafts. This paper, which isunrelated to any dissertation work, was presented at workshops at the University of Wyoming, theOhio State University, University of Nebraska. University of Connecticut and Memphis StateUniversity. We are grateful for the comments received in those workshops.

Contemporary Accounting Research Vol. 4 No. 1 pp 32-60

Executive Compensation Schemes 33

Les r^sultats semblent indiquer que les salaires et gratifications versus aux didgeants se sontaccrus, dans l'annê oii il y eut changement, au-delk des niveaux predits par les paramfetresde remuneration existant avant le changement. Plusieurs controles de validation ontegalement 6t6 effectuds, et ce au moyen d'echantillons comparatifs. De plus, ll a eteconstate que les effets de ces deux types de modification comptable etaient correies auxremunerations imprevues versees, conformement aux attentes predites selon l'hypothfese derationalite des systSmes de. remuneration.

IntroductionThe consequences of accounting choices on executive compensations have beenconsidered as infiuential determinants of managerial preferences for certainaccounting methods. In an agency setting with separation between ownership andmanagerial control, owners cannot observe managers' actions and hence, opt forthe second best solution (Holmstrom (1979)) which ties their compensation to theoutcome of their decisions. The accounting measurement of income is often usedas the performance measure to which the compensation scheme is tied. Profitsharing arrangements are often offered in order to induce incentive compatibilityso as to mitigate the manager-owner confiict of interest emanating from theirseparate pursuit of wealth maximization. Consequently, the choice of accountingmethods is expected to be of personal interest to executives to the extent that thechosen alternatives infiuence the income base used in determining the profitsharing (bonus) pool. Thus, when managers change accounting methods, they ineffect signal their preferences for different income determination rules. Theconcept that the choice of accounting methods is motivated by the effects on theexecutive's profit sharing is known as the bonus hypothesis. Under this hypothesis,executives are assumed to manage the determination of their corporate accountingincome so as to maximize their annual bonus awards.

The problemOne objective of establishing performance-contingent bonus awards is to inducetrue incentive compatibility between owners and managers in a setting where thereis information asymmetry and where real production decisions made by executivesare unobservable. In offering the managers a share in the final outcome of theirdecisions (profits), owners expect them to expand the level of effort and to makethe types of decisions that will increase the value ofthe firm. Decisions resulting incosmetic accounting changes are not necessarily included in the admissible set ofactions that lead to value maximization of the firm, unless these decisions havenonpecuniary consequences such as avoiding noncompliance with contractualarrangements like debt covenants. In general, making income-increasing cosmeticaccounting changes merely for the purpose of increasing executives' bonus awardsnegates the purpose of incentive sharing schemes, since such changes do notnecessarily generate any wealth increments for the owners. Typically, no cashflows are associated with these accounting changes, and empirical evidence hasshown that securities markets are not fooled by cosmetic accounting changes (for areview, see Lev and Ohlson (1982)).

34 A.R. Abdel-khalik C.Chi D. Ghicas

Furthermore, measuring income with the intent of enhancing executives' honusawards implies that (a) such a behavior was anticipated and was considered insetting up the bonus plans; (b) managers do not expect owners to renegotiateexecutives' incentive contracts; or (c) owners find it less costly to allow executivesto increase their income through the manipulation of measurement rules than totake actions so as to affect ex post settling up. Evidence in support of any of thesethree possibilities is lacking, although some authors have shown an inclinationtoward the assumption of high renegotiation cost (Holthausen and Leftwich(1983)). In any case, managers are assumed to be concerned with near term (oneperiod) pecuniary rewards irrespective of the form of ex post settling up thatowners or the markets for executives might take (Fama (1980)).

On the other hand, the case for ex post settling up still needs to be made. Forexample, Simon's (1986) procedural theory of rationality, which suggests thatbonus plans based on corporate profits, is made in order to underscore thesignificance of the bottom line in the manager's thinking. Thus, basing executivecompensation on firm profits serves as a constant reminder of the need to increaseincome to owners. Consequently, the psychological reinforcement of the beliefthat executives' own income is affected by the firm's income (as their effort anddecisions) leads to one important implication: executive bonus awards are tied toincome of the firm irrespective of the income measurement rules used. Indeed,Simon's procedural theory of rationality is a basis for another interpretation of thebonus hypothesis.

If, in contrast, executives are considered to be rational decision makers who aimat maximizing the expected present value of their compensations, then the validityof the bonus hypothesis might be questioned. In particular, executives cannotconsistently generate rewards by managing the measurement of accounting profitsirrespective of the consequences of such changes on the wealth of the firm. Trueincentive compatibility implies that executives are to be rewarded for the conse-quences of those decisions, including accounting method changes, that addpositive increments to the economic resources of the firm. Systems of executivecompensation that induce managers to prefer accounting methods that enhance thevalue of the firm are considered rational compensation schemes. In this frame-work, executives' ordering of preferences for accounting methods is assumed toconsider the jointness of the effects of accounting method choices on the wealth ofboth owners and executives. In particular, managers would be rewarded formaking accounting method choices that add positive increments to the wealth ofthe firm. Only then can compensation schemes be considered incentive compatible.

Research objectiveThis paper examines the consistency of the behavior of executive compensationswith the bonus hypothesis vis-a-vis the hypothesis stating that compensationschemes are rational economic systems. The environment used for the test is one inwhich the altering of preferences for accounting methods has different effects onboth net cash flows and the measurement of accounting income. The two


accounting changes chosen are real in the sense that each method change has a netpositive effect on net operating cash flows, but their accounting income effectsdiffer. The change in actuarial cost methods for pension accounting (Type A,hereafter) led to increases in both accounting income and net cash fiows. Bycontrast, due to the tax laws in the US, the change in the inventory valuationmethod to LIFO (Type B, hereafter) decreases income, but increases casb fiows.Assuming a tax rate of about 50 percent, the amount of the decline in income forthe Type B change will be about equal to the increase in net cash fiows.

Under the bonus hypothesis, the amounts of bonus awards will be based onaccounting income; hence, the amounts of awards would increase beyond expectedlevels for the Type A change, but would decrease below expected levels for tbeType B change.' In contrast, under the alternative hypothesis suggesting thatincentive compensation schemes are rational economic systems, executive com-pensations will increase beyond expected levels for both types of accountingchanges, since these incentive schemes take into account the effect of accountingchanges on net cash fiows of the firm. Furthermore, under the alternativehypothesis, unexpected compensation is expected to be significantly correlatedwith the effects of accounting changes on cash fiows.

Plan of the paperThe remainder of this paper is structured as follows. Prior research is briefiyreviewed in the next section. The research design, a brief discussion of the chosenaccounting method changes and statements of operational hypotheses follow. Thechosen samples, the data used in the analysis, and estimated compensation modelsare presented in a separate section, followed by a presentation of the results ofpredicting compensation levels and tests of the operational hypotheses. The lastsection presents a discussion of the results in view of the structure of certain bonuscontracts. Concluding remarks are also presented.

