disclosure of capital lease and stock prices

8/4/2019 Disclosure of Capital Lease and Stock Prices

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Journal of Accounting ResearchVol. 16, No. 2 Autumn 1978Printed in U.SA.

The Disclosure of Capitalized LeaseInformation and Stock PricesBYUNG T RO*

1. IntroductionThis researcb empirically examines whetber tbe recent Securitiesand Excbange Commission (SEC) decision calling for disclosure ofcapitalized accounting data for noncapitalized financing leases badany impact on tbe pricing of securities. The results of tbis researcbmay be relevant in evaluating the lease disclosure requirements setforth by tbe Financial Accounting Standards Board, as well as by theSEC.In June 1973, tbe SEC proposed to amend Rule 3-16' of Regulation

S-X calling for improved disclosure of lease information by lessees intheir 10-K reports filed witb tbe SEC.^ After evaluating many lettersof comment received in response to the proposal, tbe SEC eventuallyadopted tbe extended lease disclosure requirements in AccountingSeries Release (ASRJ No. 147 in October 1973.^ The Release waseffective as of N ov em be r 30, 1973." A ssistant Professor, Purdue U niver sity. T his paper is based on my 1976 Ph .D. the sisat Michigan State University. I am grateful to my committee, Ronald M. Marshall(chairman), Daniel W. Collins, and Richard R. Simonds. This research was generouslysupported by the Haskins & Sells Foundation. I also appreciate comments and


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3 1 6 BYUNG T. ROThe Release requires disclosure of certain lease information notpreviously required by either the SEC or other accounting regulatoryagencies. The new items of mandatory disclosure, which concern

noncapitalized financing leases, are (1) the present value (PV) of theminimum future lease commitments, (2) the interest rate(s) implicitin computing the PV and (3) the impact on net income (income effect,IE ) if such leases w ere capitalized.*Two reasons for the extended lease disclosure requirements arespecifically pointed out in the i?e/ease.^ First, the disclosure of thePVand IE numbers is considered "essential to investors," since thisinformation is felt to be "necessary to enable investors to comparemeaningfully the capital and asset structures and the operatingresults" of the companies that have lease commitments. Second, theexisting lease disclosure requirements, including APB Opinion No.31, are not sufficient to provide lease information "as needed byinvestors."These statements suggest that the capitalized lease data of the typecalled for in the Release will convey new information (important toinvestors) about the risk -return prospects of those firms which did notpreviously report such data through accounting sources. This sugges-tion, in turn, implies that the security prices of these firms will beaffected by disclosure of such data through its impact on investor'sassessm ents of the ris k-retu rn prospects of the firms.The SEC's view on the importance of capitalized lease informationis not new. Rather, it can be viewed as a reinforcement of thetraditional arguments that support disclosure of capitalized leaseinformation in financial reports. According to these arguments,* leasecommitments create property rights (assets) and liabilities on the partof lessees.Consequently, capitalized lease data will convey new infor-mation about the risk-return attributes of the lessee firms, anddisclosure of such data will improve "the information content of thefinancial s tate m en ts" of the lesseefirms.There is a counterview which asse rts th at a lease commitment doesnot give rise to an asset and a liability of the lessee firm, and thatdisclosure of capitalized lease data will not improve the informationcontent of the financial statements. As a result, capitalized lease

' For further de tails of the disclosure requirem ents, see "C. Am endm ents to Regula-tion S-X" of ASR No. 147.' See "A. Introduction" oiASR No. 147.


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CAPITALIZED LEASE INFORMATION 317

disclosure will be of little value in evaluating the market value of adisclosing firm, so tha t the costs of such disclosure may exceed itsbenefits.*Despite the importance and complexity of the issue of capitalizedlease disclosure, surprisingly little research has been und ertaken toevaluate this issue empirically. If capitalized lease data convey anynew information important to investors, as both the SEC and theproponents of capitalized lease disclosure believe, one can expect toobserve a market reaction to the disclosure of the capitalized leasedata under ASR No. 147. To tbe extent that the market reactionexists, it is possible to evaluate the information content of the

capitalized lease data and the effect of the SEC lease disclosuredecision on the pricing of securities. However, if the capitalized leasedata do not convey any new information about the risk-return pros-pects of the firms, as tbe counterview asse rts, then there should be noempirically observable market reaction to disclosure of such data. Nomarket reaction implies no information content of the capitalized leasedata and this, in turn, implies that the SEC decision had no impact onthe p ricing of securities.2. Assessment of Effects of Capitalized Lease Disclosure onSecurity Prices

This section first examines why the disclosure of capitalized leasedata, such as the PV and IE numbers, in financial reports may havean effect on the equilibrium pricing of securities, and then discussesthe research design used to detect such an effect. The framework ofthe section, including the selection of the statistical test method, isbased on the idea originally introduced by Gonedes [1975] and laterused by Gonedes, Dopucb, and Penman [1976] and Harrison [1977].POSSIBLE EFFECTS OF CAPITALIZED LEASE DISCLOSURE

Suppose that a (lessee) firm capitalizes a financing lease whichallows it to use a property under the lease contract for a certain periodof time, in return for periodic ren tal pa ym ents. Then the firm does notcharge periodic rental expenses to income. Instead, tbe firm willrecognize depreciation expense on the leased property and interest


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3 1 8 BYUNG T. ROfuture rental payments. In addition, the firm will report the leasedproperty as an asset and the lease obligation as a liability.In this case, if the amounts of periodic rental payments are not thesame as the amounts of periodic depreciation and interest expenses,which is very likely, different periodic net income will result. Thuscertain income-statement-related financial ratios, including operatingrisk measure, would be affected if the results of lease capitalizationwere incorporated into computing such ratios. Similarly, certainbalance-sheet-related financial ratios, such as financial risk measure,would also change if the leased asset and the lease liability wereincluded in calculating such ratios.It appears, therefore, that investors' assessments of the firm's risk-return characteristics, including operating and financial risks, wouldbe affected if capitalized lease data were made public. The reason isthat certain financial ratios incorporating the result of lease capitali-zation might h ave been used by investors in evaluating the risk-returncharacteristics of the firm. As a result, the effect of capitalized leasedisclosure on the assessments by investors may lead to differentevaluation of the firm's market value.^The discussion above seems to be typical of firms which havenoncapitalized financing leases. The SEC's assertion that the disclo-sure ofPV and IE num bers under ASi? No. 147 is essen tial to investorsimplies that investors would form different assessments of the risk-return characteristics of the firms with noncapitalized financingleases, if such numbers were available to them. Therefore, theirinvestment decisions conditional upon the capitalized lease numberswould differ from investment decisions that would otherwise beformed. Different investment decisions by investors would then allowfor the possibility that different equilibrium security prices mayresult.ASSESSMENT OF EFFECTS WITH CONTROL BY MATCHING

If equilibrium security prices were affected by capitalized leasedisclosure, one would expect to observe the following condition: RIO i=R where R and 6 are column vectors of ra te s of re tu rn and capitalizedlease numbers, respectively. However, it is impossible to observe thetwo types of re tu rn da ta for a firm for a given time period. To approachth is problem, a group of control firms w ere employed.


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CAPITALIZED LEASE INFORMATION 3 1 9. J ^ p lt i l lx ed -JteterU l Group 2J ^ p l t i l l x e d -J t e t

/

Honcancelablef rinanclng'lea ses \ / leasea \Voncfcpltallzed; Dmnaterlal

\

HOnc.px^ai.i.*c


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3 2 0 BYUNG T. ROFurther criteria used to select both treatment and control firms areprovided later.To control for confounding factors whicb might conceal possibleeffects of capitalized lease disclosure under investigation, treatmentand control firms were individually matched, primarily according totbe estimates of their relative risks (betas)," assuming that the well-known capital asset pricing model (CAPM) holds.'^ Tbe beta estim ates()8's) were available from Security Risk Evaluation (September 1973)publisbed by Merrill Lynch, Pierce, Fenner and Smith Inc. Thepublished /3's were estimated, via tbe market model,' as of August1973'^ using sixty prior monthly returns. In equating tbe beta esti-mates, it was assumed tbat fir = Pc at tbe individual firm level, priorto the SEC lease decision, leads to PT = Pc at tbe group level as well.A matcbed pair design of tbis type was assumed to be effective tereduce tbe undesirable effect of regression phenomenon involved inestimating /3's. (See Harrison [1977].)In matching firms individually, an attempt was made to considerindustry factors as well. However, tbe result of tbis considerationturne d out to be not as successful as it was inten ded (see Appendix A).Tbe main reason was tbat tbe relatively strict definition of control

''' The reason for using the relative risk as an attribute, as Gonedes and Dopuch[1974] and Gonedes [1975] note, is based on the fact that classifjdng firms according totheir relative risks is appropriate in investigating the effect of new information on thevalues of firms, because this attribute is generally determined by firms' production-investment and financing decisions. A theoretical discussion about how the relative riskis related to firms' production-investment and financing decisions is found in Fama andMiller [1972, chap. 7] and Hamada [1972].'-' E{R\El.Rz),E{Ru), p) = E(R,)I + [{RM) - E(R,)]pwhere (i?z) = expected re turn on a zero beta asset w ith Cov (R^ Ry = 0.

/ = an n-component un it column vector.E{Ru) = expected re turn on the m arket portfolio which consists of all securitiesin the market./3 = an n-component column vector defined by Cov (R RyyWai (fiu) whichmeasures the relative risks of the individual assets.'* The m arket model is given by:where aj, ft = regression coefficients.R.M = return on the market portfolio.ij = error term with(j) = 0and Cov iit^i) = 0 for i ^j.


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CAPITALIZED LEASE INFORMATION 3 2 1firms, as defined before, led to an insufficient number of potentialcontrol firm s. '*Based on the matched pair design, a (column) vector of monthlyreturn differences were computed as follows:

(1)The return differences were then used in investigating the possibleeffects of capitalized lease disclosure. Note that the effects wereinvestigated by focusing on total return, rather than any particularreturn component.In order to examine whether the effects of capitalized lease disclo-

sure did vary according to different levels of riskiness of firms," firmsin each group were classified into two different risk groups, high andlow, using the relative risk as an attribute.^" After the risk groupswere constructed, a two-component return (column) vector {R" R'-)was computed by tak ing the arithm etic (equally weighted) averag e ofthe component individual return differences in each of the two riskgroups. Each monthly average return difference was then used as asample unit.^' The two-component return difference vectors can beexpressed as:

'* The reason for giving some consideration to th e SIC code was based on K ing's [1966]finding that about 10 percent of the variance in returns could be explained by anindustry factor. However. Fertuck [1975, p. 847] found this to be true only for certainindustries: "In some industries, the industry effect is trivial and can be safely ign^ored.In others, it can be as large as a third of the market effect." The implication of thisstatement is that the industry effect exist s in some indu stries, but it is not as imjiortantas the market effect. Based upon this evidence, the industry classification was ignoredin favor of the general ma rket effect when there was a conflict between the m atching offirms on the SIC code and the matching of firms on /3j." Gonedes [1975] argues that the distributions of returns on different types of assetsmay be affected differently by the same source of information.