Prior researchThe accounting literature on the bonus hypothesis presents mixed empiricalresults. The studies by Hagerman and Zmijewski (1979) and Zmijewski andHagerman (1981) test, but do not support, tbe assertion that managers makeincome increasing accounting changes whenever annual bonus awards are basedon accounting income. In those as well as in several subsequent studies (e.g.,Collins, Rozeff and Dhaliwal (1981); Holthausen (1981); Leftwich (1981),Bowen, Noreen and Lacey (1981); and Hunt (1985)), the bonus contract variablewas represented by a dummy variable [1-0] designating whether or not bonus

1 This expectation is correct when the proforma income numbers of the year of the switch are usedjointly with parameters estimated in the preswitch year to predict the expected levels of com-pensation. However, we used the reported income numbers (inclusive of accounting changeeffects). The statements of hypotheses formulated in the paper are adapted to reflect this phenom-enon. The directions of unexpected compensation should be adjusted accordingly, as will bepresented below


awards are based on accounting income. Using the same approach. Watts andZimmerman (1978) examined the effect of income-based bonus contracts onFASB's lobbying behavior. They recognized the possible rationality of thesystem, however, in suggesting that managers should be rewarded for makingcertain accounting decisions such as switching to LIFO even though the switchreduces reported accounting income (Watts and Zimmerman (1978), fn. 17p. 116).

Reporting mixed results, however, is consistent with several plausible explana-tions. The choice of a 1-0 dummy variable to represent the income-based bonuscontract may have generated an inadequate research design. The studies by Antleand Smith (1985), Abdel-khalik (1985) and Healy, Kang and Palepu (1985) haveattempted to avoid this problem by examining the dollar amounts of executivecompensation. Abdel-khalik's (1985) study concluded that bonus awards did notseem to follow the income effect of accounting changes to LIFO and thatadjustments to bonus contracts (such as basing bonus on FIFO income) have beenreported by several companies at the time they switched to LIFO. Healy, Kang andPalepu (1985) examined both LIFO and depreciation switches and arrived at con-clusions that are at variance with Abdel-khalik's. The differing results generatedby those two studies might be explained by noting that the LIFO switch is a realaccounting change (i.e., it affects the firm's net worth), or by differences inresearch design and analysis. The contradictory findings, however, add to themixed nature of the earlier evidence and suggest the need for a different approachto the investigation of both managers' preferences for accounting methods and thevalidity of the bonus hypothesis.

The approach adopted here highlights the relevance of making a distinctionbetween real and cosmetic changes in generating the expectations of executivesabout consequent actions of ex post settling up. Increasing bonus awards concomi-tantly with the increase in the firm's accounting income emanating from changedmeasurement rules is likely to be done for one of two reasons: (1) the cost toowners is lower than the cost of renegotiating incentive contracts (Holthausen andLeftwich, 1983); or (2) the belief that such an increase refiects positive incrementsin the owners' wealth. While the former reason continues to be an untestedassertion, the latter implies a consistency between accounting changes and theexpected behavior under the true incentive compatibility objective. The assump-tion of costly renegotiation requires further empirical verification, although theavailable evidence does not appear to point in that direction (Abdel-khalik (1985)).Further discussion of this point is presented in the last segment below.

In contrast, the assertion of a rational compensation scheme assumes thatrewarding executives for altering their preferences for accounting methods islimited to real accounting changes (i.e., those having positive effects on cashflow). This behavior would be consistent with Fama's ex post settling uphypothesis. Fama (1980) argues that the labor market for executives (both internaland external) can discriminate between managerial skills based on the ability to


make good and bad decisions and adjusts the market price of executive servicesaccordingly. It is thus implied that owners will settle with executives who continueto he employed by the firm and, therefore, do not face the test of extemal labormarkets. In particular, the payment of higher bonus awards might be reversed insubsequent periods when owners realize that the preferred income increasingaccounting measurement methods (causing the increase in bonus awards) werecosmetic and do not add positive increments to owners' wealth. This form of expost settling up does not require executives to enter the labor market since variousmeans can be used for adjusting compensation, including making adjustments tothe salary portion of the compensations, which requires the approval of stock-holders in their annual meetings.

An altemative analysis of the relationship between markets, performance, andexecutive compensation is provided by Simon (1986). He argues that executivecompensations do not bear a relationship to the market determined "managerialworth of the executives or even that the fluctuations in profits are the result of theincentive compensation" (1986, p. 222). Simon suggests that a procedural theoryof rationality would predict that incentives would be offered to executives toincrease (not necessarily maximize) profits. The procedural theory of rationality"would assume only that executives (and corporate boards) believe tbat executivescan influence profits through their bebavior and that a bonus plan thereforemotivates them to try harder" (1986, p. 222).

If in fact Simon's description is empirically valid, then owners might reinforcethe belief that profits do count by allowing executives to reap extra bonus awardsby changing preferences for income measurement rules. In this scenario, incen-tives exist for owners to avoid taking actions that would undermine this belief.That is, if managers alter accounting measurement rules in order to increase theirown bonus awards, owners need not seek ex post settling up since doing so wouldundermine the connection between the bottom line and managers' own pecuniaryrewards. It is assumed, however, that (a) income in general is highly correlatedwith performance; (b) the proportion of income generated from altering measure-ment rules is nonsignificant, or (c) owners' cost of monitoring the partition ofincome into performance and measurement rules is high. As indicated earlier,Simon's procedural theory of rationality might in fact be another basis for assertingthe validity of the bonus hypothesis.

To summarize, there are two competing hypotheses: (a) the bonus hypothesis,and (b) the rational (cash flow based) incentive systems. The former asserts thatmanagers' rewards are consistent with changes in accounting profits, even if suchincreases emanate from altering preferences for accounting measurement rules.The bonus hypothesis is consistent with Simon's procedural theory of rationality.In contrast, the altemative hypothesis of rational (cash flow based) incentivessuggests that managers are rewarded for niaking accounting method choices thatincrease the real wealth of the firm, even if the measurement of income is adverselyaffected. Prior research has thus far provided mixed evidence.

38 A.R. Abdel-khalik C. Chi D. Ghicas

Research design and hypothesesThe approach used in this paper consists of prediction and validation of compensa-tions of top executives within a pretest, posttest research design. Accountingchanges constitute the events studied, with the year of changing accountingmethods being the test period. Executive compensation models are estimated foreach sample for the period preceding the test year. These pretest period models arethen used with the income numbers reported (inclusive ofthe change effects) in thetest year to provide the amounts of compensation expected for the test period. Thevalidity of the forecasted amounts of compensation are thus predicated on twoassumptions: (1) that performance-related parameters of compensation models arestable during the two-year period, and (2) that reported income numbers are thedeterminants of bonus awards. A test of parameter shifts in compensation modelsfrom pretest year to posttest year is also provided.

If accounting changes do not lead to change in the parameters of managementcompensation plans, we should expect the two assumptions to be valid and theunexpected compensation (excess of actual over predicted) to be equal to zero. Onthe other hand, if because of accounting changes, the parameters of managementcompensation plans change, the unexpected portion of compensation should besignificantly different from zero. The bonus hypothesis would be sustained if thelevel of unexpected compensation generated from proforma income (i.e., as if theaccounting changes were not made) is positive for pension and negative forinventory changes. Alternatively, the bonus hypothesis can be said to hold if theunexpected compensation generated from reported income is about zero for bothchanges. These expectations can be expressed by the following hypotheses (usingpredictions generated from reported income, inclusive of effects of accountingchanges):

1 Hypotheses consistent with the bonus hypothesis:

HBP\ UC = 0 for pension change, and

HBL- UC = 0 for inventory change.

2 Alternative hypotheses consistent with the cash flow based (rational) compen-sation systems and assuming stationary compensation parameters:

HRP: UC = 0 for pension change, and

fl/fi,: UC > 0 for inventory change.

3 The alternative hypotheses assuming that the proportion of income paid ascompensation increased in the year of the accounting change over the pretestyear:

H/ip: UOO for pension change, and

HKL: UC > 0 for inventory change,

where UC is the unexpected compensation component, P and L stand for pension


and LIFO switches, respectively; B stands for bonus hypothesis and R stands forthe cash fiow (rational) hypothesis. However, the link between the unexpectedcompensation and the effect of accounting changes on income (cash fiows) needsto be established in order to further validate either the bonus or cash fiow (rational)hypothesis. Obtaining positive correlations between unexpected compensationamount and the cash flow effects of accounting method changes would beconsistent with the hypothesis that managers were rewarded for having made thesereal accounting changes. However, the cash flow and income effects of theseaccounting changes are (1) consistently positive for Type A, and (2) opposite forType B, where positive net cash flows correspond to negative effects on reportedincome. Consequently, an additional approach for evaluating the rationality ofcompensation systems (or cash flow hypothesis) is to examine the correlationbetween unexpected compensation (UC) and the accounting income effects (EI)of these accounting changes.

Since the pension switch (Type A) increases both net income and cash flows inour test sample, and since UC is based on reported (as opposed to proforma)income, a zero correlation between UC and EI would be consistent with both thebonus and rational (cash flow) hypotheses if model parameters (the percentage ofincome paid as compensation) did not change. If the parameters of compensationincrease concurrently with changes in the accounting method, then a positivecorrelation between UC and EI would be expected. (A priori one would not expectdecrease in compensation.) However, the Type B sample provides the opportunityto distinguish between the two hypotheses, since the effects on income and cashflows are exactly opposite for those firms switching to LIFO. A zero correlationbetween EI and UC is consistent with the bonus hypothesis since UC is based onreported income (i.e., inclusive of £/). However, a negative correlation betweenUC and EI is consistent with the rational (cash flow based) hypothesis underconditions of both stable and increasing parameters.

Two additional steps are taken for further validation. The compensation modelsused for prediction in the test year were also estimated from two other samples thatdid not make accounting changes during the corresponding test periods. Each ofthese samples was selected (as discussed below) for comparison with one of thetwo switch samples. Hence, unexpected compensations used in the analysis wereestimated twice; once using the experimental samples' own models, and onceusing the models generated from the comparison samples.^ Furthermore, since the

2 Using compensation model parameters of the control samples to predict expected compensationfor the corresponding switch samples was reasoned as follows. It is assumed that the labor marketfor executives is efficient, such that executives are paid close to the worth of their marginal pro-duct. As a consequence, a consistent relationship between an executive's pay and his perform-ance would be expected to hold in a given year for a similar cross-sectional sample of companies.Consequently, the compensation model estimated for the comparison sample can proxy for theparameters for a firm selected at random in the same time frame. This reasoning can be extendedto a set of firms such as the switch sample. Changing accounting methods, however, breaks theconsistency of relationships between compensation and performance, and the use of parametersgenerated from another cross-sectional sample gives the effect of accounting changes on compen-sation the greatest chance to be detected.


results attributable to test periods should not be expected to hold for other periods,the posttest year was also used for validation. More discussion of the design ispresented later, but for the purpose of this section, the hypotheses tested in thisstudy can be stated as follows:

4 The null hypothesis consistent with the bonus hypotheses for both Type A andType B changes:

Ho: r(UC, El) = 0.

5 Alternative test hypotheses consistent with the rational compensation schemes(cash fiow based compensation) under conditions of stable parameters:

HRP: rp{UC, El) = 0

6 Alternative test hypotheses consistent with the rational compensation schemes(cash flow based) under conditions of increasing compensation parameters con-currently with accounting changes:

H^p: rpiUC, El) > 0

HRI^: rdUC, El) < 0,

where

r = the correlation coefficient;

UC = unexpected compensation;

E! = effect of accounting change on income;

and other notations are as defined earlier for hypotheses (l)-(3).The signs between HRP and Hî^ are different due to the perverse effect of the

inventory switch (in the US) on accounting income - i.e., depending on the taxrate, the decrease in income is about equal to the increase in cash flows. Thus,obtaining a negative correlation between unexpected compensation and the incomeeffects of Type B accounting method changes is equivalent to obtaining a positivecorrelation if the effects of the change on cash flows were used instead of effecton income.

Empirical analyses

Samples

Type A changeThe first sample is for the pension accounting change. Prior to the issuance ofSFAS No. 87 (1985), accounting for pensions was governed by APB Opinion No.8 (1966). In general, the normal cost component of the pension represents the


current period's shareof the present value of pension benefits. APB OpinionNo. 8allowed the use of an appropriate actuarial method in the determination of normalpension costs. Two broad classes of actuarial methods have been suggested; (1)the benefit allocation methods, and (2) the cost allocation methods.^ The behaviorof normal costs differs under these two categories (see Winklevoss and McGill(1979)) in that the normal cost of any cost allocation method exceeds the normalcost of any of the benefit allocation methods in the early periods of work cycles,but the relationships reverse in later periods.** Accordingly, under certain condi-tions , the change in actuarial methods could lead to a reduction in the measurementof normal pension cost and the attendant reduction in the cash outflow required forfunding.

The sample selected for this accounting change (Type A) consists of companies

3 The benefit allocation methods allocate the benefits of the plan to the benefit years and use theapportioned benefits for the determination of the annual cost. The most common benefit alloca-tion methods ate (1) accumulated plan benefit, (2) level dollar benefit, and (3) level percentagebenefit. Those cost allocation methods allocate the costs of prospective benefits to the planyear without directly allocating the benefits. The most common cost allocation methods are (a)level dollar cost method, and (b) level percentage cost method.