'" The classification of firms into two risk lev els w as a simplification; how ever, it w asbelieved that the tests with the two risk levels would be sufficient to detect any risk-dependence of lease disclosure effects. It was a lso assumed that further classification offirms by risk leve ls would not be desirable because of the relatively sm all sam ple size." It was asumed that monthly average returns were independent over time, implyingthat returns (R" and.R'') were identically distributed in every month. This independ-ence assumption may be questioned, since the behavior of returns in different months(for example, March and September) may not be the same. (For empirical evidence ondifferent behavior of returns in different months, see Beaver [1968. p. 41].) However,


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322 BYUNG T. RO

The distribution of return differences computed above, as well as therelated raw rates of return, was assumed to be multivariate normal.TESTING HYPOTHESIS AND EVALUATION METHOD

Based upon the matched pair design, along with the normal i tyas sumpt ion , the main concern was w h e t h e r the mean va lue of r e t u r ndifferences was equa l to zero. Thus the nul l hypothesis was set up inthe following form:

where fxa is the population mean of return differences.Given the null hypothesis, one may use a univariate test method to

see if each of the component (high and low) mean return differences isequal to zero. However, one problem with a univariate approach isthat the joint level of significance (or confidence interval) which isdesired cannot be determined exactly by the repeated use of a univar-iate test for each component. In order to ensure a desired joint level ofsignificance, one needs a multivariate test method which simultane-ously examines the equality of all the component mean differences tozero. ^ Moreover, a multivariate approach is consistent with themultivariate normality assumption which was introduced earlier.

The null hypothesis was tested by using Hotelling's [1931] T^statistic,^^ the form of which is given by:

(3)| / | S J ) - 1

where d = the sample mean of return differences.t(W) = W^d/V(W^SdW)/iV which is the standard f-test

statistic incorporating a weight (column) vector= the 2x 2 covariance matrix defined as


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CAPITALIZED LEASE INFORMATION 3 2 3Number of

Group Subgroups ^ Level Matched Pairs3 030Group 220

L 19F I G . 2. Grouping of firms.

" I I " = '


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3 2 4 BYUNG T, ROWith the groups formed, an overall test was first conducted combin-ing the PV and PV-IE groups into a single analysis. The ovei*all nullhypothesis for this test can be expressed in terms of a four-element

mean return difference vector, the elements consisting of PV-H,PV-L, PVIE-H, and PVIE-L, with a related 4 x 4 covariance matrix.However, the res ults of the overall test would not reveal w hat effect(if any) the disclosure of the PV numbers alone had on the pricing ofsecurities and whether there w as any join t effect of disclosing bo thPVand IE numbers at the same time. In addition, the result would notshow which component (high or low) of the mean return vector for thePV group (or the PV-IE group) primarily contributed to a rejection ofthe overall null hypothesis.In order to address these issues, separate tests by subgroups and byrisk levels were undertaken as a second step of the analysis. For thepurpose of examining any risk-dependence of lease disclosure effect,three different values of the weight vector were employed: the weights(W,) implicit in com puting th e value of T^, W2 = [1 0]^ for the high-risk group, and W3 = [0 1]^ for the low-risk group. The weight vector

W2 was employed to determ ine if the high-risk mean r etu rn differencewas significantly different from zero, while the weight vector W3 wasused to investigate if the low-risk mean return difference was signifi-cant , ^TEST PERIODS

The following events took place during the course of the SEC leasedisclosure decision: (1) the publication of the ASR No. 147 proposal;(2) the adoption of the proposal; (3) the effective date of the Release;and (4) the first public disclosure of the actual PV and IE num bers byindividual firms according to the Release.A twenty-one-month test period (1/73-9/74) was adopted to investi-gate possible market reactions to these events. This period runs fromsix months before the first event (June 1973) to six months after thelast event (March 1974), ^ In an effort to identify more precisely when^ According to Gonedes [1975], a value of the weight vector can be viewed as theproportions of investments in different types of assets. For exam ple, the w e i ^ t vectorW2 implies the hypothetical situation in w hich one places his entire investm ent in thehigh-risk securities. Likewise, W3 implies an entire investment in the low-risk securi-ties." The data collection showed that about 95 percent of the sample firm s filed their 10-


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CAPITALIZED LEASE INFO RMA TION 325the most intensive market reaction took place, tbe twenty-one-montbperiod was divided into five subperiods on tbe basis of tbe criticalevents. H owever, tbe effective date (November 30, 1973) of ASR No.147 was deleted from consideration on tbe grounds that there mightbe little to differentiate it from tbe adoption date, in terms of poten tialsignificance to investors. Botb tbe ev ents and tb e selected test periodsare summarized in figure 3. All statistical tests, including tbe overalltest, were conducted for eacb of the six different test p eriods.3. Data CollectionDISCLOSURE FIRMS