4 A basic difference between the two mam approaches is that the cost allocation methods assign alarge amount of normal pension costs to the early years of a pension plan, while the benefitallocation methods allocate large amounts of normal costs to the later years of a pension plan.Winklevoss and McGill (1979) reported simulated results for the five different methods (three forbenefit allocation methods, and two for cost allocation methods) and reported the cost of pensionsas a percentage of payroll and the number of years of service. The contrast between the vanousmethods is shown in the following graph.

4.3

I 3.7

Cu

I3.0

2 72.6

25 50Years of service

A = Accumulated plan benefit method (unit credit).B = Level percentage benefit method (modified unit credit).C = Level dollar benefit method (modified unit credit).D = Level percentage cost method (entry age normal).E = Level dollar cost method (entry age normal).

Source: Winklevoss and McGill (1979, p. 231).


that made the change in actuarial methods so as to reduce the amounts of accruednormal cost, which results in increasing both income and net operating cash fiowsby the reduced funding requirements. Using the recent mandated (SFAS No. 35(1980)) disclosure in financial statements, a sample of 30 companies was selected(13 changed in 1982,10 in 1983 and seven in 1980-81). These companies use thesame actuarial methods for both funding and expense accruals. Changing actuarialcost methods has reduced the pension expense by an average of 32 percent and hasincreased net income by an average of 16 percent.' These effects were identified inboth annual reports and Form 5500 of the Department of Labor, which is requiredby ERISA.

A comparison sample was selected to provide a benchmark for studying theswitch sample.* The comparison sample consists of 32 companies that did notdisclose changes in actuarial methods in the years of analysis for which data for theswitch sample were collected (1980-83).

Type B changeThe second pair of samples was selected as (test) experimental and control samplesfor the inventory valuation method change (Type B). Given the US tax code, theaccounting valuation change to LIFO is accepted for tax purposes only if it is usedconcurrently for external reporting. Hence, under conditions of increasing inputprices, growth, and positive incremental tax rate, the switch to LIFO reduces thetax burden, but also results in reporting lower accounting income. From a largenumber of companies that made the switch in 1974-75, a sample of 89 companieswas selected. The data requirements led to a reduction of the inventory sample to74 companies, of which 68 made the switch in fiscal period 1974.

Initially, a comparison (FIFO) sample of companies of equal size was alsoselected, but data requirements reduced this sample to 63 companies. These were

5 A 16 percent average increase in income due to the pension accounting change appears to be toohigh. The reason for this relatively high percentage is that several firms in the switch samplehad relatively low income in the year of the change. This is why the average percent change issignificantly greater than what could be reasonably expected.

6 An initial randomly selected sample did not serve as a good comparison group due to significantdifferences in size as measured by total assets. Consequently, the initial random sample wascombined with another sample of firms for which proxy statements were available in the Account-ing Research Center at the University of Florida for the relevant period, 1980-83. From thecombined samples, a smaller sample of 32 companies was selected as in pair-matching by select-ing a comparison sample that had the least difference in total assets. Subsequent to selecting thecomparison sample, two firms of the switch sample were deleted for lack of data.

The industry composition of the switch and control samples (using four digits) is as follows:

Industry digits

Switch sampleComparison sample

1000

35

2000

106

3000

1114

4000

2

5000

24

6000-1-

41


companies that did not make the inventory change in 1974-75 and for which allrequired data were available from public sources.^

The data required for estimating compensation models were gathered fromcompanies' proxy statements, annual reports, the Compustat tape and the ValueLine Investment Survey. Descriptive statistics showing relevant distributionalcharacteristics of these samples are presented in Table 1 and Table 2.

Basic compensation modelIncome based performance compensation plans provide a direct relationshipbetween corporate income and executive compensation. Thus, a simple modelmight be constructed: bonus = / (income). This function, however, cannot beestimated directly, due to the difficulty of obtaining information about the size ofannual bonus awards for most companies. In general, proxy statements providecombined numbers for salary and cash bonus awards, and no information isprovided to separate the two components. The dependent variable then becomesthe combined salary plus bonus, which will be termed P.*

Prior research (e.g., Ciscel and Carroll (1980); Murphy (1985); and Abdel-khalik (1985)) found that the income and compensation data have skeweddistributions and suggested that logarithmic transformation (or some other form ofBox-Cox transformation) should be used. Thus, the basic compensation modeltakes the form ln(P) = / (In Income), where In is the natural log. This model,however, is an incomplete specification of the compensation function due toomitted variables. Two possible omitted variables are: (1) the extent to whichinsiders own shares in the stock of the companies they manage, and (2) the degreeof attractiveness of the stock to institutional investors.

Income-based bonus schemes are established in order to motivate incentivecompatibility in cases where ownership and control are separated. Thus, themotivational aspect of this incentive scheme is greater when executives generatemore of their income from employment than from stock ownership. To capture thiseffect, researchers have dichotomized companies into owner-controlled andmanager-controlled using a cut-off point generally set at ten percent for totalinsider ownership. The association between this variable and executive compensa-

7 The initial comparison sample ofthe Type B switch was selected from the Forbes 1000 based onindustry matching with the switch sample. The Forbes list was chosen because all ofthe switchfirms were also on that list. In selecting the comparison sample, recognition was given to theindustry classification only. However, eliminating some firms due to incomplete data resulted inless than complete industry matchmg as shown in the following tabulation:

Industry digits

Switch sampleComparison sample

1000

53

20OO

3226

3000

2816

4000

63

5000

112

6000H-

3

8 Less than five percent of the proxy statements examined have reported separate numbers forsalary and bonus awards.

I6

V3

00 ^ - - ^ f*^ — - ^

ex,

II <« II „ „ o <

S

S I

, r-- .-H ..- -M . . - „.-

00 — — —

T3 MS

O S K u•S o g_ c.

IflflIliiiif-'ri


TABLE 2Additional descriptive statistics for the four samples for the preswitch year

Item

Type A: switch

Inc = Income'P = Salary + bonus"TA = Total assets^

In Inc.\nPlnTA

Type B: control

Inc = Income"P = Salary + bonus*"TA = Total assets'

In Inc.lnPinTA

Type B: switch

Inc = Income'P = Salary + bonus""TA = Total assets'

In Inc.\nPInTA

Type B: controt

Inc = Income'P = Salary + bonus""TA = Total assets

In IncinPInTA

Mean

68389914

10.315.87

12.94

51378918

10.305.86

13.11

95250

1333

11.015.58

13.75

65249980

10.725.43

13.47

S.D.

87179

1270

1.430.441.28

59144

1177

1.130.421.11

136100

1391

0.860.420.83

71101895

0.850.440.82

Min

0.5013836

6 204.93

10.50

1.4010042

7.204.61

10.63

9.40100157

9.154 61

11.97

7.906394

8.974.14

11.45

Max

361935

4521

12.806.84

15.32

262741

4439

12.506.61

15.31

844600

9082

13.646.90

16.02

413581

4836

12.936.36

15.39

Notes:' m million dollars"" in thousand dollarsIn = natural log.

tion has been mixed (see, for example, Abdel-khalik (1985); and Hunt (1985)).Although a possible explanation for these mixed results is the loss of informationentailed in dichotomizing the ownership variable, the use of percentage of insiderownership as a continuous independent variable did not perform differently.^

9 The issues about which measure (either percent of insider ownership, or a dichotomous variable)to use have not been fully explored in the literature. For example, if it is a matter of insiders'control, the degree of control may not be a monotonically increasing function of the percentage ofownership. That is, top managers either do or do not have control. Similarly, the trade-off be-tween expected earnings from stock holdings versus expected increases in bonuses requires moreinvestigation.


Concentration of ownership in the hands of institutional investors is the secondindependent variable used to augment the basic compensation model. A relativelylarge institutional holding is an indication of a greater degree of confidence bysophisticated investors in the company's financial prospects. Further institutionalinvestors are likely to invest in firms that are well managed by highly skilledmanagers who in tum will demand higher compensation. One implication is thatthe brighter the prospects of a company, the more its executives will be willing totie a larger percentage of their compensation to its income.

Augmented compensation modelWith the inclusion of the percentages of insider ownership and institutionalholding, the compensation model takes the form:

In /'y = ay + Ply ^ InCy + PlylOy + ^3ylUy -i" Cy , (1)

where:

In = natural logarithm;

P = Salary plus annual cash bonus (botb of which are voted upon by stockholdersannually);

Inc = accounting income;

10 = insider ownership;

IH = institutional holding;

a = constant term;

P = estimated coefficients;

e = error term with expectations zero;

y = year of estimation.

Based on the above discussion, the directions of the estimated parameters areexpected to be as follows:

7 a>0

8 0<p, <1

9 P2SO

10 0<P3.

Model estimationTwo approaches were used for estimating equation (1): ordinary least squares(OLS), and weighted least squares (WLS). OLS regressions were estimated foreach sample separately for each year: the year prior to the switch iy= —I), and forthe year of the switch (y = 0). Since differences in company size could generate


significant levels of heteroscedasticity, each sample was rank ordered by the sizeof ln Inc. The top and bottom third of the sample were used to test for homogeneityof variance by testing the equality of their sum of squared errors using the Golfieldand Quandt F-test (Johnston (1972, p. 218)). The results indicated that three ofthe eight regressions did not satisfy the homogeneity of variance assumption.Consequently, the weighted least squares (WLS) method was applied using thevariance of the independent vadable, ln Inc, as the weight. However, none of thefindings obtained with OLS were altered using WLS.'°

Additional diagnostic checks applied were related to the effect of outliers onestimated results. Cook's D statistics" were generated for both types of theestimation procedures, OLS and WLS. The results indicated no significant Dstatistics for the sample observations reported here. ^ This conclusion was thesame for both the OLS and the WLS estimations. Because of this consistency, onlythe results based on OLS are reported in this paper. The OLS estimates arepresented in Table 3 and Table 4 for Type A and Type B accounting methodchanges, respectively.

The estimated models suggest the following relationships:a Consistent with expectations, the coefficient of income. Pi, is significantly

greater than zero but less than unity (at p < 0.01);b Also consistent with expectation, the intercept term is significantly positive (at

p < 0 . 0 1 ) ;c The coefficient of IH was significantly positive for the LIFO switch sample, but

not for any otber;d The coefficient of 10 was significantly negative for the FIFO sample, but not for

others;e All the models have reasonable levels of explanatory power resulting in signi-

ficant F-statistics (at p < 0.001).Notwithstanding the consistency of expected and obtained directions for a and

Pi, the coefficient of income, Pi, cannot be used to evaluate the effect ofaccounting changes on bonus awards for two reasons: (i) the models are multipli-cative, which means that the intercept term does not actually measure the fixedcomponent (e.g., salary) in the compensation function; and (ii) it is observed thatevery increase (decrease) in a is accompanied by a decrease (increase) in PJ .Therefore, it is important to examine the direction of changes in the compensation

10 The similarity of the results is essentially in terms of significance of the coefficients and func-tions; the values of parameters were similar but not identical.

11 Cook's D examines the extent to which deletmg an observation (a company) from the sampleaffects the estimated function. A discussion of the test is presented in Belsley, Kuh and Welsch(1982). A rule of thumb is that the observation is influential in estimating the model if Cook's DIS about one.

12 Initially, the comparison sample for the inventory change had 66 companies, but Cook's D wassignificantly large for three companies. Consequently, those three companies were deleted and theremaining 63 companies were used in the analysis reported. However, the use of 66 companies,through altered diagnostic statistics, did not change the findings from what is reported here forthe 63 companies.


TABLE 3OLS estimation results of compensation model Ml for the pension change and comparison samples

Statistics

Interceptit)

PI; Coefficient of income

P2: Coefficient of insiders(0

P3: Coefficient of inst. hold(1)

F-statisticAdj. R^

* Significant at p < 0.01

Model; In P,y = a,j. + pi,, Inc,,

Sample; Pension switch sample(n = 30)

Year= - 1

4.09*(6.6)

0.174*(2.42)

-0 .18(-0.50)

0.0003(0.01)

4.76*0.28

+ p2lO,v + P3II

Year of acctg.changey = 0

3.45*(6.27)

0.230*(4.13)

-0.15(-0.39)

0 0049(1.11)

8.70*0.44

H,y + e,y.

Comparison sample(n = 32) (Random)

Corresponding to

Year -1

3.36*(5.50)

0.220*(3.88)

0.62(1.67)

0.0020(0 54)

8.27*0.41

YearO

3.48*(7.50)

0.250*(4.99)

0.33(0.98)

-0.0040(-1.11)

10.07*0.47

TABLE 4OLS estimation results of compensation model Ml for the LIFO and FIFO samples

Statistics

Intercept(f)

Pi: Coefficient of incomeit)

P2: Coefficient of insiders(0

P3: Coefficient of mst. hold(0

F (3:70) and (3:59)Adj R^:

* Statistically significant at p

Sample: LIFO samplen = 74

Yearbefore thechangey=-l

2.686*(5.58)

0.244*(5.70)

-0.0021(-0.90)

0.0130(2.00)

13.40*0.34

< 0.01 (one tail)

Model: In P^ = a,, + pw. In Inc., + P2r,IO,j. +

Year ofthe acctg.change

3.381*(7.06)

0.190*(4.45)

-0.0008(-0.33)

0.0110(2.10)

8.92*0.25

P3iyIH,, + e,y.