Tbe disclosure firms (group 2) did report tbe capitalized data ofnoncapitalized financing leases in tbe ir 10-K rep orts according to ASi?N o. 147 for 1973 and 1974. In addition, tbey met the following selectioncriteria. (1) Firms must be registered on tbe New York Stock Ex-change (NYSE). (2) No capitalized lease da ta were reported on the 10-K or tbe an nua l repo rts prior to ASi? N o . 147 (3) Fiscal yea r m ust endDecember 31. Tbese crite ria were checked by reading tbe 10-K andthe annua l reports for the thre e y ea rs, 1972 throu gh 1974.Table 1 provides the statistics of tbe PV and IE ratios as defined inthe Release. Several interesting points appear. First, tbe mean of tbePV ratios {PV/L-TC) in each year is much greater for tbe PV-IE firmsthan for the PV firms. Note tbat tbe former firms also had lsirger IEratios (IE/AN I) tban tbe latter firms, wbose IE numbers were notdisclosed because of the immateriality of such numbers. The greaterPV and IE ratios imply a greater potential impact on financial ratios,suggesting that the SEC decision might have affected tbe securityprices of the PV-IE firms more tban tbe security prices of the PVfirms.Second, tbe mean and tbe standard deviation of tbe PV ratios arefar greater for tbe low-risk group of tbe PV-IE firms than for tbecorresponding high-risk group, altbougb tbis is not true for tbe PV

A n c o u n c e o e n to f the p r o p o w l1773

A d o p t i o n oft h e p r o p o s a l E f f e c t i v ed a t e FirstdisclosureJ u n 1 5 73 O c t 1973 U o v 1973 M a r 197' 3 e p | 1 9 7 l t


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326 BYUNG T. RO


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CAPITALIZED LEASE INFORMATION 3 2 7firms. This would indicate that the low-risk PV-IE firms' financialratios were affected to a greater extent by the capitalized leasedisclosure than the same financial ratios of tbe high-risk PV-IE firmswere. As a resu lt, th ere is the possibility th a t the security prices of thelow-risk PV-IE firms were more affected by the SEC decision thanwere the security prices of the high-risk PV-IE firms, if tbe effect ofthe PV disclosure was significant and dominant over the effect of/disclosure.Finally, the directions of changes in the two ratios, PV andlE, arehighly consistent. Except for one firm, all bad positive changes in tbePV ratios, while m ost of the PV-IE firms revealed negative changes inthe IE ratios. It should be noted, however, that such consistentdirectional changes in these ratios might not necessarily have causedsimilar directional changes in the behavior of security returns.^*NONLEASE FIRMS

The nonlease firms (group 1) were those which had no noncancelableleases for 1973 and 1974, and which m et tbe same three c riter ia usedin selecting the disclosure firms. Information on these firms wasobtained from footnotes on leases and sucb sections as "Summary ofSignificant Accounting Policies" and "Supplementary Income State-ment Information" in the 10-K and annual reports of 1972 through1974.As indicated ear lier, each firm in group 1 was paired w ith everyfirm in group 2 on tbe basis of )3's. Table 2 gives the statistics thatwere calculated using the individual /3's provided in Appendix B. Notethat between tbe two groups, tbe mean values (1.366 and 1.328) of )8'sare fairly equal. The variances (.105 and .092) associated with thesemean betas are relatively sm all.Tbe approximate equality of tbe means (and variances) of j8'sbetween tbe two groups was maintained even when each of the twogroups was divided into two subgroups by the type of lease disclosureand then into the two risk groups. Note also that tbe PV-IE firmstended to have higher mean and variance of ^'s relative to the PVfirms.

The results of the matching suggest that the purpose of matchingfirms was reasonably achieved, in the sense that the treatment and


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3 2 8 BYUNG T. ROT A B L E 2Statistics of Beta Estimates Used for M atching

Overall Group 2 Group 1Group 2 Group 1 PV PV-IE PV PV-IE

M e a n 1^366 1^328 1^09 1^4153 1^276 1^409(H)* ( 1 . 535 ) ( 1 . 755 ) ( 1 . 4 77 ) ( 1 . 6 7 8 )(L ) (1.082) (1.135) (1.075) (1.126)Var iance 105 .092 .077 .139 .059 .134iH ) ( .035) ( .049) ( .022) ( .083)(L) ( .016) ( .034) ( .015) ( .029)Number of/3 99 99 60 39 60 39(H ) (30) (20) (30) (20)(L ) (30) (19) (30) (19)* H = High risk, an dL = Low risk.

control firms were in the homogeneous risk class and had the samefirst mom en ts of return distribution s durin g the pretest (beta estima-tion) period. In an effort to evaluate this fact, a separate statisticaltest was conducted for the period of on e yea r (1972) selected as pa rt ofthe pretest period. The results from this test suggest th at the re was n ostatistically significant difference in mean return vectors between thetwo groups offirms.^SECURITY RETURN DATA

M onthly return s, used as a stock price variable, were the continu-ously compounded rate s of retu rn computed as:(5)

where P an d D denote stock price and dividend, respectively. All rawreturn data (before continuously compounded by means of equation (5)above) were obtained from a 1975 edition of the CRSP tapes.4. Analysis of Test Results