HFO sample n =

Corresponding to

Yeary=-\

1.910(3.60)

0.340*(6.78)

-0.0063*(-2.50)

-0.0026(-0.46)

20.96*0.49

63

Year

1.910(3.22)

0.340*(6.24)

-0.0076*(-2.50)

-0.0023(-0.50)

19.06*0.47


parameters from the prechange year to the year of accounting change. It is alsoinstructive to report the results of the same test for the postswitch year.

Test of changes in compensation parametersTwo tests were conducted: (1) intercept-restriction: estimating the compensationmodel (equation 1) for the year of the switch with the intercept restricted at thelevel estimated in the preswitch year; and (2) slope-shift parameter: estimating acompensation function that contains two years' data, but allows for one interceptonly and for a slope-shift parameter. Both approaches were applied to the basic andaugmented compensation models. Since the relevant independent variable for thisstudy is corporate income and testing the slope-shift parameter provided the sameresults for the basic and the augmented regression models, only the results of thebasic model are reported here. Furthermore, the results of intercept restriction andthe slope-shift approaches were similar. Thus, only the findings of regressionestimates for the slope-shift approach are reported here.

Estimating the slope-shift parameter for the basic compensation model took thefollowing form:

In Py,y_, = a + Pi lnlncy,y.-i + \i D* lnlncy,y_i + e (2)

where y and y — 1 are any two-year combinations of ( -1) , preswitch year, (0), theswitch year, and (+1) the postswitch year; D is a dummy variable representing theyear in the pair of years used to test the parameter stability (D = 0 for the earlieryear in the pair, and D = 1 for the later year in the pair); d is a common intercept;X.1 is a slope-shift parameter for Pi; and other terms are as before.

For each of the four samples used in the study, equation (2) was estimated threetimes to test the parameter stability (S) of the compensation function for the pair[y, y — l]as follows:

S[y, y — l] = S[0,1] to test the slope-shift over a two-year period from the pre-switch year to the year of making the accounting change.

S[y> ^ — 1] = 5[+1,0] to test the slope-shift during the two-year period from theyear of accounting change to the subsequent year.

S[y, y — 1] = 5[+1, — l ] to test the slope-shift during the three-year period fromthe preswitch year to the postswitch year (the middle year is excluded).

The regression results of estimating equation (2) for the Type A (pension)accounting change and for its comparison sample are reported in Table 5. Table 6contains the corresponding results for the Type B (inventory) accounting changeand for its comparison sample. As shown, the slope-shift parameter, \ , isstatistically significant (at p < 0.05) for the third test of each sample coveringS[4-l, —I]. Since this test covers a three-year period (with the middle yearomitted), it is not surprising that the small accumulation of changes in adjacentyears shows statistical significance over the three-year span. In contrast, the


TABLE 5Test of income slope-shift of compensation models for the pension (Type A) switch and comparisonsamples

Type A: switch

SIO,-l]Coef,(0

5[+l,0]Coef,(0

5[+l,-l]Coef.(0

Type A: comparison

5[0,-l]Coef.(0

5[+I,0]Coef.(0

5[+l,-l]Coef.it)

a

3.70(10.3)*

3.70(10.1)*

4.00(11.5)*

3.55(10.3)

4.03*(13.0)*

3.93(11.6)*

P.

0.21(6.1)*

0,22(6,3)*

0,18(5.4)*

0.22(6.7)

0,19(6,3)*

0,19(5,7)*

X,

0.014(1.5)'

0.005(0.6)'

0.020

0,013(1,7)"

0,005(0,7)

0,018

Adj«^

0,40

0.39

0,36

0,44

0.39

0.39

F(n;dn)

20.5(2; 57)

20.6(2; 57)

17.5(2; 57)

25.3(2; 61)

21.3(2; 61)

20.77(2; 61)

Notes:S[y, y - 1] = Stability test between year (y - 1) and year (3-).(*) = Coefficient is statistically significant at p < 0.01.(a) = Significant slope-shift (p < 0.01) covering a three-year span,(b) = Significant slope-shift (p = 0.10) covering a two-year span.(F) = All F-statistics are significant at p < 0.001,(c) = Not significant slope-shifts,(n; dn) = Degrees of freedom for numerator and denominator.

slope-shift parameters for adjacent periods were either statistically not differentfrom zero or marginally significant. However, only for the Type A comparisonsample, and for Type B switch sample did the slope-shift parameters showstatistical significance (at p = 0.10, two-tailed) for the switch year in comparisonwith the preceding year. In general, these results suggest that the shift incompensation model parameters on income, though positive, were marginallysignificant (greater than zero) over any two adjacent years.

Results

Unexpected compensationExpected salary plus bonus in the switch year was estimated for the realizations ofthe independent variables in the switch year (e.g., using reported income), butusing the parameters estimated for the year preceding the accounting change. As


TABLE 6Test of income slope-shift of compensation models for the inventory (Type B) switch and comparisonsamples

Type B: switch

5[0,-l]Coef.(0

5[+l,0]Coef.(0

5[+I,-l]Coef.(0

Type B; comparison

S[0,-l]Coef(0

5[+l,0]Coef.(0

5[^fl,-l]Coef.(0

a

3.04(8.9)*

3.57(12.1)*

3.03(11.1)*

1.60(4.1)

2.40(6.3)*

2.40(6.6)*

P,

0.22(7.1)

0.18(6.7)*

0.19(7.2)*

0.36(9.9)

0.29(8.4)*

0.29(8.5)*

0.008CLZ)"

0.006(1.3)'=

0.014(3.n°

0.004(0.8)'

0.006(1.2)'

0,012

Adj.^

0.28

0.24

0.31

0.45

0,36

0.40

F(n;dn)

28.9(2; 145)

24.1(2; 145)

33,8(2; 145)

52,2(2; 123)

36,6(2; 183)

43.1(2; 123)

Notes:S[y, >' - 1] •= Stability test between year iy - I) and year (y).(*) = Coefficient is statistically significant at /) < 0.01.(a) = Significant slope'Shift (p < 0.01) covering a three-year span.(b) = Significant slope-shift (p = 0,10) covering a two-year span,(F) = AU F-statistics are significant m /> < 0,001,(c) = Not significant slope-shifts.(n; dn) = Degrees of freedom for numerator and denominator.

has been shown, it is reasonable to infer that the parameters ofthe compensationmodels are stable over adjacent years. Therefore, obtaining significant deviationsof actual from predicted compensation levels (UC) for the switch year would beconsistent with one of two possibilities; (i) changes in compensation parametersthat are unrelated to accounting income or the accounting method change, or (ii)changes in compensation parameters that are precipitated by the accountingchanges. Tests of rationality of compensation schemes (i,e,. in terms of cash floweffects) are to be based not only on whether the UC is positive for the switchsamples, but also on the significance levels of the correlations stipulated by thealternative hypotheses in (5) or (6) indicated above.