According to the test results discussed in detail below, it appears thatthe lease information effects, measured by changes in the conditionalexpected returns for the lease disclosure firms, did exist. However,evidence for the information effects was present only for the PV-IEdisclosure firms, and n ot for the firms w hich disclosed the PV n umbers


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CAPITALIZED LEASE INFORMATION 3 2 9RESULTS OF OVERALL TESTS

The T" values in table 3 were calculated using equation (3), while th eF-values were computed according to equation (4), The computed F-values for all test periods, except for TP4, were significant at the ,10level. Therefore, th e nu ll hypothesis of no re tu rn difference between thelease disclosure firms and th e nonlease firms can be rejected a t the levelof significance for each of those five test periods, including the entiretwenty-one-month period.The fact th a t th eF -value (2,020) for TP4 turned out to be insignificantis somewhat surprising. It may be that the lack of significance for thisperiod was simply attrib uta ble to a samp ling variability which could beexpected under tiie null hypothesis with a chance of 10 out of 100, Theinterpre tation would be reasonab le, since the period (TP4) covered theevent of the first public disclosure of the PV and IE numbers underASR No. 147, the event which was presumably most important for thelease disclosure. Moreover, the test result for TP6, which was not onlya subperiod of TP4, but also covered the same disclosure event, turnedout to be significant. Despite the ambiguous implication of the testresult for TP4, the overall test results strongly suggest that there wassignificant difference in return behavior between the firms whichdisclosed the ir capitalized lease information as required by the SEC andthe firms which did not have leases to be capitalized.As indicated earlier, however, the results of the overall tests do notreveal which group (PV or PV -IE) and/or which risk class (high or low)primarily contributed to the rejection of the overall null hypothesis.Some insights into th is issue a re obtained from the resu lts of additionaltests described below.

TABLE 3Results of Overall T estsMonths

TPl: 21TP2: 16TP3: 12TP4: 9TP5: 14TP6: 7

r ' Value11.38412.83617.34712.92816,16947.992

Computed F-V alu e'2.419*2.567*3.154*2.0203,109*5.999** Significant at the . 10 level.


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330 BYUNG T. ROT A B L E 4

Test Results for PV Firms

MoTPl:

TP2:

TP3:

TP4:

TP5:

TP6:

mths212121161616121212999

141414777

Average Return DifferenceHigh

- . 0 0 5 9(.0159)'-

- . 0 0 5 8(.0155)

- . 0 1 0 2(.0133)

- . 0 0 9 5(.0131)

- . 0 0 2 9(.0164)

- . 0 1 1 7(.0142)

Low- . 0 0 3 5(.0222)''

- . 0 0 5 4(.0246)

- . 0 0 4 7(.0240)

- . 0 0 5 4(.0213)

- . 0 0 4 1(.0238)

- . 0 0 2 2(.0204)

7-2

3.3572.847

.5153.0502.248

.7657.0237.019

.4665.6634.829

.572

.852

.449

.4224.8484.753

.080

ComputedF-Value"

1.5951.352

.2451.4231.049

.3573.192*3.191*

.2122.4772.113

.250

.393

.207

.1952.0201.981

.033

WeightsHigh

.76710

.72810

1.01310

.80110

.60010.91110

Low.233

01

.27201

- . 0 1 301

.19901

.40001.08901

* Significant at the .10 level." Degrees offreedomfor theF-values are (2, 19) forTPl, (2, 14) for TP2, (2, 10) for TP3, (2, 7) forTP4, (2, 12) for TP5, and (2, 5) for TP6." The first line of the weight vector column in each test period stands for the implicit weights(implicit incomputing the value ofT^)." Standard deviation.Selected fractiles of the /"-distribution and the exact F-values are:

d.f. Fractiles F-Value d.f Fractiles F-Value(2 , 19)(2 , 14)(2 , 12)

.900.950.900.950.900.950

2.613.522.733.742.813.88

(2 , 10)(2 , 7)(2 , 5)

.900.950.900.950.900.950

2.924.103.264.743.785.79TEST RESULTS FOR PV DISCLOSURE FIRMS

For the PV firms, tbe means of return differences {RT" Re''example), computed over tbe number of montbs in eacb of tbe testperiods, are shown in columns 2 and 3 of table 4 by the risk classes.


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CAPITALIZED LEASE INFORMATION 331A glance at the F-value column reveals that the only test period

yielding the significant F-value (3.192) at the .10 level is the twelve-month period {TP3), October 1973 (the month in which the SEC decisionwas formally announced) through September 1974 (six months after thefirst public disclosure of the PV and IE numbers). Moreover, thesignificant F-value for the period is shown to be primarily attributableto the high-risk firms in group 2 (see the observed F-value (3.191)).

However, the failure to reject the null hypothesis for the other testperiods, including the entire twenty-one-month period, casts doubt onthe implication of the test result for the twelve-month period. Since thenull hypothesis for the overall period was not rejected, it seemsreasonable to assume that the result for the twelve-month period turnedout to be significant by chance. This assumption is consistent with theimplication of figure 4, wherein none of the CARD curves tends todeviate substantially from the horizontal zero return line.^'

To summarize, evidence from the test results for the PV disclosurefirms suggests that the disclosure of capitalized lease information asrequired by the SEC had no significant effect upon the pricing ofsecurities. This implies that the PV numbers of noncapitalized financ-ing leases, as disclosed under ASR No. 147, did not carry newinformation to investors for assessing the risk-return attributes of firmswith such leases. There exists evidence (especially the average returndifferences for TP3 and TP6) for some degree of market reaction to theannouncement of the SEC lease decision and the disclosure of the PVnumbers. However, the degree of the market reaction was not sufficientto lead to the conclusion that the SEC decision affected the pricing ofsecurities significantly.