Predicting compensations for the switch year is essentially a one-period aheadforecast, which is expressed as [—1,0]. By way of validation, two-period aheadpredictions were made for the period after the switch, which are denoted [ - 1 , +1 ].It is expected that prediction errors will be greater for the longer prediction


horizon. In both cases, the first number indicates the year for which the parameterswere estimated, and the second number indicates the year for which compensationforecasts were developed. Finally, another set of one-period ahead compensationpredictions was developed using the models estimated in the year of the switch,which are denoted [0, + l ] . The objective of this last procedure is to evaluatewhether the effect of the accounting change on compensations persists in futureperiods. Obtaining an average UC not different from zero would imply a lack ofcontinuity of the effect ofthe accounting change on compensation.

A further validation is generated by using the parameter estimates of thecompensation models ofthe comparison sample in predicting expected compensa-tion levels for the switch (test) samples. In this case, the compensation models ofthe comparison samples are used as estimates of instrumental variables that areneutral to the impact ofthe accounting changes studied. In either case, predictionsare generated for the test (experimental) sample, since these are the samples forwhich accounting method changes had an impact on income. Thus, for anyprediction interval, there are two underlying models, depending on whether theestimated parameters were generated from the test (J) or comparison (C) sample.Consequently, six different expectations are generated:

one-period ahead: r[ - 1 , 0 ] ; r[0, +1 ]; two-period ahead: 71 - 1 , +1 ]

one-period ahead: C[ - 1 , 0 ] ; C[0, -t-1 ]; two-period ahead: C[ - 1, +1 ]

where the numbers 0,-1,-1-1 indicate the year of the switch, the year before, andthe year after the switch, respectively.

For each of the six combinations, unexpected compensation, UC, was com-puted as: UC = [actual lnP - Predicted lnP].

The results of testing the hypotheses about mean unexpected compensation{UC) are reported in Table 7. Based on the f-test, the following observations canbe made:a Average UC for the one-period ahead prediction is less than the average UC for

the two-period ahead prediction. This observation is consistent with the resultsof tests on the slope-shift parameter.

b The unexpected compensations for the switch year are significantly greater thanzero for both types of accounting changes when the switch samples' models areused. As shown, for the Type A switch the r-statistic is 1.73 {p < 0.05,one-tailed), and is 2.0 {p < 0.025, one-tailed) for the Type B switch. Both arefor the 7T-l ,0] prediction.

c Average unexpected compensation for the postswitch year was significantlygreater than zero only when the prediction covered three years, i. e., models forthe preswitch year were used in prediction ([—1,-1-1]). However, averageunexpected compensation was not different from zero for the one-period aheadprediction, i.e., when the switch period models were used {T[0, -f 1], orC[0,+1]).


TABLE 7Significance tests for unexpected compensations (UC) of the switch samples

Type A: Pension switchUsing test models

for prediction

Using com. modelsfor prediction

Type B: Inventory switchUsing test models

for prediction

Using com. modelsfor prediction

Mean UCU)

MeanUC(.t)

Mean UC(t)

Mean UCit)

Prediction interval

[-1.0]

0,12(1.73)"

0.12(1.60)

0.08(2.00)"

0.00(0.00)

[O. + l]

0,07(1.03)

0.08(1.08)

0.08(0.35)

0.04(1.00)

[ -1 , + 1]

0.19(2.68)»

0.18(2.53)'

0.16(4,08)^

0.09(1.98)'

(a) Significantly greater than zero at p < 0.025 (one tail test).(b) Significantly greater than zero at p < 0.05 (one tail test)(f/C) is measured as (In actual compensation — In predicted compensation).(0 is student /-statistic for difference of mean UC from zero.

d The combined results reported above suggest a significant unexpected increasein the compensation of top executives in the year of the switch. It also suggeststhat UC is inconsistent with the bonus bypothesis as posited by HBP and HBL,and is consistent with the alternative hypotheses as posited by H^p and H^L-Results of the correlation tests concerning the extent to which these positive

unexpected compensations in the year of the switch are related to accountingchanges (hypotheses (4) through (6)) are reported in Tables 8 and 9. Also includedis information about the effect of accounting changes on income. As indicated, theincome (and cash fiow) increased by an average of about $2 million for pensionfirms, while income decreased (and cash fiow increased) by an average of about$21 million for inventory switch firms.

Hypotheses test resultsThe /-statistics of UC provide the evidence concerning testing hypotheses (1), (2)and (3) to see if the unexpected compensations are, on average, significantlydifferent from zero. As indicated, the Mests for UC in the switch years aresignificantly positive (p < 0.05) for 7T"~1,O] for both changes. That is, the/-statistics reject Ho (at P < 0.05), suggesting that executive compensationincreased concurrently with the two accounting changes. In order to further exam-ine the validity of the bonus hypothesis versus the rational scheme hypothesis, weneed to test hypotheses (4)-(6). These tests are reported in Table 8 for the pension(Type A) change, and in Table 9 for the inventory (Type B) change.

As reported in Table 8, tbe correlation between (one-period ahead) unexpectedcompensation and the effect of the accounting change on income (r(UC, El)) is

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about -(-0.37 for T[-l,O], and is about -1-0.38 for C[-1,0] . Both are significantlydifferent from zero (at p < 0.05). That is, irrespective of which models are used inpredicting compensation, the unanticipated increases in compensation are signifi-cantly positively correlated with the increases in income (and cash flow) attribut-able to the accounting change. The possible causal link implied by the aboverelationship draws further support from the absence of a similar pattern for thepostchange year for both the one-period and two-period ahead forecasts. Thecorrelations between the effect of accounting changes on income and unexpectedcompensation resulting from two-period ahead and one-period ahead predictionsfor the postswitch year are reported in the last two columns of Table 8. None of thecorrelations (between unexpected compensation and the effect of the change onincome) for each combination of model/period (T[-l, + l], T[0, -f-1], C [ - 1 , +1]and C[0,+ 1]) is significantly different from zero at conventional levels ofsignificance.

The results related to the inventory (Type B) change are reported in Table 9. Asexpected, the correlation between (one-period ahead) unexpected compensationand the effect of the accounting change on income (cash flow) is negative(positive) and is different from zero (-0.28 for C[-1,O]) but marginal (-0.14)for T[-l,O]. While the former is significantly less than zero (at p < 0.05, one-tailed test), the latter is not (p = 0.11). Furthermore, the negative correlationsbetween unexpected compensation and the effect of accounting changes on incomepersisted in the postchange period. As shown, the correlation for C [ - l , + l] isequal to -0.23 (significant at;? < 0.05), and for C[0, +1] is about -0.24 (p <0.05). In contrast, when the test sample's own models were used in makingtwo-period ahead predictions, the correlation coefficients were not different fromzero. Since all correlations are for the switch sample, the difference in results bythe type of model used in predicting compensations implies that executivecompensation for the switch firms has experienced an upward shift concurrent withthe accounting change that was not temporary and that was not fully captured bythe models presented here.