The above finding appears to imply that the balance sheet effect oflease capitalization (including the effect of changes in leverage ratios)values of th e T^ and F -statistic. All values of T^ and F associated with other weightvectors should be less than or equal to the values of T^ and F related to the implicitweight vector. Notice that this is true for all test periods, as seen in table 4." The overall CARD curve was constructed using eq. (4-2) and th e CARD curves bydifferent risks groups using eq. (4-4) below:

14-1)


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332 BYUNG T. ROT A B L E 5Test Results for PV-IE Firms

M rIVIU

TPl:

TP2:

TP3:

TP4:

TP5:

TP6:

inths212121161616121212999

141414777

Average Return DifTerenceHigh

- . 0 1 8 0( .0329)

- . 0 1 8 6( .0294)

- . 0 1 9 4( .0272)

- . 0 1 7 7( .0305)

- . 0 1 3 0( .0362)

- . 0 2 8 0( .0244)

Low- . 0 0 6 5( .0316)

.0011( .0302)

.0094( .0265)

.0142( .0285)

- . 0 1 7 8( .0251)

.0161( .0326)

6.378

6.268.876

7.0816.445

.0218.1926.1111.5315.1733.0482.2277.074

1.7996.992

11.7689.1951.714

ComputedF-Value3.030*2.978*

.4163.304*3.007*

.0103.724*2.778

.6962.2631.334

.9753.265*

.8303.227*4.903*3.832*

.714

WeightsHigh.872

101.412

102.356

1010.704

10- . 0 9 5

101.61010

Low.12801

- . 4 1 201

- 1 . 3 5 601

- 9 . 7 0 401

1.09501- . 6 1 001

* Significant at the .10 level.

on security prices may not be as great as has generally been believed.Alternatively, it seems that information conveyed by the PV numbersreleased under ASR No. 147 might already have been impounded insecurity prices through nonaccounting sources, even prior to the SECextended lease disclosure decision of 1973. This explanation is probablyreasonable, provided that lease capitalization generally affects thefirms' leverage ratios. It also appears to be consistent with whatHamada [1972] proved conceptually and empirically. He proved thatthere is a relationship between the firms' leverage profiles and theirrelative risks.TEST RESULTS FOR PV AND IE DISCLOSURE FIRMS


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CAPITALIZED LEASE INFORMATION 333CARD

I 1 I I r I

A l l

I IJan J u n O ct Mar1973 19 73 1973 197'* Sept

CARD

-High

I I I I t

lowHigh

I IJan J u n O c t Kar Sept197 3 1 9 7 3 1 9 7 3 197'*- 197'*

FIG. 4.Cumulative average return difference: PV firms. All = all firms; Highhigh-risk firms; Low = low-risk firms.figure 5. The absolute values of average return differences for the PV-IE firms tend to be much greater than those for the PV firms. Inaddition, the above finding is also consistent with the implications ofthe PV and IE ratios (see table 1). The mean values of both PV and IEratios for the PV-IE firms were shown to be far greater than the meanvalues of the sam e ratios for the PV firms.^*


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3 3 4 BYUNG T, ROC A R D

AU

L LJ a n J u n O c t M ar S e p t1 9 7 3 1 9 7 3 1 9 7 3 1 9 7 ^ 1 9 7 *

CARD

M 1 1 1 'J a n)an Jun Ocz1973 19 73 197 3

FIG. 5. Cumulative average return difference: PV and/ firms.

decision was initiated about thre e m onths p rior to the p ublication of theASR No. 147 proposal. Moreover, there was a further market reactionup on the disclosure of the actual P V and IE numbers .The above findings, coupled with the results in table 4, suggest thatthe joint effect of both PV and IE disclosure upon security prices wassignificant, and m uch m ore so than th e singular effect of PV disclosurealone. Evidence from table 5 also indicates that the extent to whichsecurity prices were affected by the SEC decision was not the same forthe firms in different risk classe s. The observed F -value s of the high-risk group associated with the weight vector [1 0] for a ll test periods,


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CAPITALIZED LEASE INFORMATION 3 3 5risk group shows a downward deviation from the horizontal zero lineover almost the entire twenty-one-montb period, while the CARD curvefor tbe low-risk group starts down but tben turns up in tbe second balfof tbe period. Tbis caused a reversal of tbe two curves, higb and low,around March 1974. The interesting behavior shows up also in tbelarger magnitude of average return difference (- .0178) for tbe low-riskgroup for TP5 (the fourteen-month period), and in tbe significant F-value (3.227) associated witb tbe weigbt vector [0 1]. However, itappears tha t the relative ly larger dev iation of tbe low-risk CARD curvefrom the zero return line during tbe fourteen-month period and itssubsequent upward cbange around Marcb 1974 may merely be causedby a sampling variability, since, by definition, tbe bigb-risk firms mustbe riskier tban tbe low-risk firms. Therefore, it is not likely tbat tbelatter firms were, in fact, more adversely affected by tbe capitalizedlease disclosure tban the former firms.^^