The combined results of these tests suggest that the changes in the compensationof top executives of the switch firms appear to be consistent with the effect ofaccounting changes on cash flows. For both types of accounting changes,executives appear to have been rewarded for making the switch.

In summary, the following points can be made:a Executive compensation of the switch firms (for both Type A (pension) and

Type B (inventory) changes) has increased in the year of the accounting changebeyond what can be anticipated using the models of the preswitch year. Thisobservation holds for both types of accounting changes, even though theyaffected accounting income in opposite directions,

b The unanticipated increases in compensation for both of the switch samples arepositively correlated with the effect of such changes on cash fiows.

c Concurrent with the switch period, an upward shift in the income parameters ofcompensation models is observed, but the shift is not always significantly


different from zero. Thus, rewarding executives for making these two account-ing changes is not systematically captured by the compensation models used.The results, however, tend to be consistent with the hypothesis of rationality ofcompensation schemes.

DiscussionInformation asymmetry in organizations leads to the design of compensationsystems based on observable outcomes such that owners and managers have trueincentive compatibility. The owners' objective is assumed to be the maximizationof the value of the firm. Since income represents a measure of the increments inowners' wealth, accounting income numbers are often used as the outcomemeasures upon which managers' performance is evaluated. In a real sense, thewealth increments that owners seek to maximize are those real (as opposed tomeasurement-induced) changes in the economic value of the firm. Thus, profitsharing compensation schemes would be effective systems for inducing incentivecompatibility to the extent that accounting income numbers are indeed measures ofthe outcomes of economic decisions only. However, the significant degree ofmanagerial judgment allowed under present accounting standards introduces thepossibility of managing income such that the correspondence between accountingincome and economic increments in wealth is distorted. Nonetheless, researcherscontinue to assert that executives do consistently manage the measurement ofincome so as to maximize their own annual bonus awards. If managers do soconsistently, does it follow that owners can be fooled into paying extra amounts inbonus awards for managed measurement of income? The evidence provided in thispaper adds to growing evidence suggesting that income-based bonus awards arenot necessarily the result of a mechanical application of a bonus formula toaccounting income numbers. The evidence suggests that positive unexpectedbonus awards are associated with making accounting changes that increase realincrements in owners' wealth, even if such changes decrease accounting income.

In fact, an examination of several proxy statements suggests that well-managedcompanies seem to guard against the capricious manipulation of bonuses byproviding for an additional built-in bonding. Contracts stipulating that annualbonus awards are to be disbursed over several future years are examples of this. Toillustrate, the 1984 incentive system of RCA Corporation requires that the annualcash bonus awards related to the performance of any given year be paid in twoequal installments over the two following years (see, RCA's 1984 Proxy State-ment). A similar practice is followed by GM: the bonus related to any year ispaid over a period of three years, including the year of the award. In its Notice ofAnnual Meeting of Stockholders and Proxy Statement, General Motors disclosedthat the cash bonus awards relative to 1983 are payable in three installments: in1984,1985, and 1986 (pp. 10-11). It was also disclosed that the Bonus and SalaryCommittee had met 12 times in 1983 to consider matters related to the incentivecompensation system: amounts, eligibility and timing of disbursement.

Furthermore, an executive's eligibility to participate in some incentive plans


can be left to the discretion ofthe bonus committee, which presumably considersreal performance. Consider the following statement from the GM plan:

The purpose of the Performance Achievement Plan is to provide employees in approxi-mately 500 positions of major responsibility with incentive compensation related to theachievement ofthe Corporation's long-term performance goals. Participation in the Plan byeligible executives is not automatic but is determined by the Bonus and Salary Committee,except that the Committee has the sole discretion with respect to participation by officersand directors (emphasis added, GM Proxy Statement, May 25, 1984, pp. 12-13).

Two important issues are indicated by this quote: (i) safeguards are providedagainst the possibility of making bonus awards to executives if their actions do notcontribute to real achievement of corporate goals, and (ii) the discretion built intothe incentive system renders it fully adaptable to changes in eligibility and amountsof bonus awards without renegotiation of contracts. Indeed, the lack of evidenceabout the cost of contract renegotiation cannot be underestimated.

Finally, under the bonus hypothesis the assumption is made that offering bonusawards to a few top executives leads to the process of making accounting methodchoices. However, the cost to the company of making an income-increasingaccounting choice extends far beyond the amounts of bonus awards made to topexecutives. In the case of GM, for example, the total performance-related cashbonus awards made in 1983 amounted to over $90 million, but only about twopercent of that amount ($1.8 million) was paid as bonus awards to the top fiveexecutives. Thus, even if owners preferred an income-increasing accountingchange that would double the amounts of bonus awards for the top flve executives,that expectation alone should not be the sole factor considered in making theaccounting change, since the full cost ofthe bonus system resulting from adoptingsuch a change would most likely exceed the amount awarded to the top fiveexecutives. That is, if the bonus system at GM remains unchanged when anaccounting method change is made, every one dollar increase in the bonus awardto the top five executives in GM could conceivably cost the company up to 50additional dollars in bonus awards to the other (500) executives sharing in thebonus incentive scheme. Thus, the cost of bonus awards to top executives alonecannot be a good indicator of the bonus cost of an accounting method choice.

ConclusionIn order to reduce agency costs, true incentive compatibility implies that managersshare in actual increments in owners' wealth. Consequently, it is not clear thatmanagers can consistently increase their bonus awards by manipulating themeasurement of accounting income irrespective ofthe effects on real wealth. Thetest results provided here are consistent with this implication: top executives'compensation changes in a manner consistent with the real (instead ofthe income)consequences of accounting changes. In addition, it has been suggested thatseveral bonding factors act to mitigate against executives' ability to consistentlymanage the measurement of income for the purpose of increasing bonus awards.


For example, executives must be cognizant of potential recourse by owners whenthey realize that executives have caused the increase in their bonus awards byartificially inflating income.

The realism of ex post settling up is questioned, however. As indicated earlier,Simon's procedural theory of rationality implies that owners might want managersto generate bonus awards from income increases even if such increases came aboutfrom managing measurement rules. The reason for this implication is that ownerswish to reinforce the belief that corporate income and executive rewards arerelated.

Finally, several corporations do defer the payment of bonus awards and holdthem in "escrow" by requiring that bonus awards of any year be paid over a numberof future years. An outcome of this additional bonding is that managers would beless likely to shirk their responsibilities which would be the outcome of enhancingtheir own bonus awards from manipulating the measurement rules of reportedincome.

ReferencesAbdel-khalik, A.R., "The Effect of LIFO-Switching and Firm Ownership on Executives'

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