To summarize, tbe test results for the PV-IE firms suggest tbat theinformation effects of the various events of tbe SEC lease decision werepresent when the effects were measured by tbe changes in th e expectedvalues of return distributions. The results also reveal tbat tbe marketreaction began as early as Marcb 1973, indicating th at tbe effects of tbecapitalized lease disclosure (as required by tbe SEC) were anticipated

'' An alternative explanation may be possible-that the security prices of the low-risk PV-IE firms, relative to those of the high-risk PV-IE firms, were in fact moreadversely iffected by the predisclosure events of the SEC lease decision. They, then,experienced substan tial upward adjustments upon the disclosure of the capitalized leaseinformation. First, the possible reason for the larger adverse effect is that investorsmight have been concerned primarily with the impact of lease capitalization on thefirms' capital structures (or on the balance-sheet-based financial ratios, such as debt-equity ratio), rather than with the impact on the firms' operating results as reflected inthe eaiTiings-per-share figures. Moreover, such an impact was anticipated by investorseven prior to the release of thePV numbers which m ight have been used by investors assurrogates for the measures of capital structures. In this case, it is likely that thesecurity prices of the firms with higher PV ratios could be more affected by thepredisclosure events of the SEC decision than the security prices of the firms with lowerPV ratios. As seen in table 1, the m ean values of theP V ratios were higher for the low-risk PV-/E firms (.264 for 1973 and .282 for 1974) than for the high-risk PV-IE firms(.203 for 1973 and .210 for 1974).Second, the reason for the upward revision of investor's expectations about the low-risk PV-/E firms may be tha t, when the PV and IE numbers (especially the IE numbers)


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3 3 6 BYUNG T. ROby the investment community prior to the disclosure of the capitalizedlease numbers through accounting sources.The re sults also show th a t the capitalized lease disclosure did, overall,have an adverse effect on the valuation of the firms, although a slightupward readjustment of the security prices upon the disclosure of thecapitalized lease data was observed for the low-risk firms. This upwardreadjustment may indicate that the negative effects of the PV and IEnumbers (especially the latter) on various financial ratios were not asbad as investors had originally anticipated. This evidence of adverseeffects on the valuation of the firms is consistent with the result of astudy by Nelson [1963], who found that lease capitalization adverselyaffected the financial ratios examined in his study.

There is also strong evidence of risk-dependence of the lease disclo-sure effects. In general, the h igh -riskPV -/E firms were more adverselyevaluated by investors, given the capitalized lease disclosure by theSEC, than were the low-risk PV-IE firms. The existence of such riskdependence is inconsistent with the situation found by Gonedes [1975],where no such dependence was shown to exist between the relative riskof firms and th e effect of special items th a t he investiga ted.5. ConclusionsGiven the empirical findings of this study, it seems reasonable todraw the following conclusions. (1) The capitalized lease disclosurealong with income effect disclosure, required by the SEC, did signifi-cantly change the distribution of security returns. Hence, the SEC'sextended lease disclosure decision of 1973 had an impact on the pricingof securities for the firms which were affected by the decision. (2) Theobserved price effect was less significant for the firms which made onlyPV numbers public, without disclosure of IE numbers as well. Thisseems to imply that the balance sheet effect itself of capitalized leasedisclosure on the pricing of securities may no t be as grea t as is generallybelieved. Alternatively, it appears that the balance sheet effect (if any)was already impounded in security prices prior to the SEC leasedecision, possibly a t or around the tim e w hen the con tracts of the leasescapitalized under ASR No. 147 were initially established. (3) Theinformation effects of the various events of the SEC decision tended tobe risk-dependent, in the sense tha t the exten t of the m arket reaction tothose events varied according to the firms' riskiness. In general, thesecurity prices of the high-risk firms were more affected by the SEC


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CAPITALIZED LEASE INFORMATION 3 3 7

The above conclusions must be qualified because of the sm all samplesize, the lack of a test for the homogeneity of control group firms, andthe sample firm selection criteria (NYSE firms which have the samefiscal year ending December 31). These potential sources of limitationsopen areas for future research on tbe subject. In addition, futureresearch m ight consider grouping firms on the basis of the ran ks of thePV ratios and/or the IE ratios . Then, some sort of statistical significancetests could be conducted. Also, the research design of the present studycould be refined through a better matching of tbe treatment and controlfirms based on the standard errors of individual beta estimates andresidual re turn s, in addition to beta estimates and the SIC code.

APPENDIX ASample Firms Matched by Industries

12

4

SIC Code

& 327282933

35

39

281324331343366367371394

IndustryMining and constructionManufacturingPrinting, publishing, and alliedindustriesChemicals and allied productsIndustrial inorganic chemicalsPetroleum refining and related industriesCement and hydraulicPrimary metal industriesBlast furnaces, steel works, and

rolling and finishing millsHeating equipmentMachinery except electricalCommunication equipmentElectronic components and accessoriesMotor vehicles and equipmentMiscellaneous manufacturing industriesToys and am usement, sporting andath letic goodsTransportation and public utilitiesRailroads

No. ofPairs1

19111111

1141121

121


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3 3 8 BYUNG T. ROAPPENDIX BIndividual Beta Estimates Used in Matching FirmsFirm No. Group 2 Group 1 Firm No. Group 2 Group 1

12345678910111213141516171819202122232425262728293031323334353637383940

1.401.411.481.78.83.871.761.061.481.391.401.121.041.292.061.401.861.111.031.221.011.841.161.151.401.211.651.031.031.461.281.651.221.791.45.931.371.331.03

1.401.361.461.65.83.871.621.071.471.371.411.081.041.301.921.341.741.101.031.231.001.711.151.131.361.201.581.021.041.431.251.481.201.681.42.93

1.301.311.04

535455565758596061626364656667686970717273747576777879808182838485868788899091

1.94.941.281.281.781.662.111.121.161.421.761.391.141.261.211.661.37.761.501.171.361.031.24.95.99.871.711.921.262.251.761.641.941.531.131.301.241.061.65

1.88.941.271.251.651.581.981.111.131.421.641.331.131.241.201.511.32.751.461.161.311.041.20.961.00.871.591.801.232 . 6 71.651.521.861.481.131.301.241.071.51


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CAPITALIZED LEASE INFORMATION 3 3 946 1.06 1,04 98 1,74 1,5047 1,16 1.16 99 1.75 1.6648 .91 ,9249 1,80 1,69 M e a n of ^ = 1,366 1.32850 1,62 1,5051 .99 .99 V ar ia nc e of ;8 = ,105 .09252 ^98 ^ 8

REFERENCESAxELSON, K. S. "Needed: A Generally Accepted Method for Measuring Lease Commit-ments." Financial Executive (July 1971): 40-52.Beaver, W. H. "The Information Content of Annual Earnings Announcements."Em pirical Research in Accoun ting: Selected Studies, 1968. Supplement to Journal ofAccounting Research 6: 67-92., P . KETTLER, AND M . SCHOLES. "The Association between Market Determinedand Accounting Determined Risk Measures." The Accounting Review (October 1970):654-82.BoLCH, B. W., AND C. J. H U A N G . Mu ltivariate Statistical Metho ds for Business andEconomics. Englewood Cliffs, N. J.: Prentice-Hall, 1974.BUFF, J. H. "Reporting of Leases in Finan cial S tate m en ts - Th e Investor's Viewpo int."Retail Control (Ju ne /Ju ly 1971): 19-24.COOK, D. C . "The Case against Capitalizing Leases. ' Harvard Business Review (Janu-ary /Fe bru ary 1963): 145-62.DEFLIESE, P . L . Should Accountants Capitalize Leases? The Economic Case against It atPresent. New York: Coopers & Lybrand, 1973.DONALDSON, G. "New Framework for Corporate Debt Policy." Harva rd Business Review(March/April 1962): 117-31.F AM A, E . F . , AND M . H . M I L L E R . The Theory of Finance. Hinsdale, 111.: Dryden Press,

1972.FERTUCK, L. "A Test of Industry Indices Based on SIC Codes.' Journal of Financial andQuantitative Analysis (Decem ber 1975): 837-48.GONEDES, N . J. "Risk, Inform ation, and the Effects of Special Item s on Ca pital M ark etEquilibrium.' Journal o f Accoun ting Research (A utu m n 1975): 220-56., AND N. D O P U C H . "Capital Market Equilibrium, Information Production, andSelecting Accounting Techniques: Theoretical Framework and Review of EmpiricalWorks." Studies o n Financial Accoun ting Objectives. Supplement to Journal ofAccounting Research 12 (1974): 48-129., AND S. H. P E N M A N . "Disclosure Rules, Information-Production, and CapitalMarket Equilibrium: The Case of Forecast Disclosure Rules. ' Journal of AccountingResearch (Spring 1976): 89-137.HAMADA, R . S . "The Effect of the Firm's Capital Structure on the Systematic Risk of

Common Stock.' Journal of Finance (May 1972): 435-52 .HARRIS , R . J . A Primer of Multivariate Statistics. New York: Academic Press, 1975.


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3 4 0 BYUNG T. ROMORRISON, D. F. Multivariate Statistical Methods. New York: McGraw-Hill, 1976.MYER S, J. H. Accounting Research Study No. 4: Reporting of Leases in F inancialStatements. New Y ork: AICPA , 1962.N E L S O N , A. T. "Capitalizing Leases-The Effect on Financial Ratios." Journal of

Accountancy (July 1963): 49-58.NuRNBERG, H. "Leases, Purchase Commitments, and Pensions Revisited." CPA Journal(May 1973): 375-89.Ro, B. T. "The Disclosure of Capitalized Lease Information and Stock Prices." Ph.Ddissertation, Michigan State University, 1976.SUNDER , S . "Relationship between Accounting Changes and Stock Prices: Problems ofMeasurement and Some Empirical Evidence." Empirical Research in Accounting:Selected Studies, 1973. Supplement to Journal of Accounting Research 11:1-45.VATTER , W. J . "Accou nting for Lea ses." JournaZ of Accounting Research (Autumn 1966):133-48.ZiSES, A. "Disclosure of Long-Term Leases.' Journal of Accountancy (February 1961):37-47.. "The Pseud o-Leas es-Trap and Time Bomh." Financial Executive (August 1973):20-25.


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disclosure of capital lease and stock prices

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