International Review of Economics and Finance 9 (2000) 139–156
Self-selection and the effects of poison put/call
covenants on the reoffering yields of corporate bonds
Khalil M. Torabzadeh
a,*, John Roufagalas
b, Criss G. Woodruff
c aFaculty of Management, The University of Lethbridge, Lethbridge, AB T1K 3M4, Canada bDepartment of Economics, College of Business and Economics, Radford University, Radford,VA 24142, USA
cCollege of Business, Texas A&M University-Corpus Christi, Corpus Christi, TX 78412, USA
Received 25 September 1998; accepted 7 April 1999
Abstract
This study examines the effects of poison put/call covenants on the reoffering yields of corporate bonds issued during 1986 through 1990. The analysis controls for possible biases due to the endogeneity of the decision to issue the bond and the decision to include a poison covenant. The findings indicate that the presence of a single poison call provision has a positive effect on the yield differential, increasing the yield by up to 54 basis points. This is consistent with the managerial entrenchment hypothesis. On the other hand, the inclusion of a poison put provision reduces the yield by 58 to 78 basis points. Furthermore, we find that the imposition of an additional stringent provision to transfer a simple poison put to a super poison put does not significantly decrease the yields further. 2000 Elsevier Science Inc. All rights reserved.
JEL classification:G32
Keywords:Poison put/call covenants; Debt financing; Bonds; Reoffering yield
1. Introduction
The explosive growth in leveraged buyouts and other leverage-increasing takeovers during the second half of the 1980s resulted in substantial wealth losses to bondholders (Asquith & Wizman, 1990; Crabbe, 1991; Warga & Welch, 1993). The main source of these losses was the increase in risk associated with the large amount of debt needed to finance the transaction. The RJR Nabisco leveraged buyout made these wealth
* Corresponding author. Tel.: 403-329-2249; fax: 403-329-2038.
E-mail address: [email protected] (K.M. Torabzadeh)
losses painfully obvious to all market participants. Consequently, bondholders started requesting protection against takeover-related events that would damage the credit quality of the firm. The protection demanded by bondholders, in turn, provided opportunities for incumbent managers to launch new defensive mechanisms in an attempt to deter unwanted takeovers.1 Hence, investors were offered bonds with a
variety of protective covenants.
One category of covenants designed to protect bondholders against takeover-related losses includes the so-called poison call and poison put covenants. These covenants, similar to poison pills, entitle their holders to special rights if an unanticipated event occurs. A poison put provision gives bondholders the right to require the firm to redeem the bond at par (sometimes at premium) upon the occurrence of a designated event or a sequence of events that would likely impair bond values. A poison call covenant, on the other hand, gives the company the right to call the bond if a risk event occurs. Some bond indentures contain both poison put and call provisions. The designated events by which a call or put can be triggered are generally takeover related. In fact, almost every poison put/call that was issued prior to the 1988 RJR buyout had a “change of control” as its designated event.2The RJR buyout and the
subsequent downgrading of its bonds intensified the tension in the corporate bond market and prompted bondholders to demand stronger and wider coverage. Bond issuers responded by offering stronger covenants called super poison puts. A super poison put allows bondholders to redeem their bonds (or reset the coupon to restore market value to par) when a change in control happens and/or the bonds are down-graded by either Standard and Poor’s (S&P) or Moody’s, or both. An example for each type of poison covenant is presented in the Appendix.
The specifics of each covenant differ. Some super poison puts require downgrading from investment to speculative grade for the puts to be triggered, while others are triggered by downgrading within investment grades. Some provisions contain other risk events in addition to change of control. These differences led the S&P to introduce a five-scale event-risk rating scheme in July 1989. The scale ranges from E-1 (strong protection) to E-5 (insignificant or no protection).3 Moody’s Investor Service, on
the other hand, directly incorporates event-risk in its overall assessment of bond creditworthiness by assigning lower ratings to bonds with weaker protection.
levels of covenant protection. They observe that bonds with strong protection gain, whereas those with weak or no protection lose value.
In addition to small sample size, Crabbe’s (1991) use of two categorical dummy variables (firms with or without super poison puts) produces a potential bias. The source of this bias is the self-selection problem. His model yields consistent estimates only when the decision to include a covenant is exogenous or independent of the decision to issue a bond. Given bondholders’ demands for stronger protection in the latter part of the 1980s, the condition of exogeneity was violated. In this peculiar period, the inclusion of a poison covenant in the bond indenture became a major deciding factor affecting corporate financing strategy. The decision was particularly important for those corporate borrowers that were viewed as possible takeover targets.4
They had to decide whether to stay in the bond market and possibly meet bondholders’ demands, or to pursue an alternative form of financing. Those firms that decided to stay in the bond market and at the same time decided not to include poison covenants and, as a result, were forced to offer bonds at higher yields, formed a distinct category in the universe of bond issuers. Combining this group of firms with those firms for which investors demanded no poison covenants and comparing their yields with those of issues with poison covenants, as in Crabbe (1991), yields biased estimates. Lehn and Poulsen (1991), commenting on Crabbe’s results, identify this bias as they state, “unfortunately, these are conservative estimates since the presence of event-risk cove-nants is determined endogenously (that is, firms that are not likely to receive leveraged buyout offers are less likely to issue debt with event-risk covenants than other firms).” A consistent estimate of the effect of the poison covenants on the reoffering yields should compare bonds with and without covenants issued by firms with similar proba-bilities of takeover. Econometric methods to deal with the self-selection bias are described in Maddala (1983) and Greene (1993) and have been applied, among others, by Smith (1987) in his study of the choice of issuance procedures.
This study extends that of Crabbe’s (1991) in a number of ways. First, this analysis uses a much larger sample than previous analyses. The sample includes most of the industrial bonds with maturity of no less than seven years issued during 1986 through 1990. Second, the study examines the effects of all categories of poison covenants (i.e., simple poison call, simple poison put, and super poison put). Third, because of the takeover defensive properties of poison put/call provisions, the study addresses the issue in conjunction with managerial entrenchment hypothesis and its implication for bondholders. Finally, this analysis controls for possible self-selection bias to provide consistent estimates of the effects of the poison covenants on the reoffering yields.
The results of the study show that the inclusion of a simple poison call has positive effect upon reoffering yields, increasing the yield by up to 54 basis points. This is consistent with the managerial entrenchment motives. On the other hand, the inclusion of a poison put reduces reoffering yields by 58 to 78 basis points. Interestingly, the imposition of an additional stringent provision to transfer a simple poison put to a super poison put does not significantly decrease the yields further.
on the bond prices. Section 3 defines the sample, data, and methodology. Section 4 presents and discusses the results, while Section 5 concludes the paper.
2. Hypotheses
The decision to finance via a bond issue and the type of covenants to be included involves at least three parties: shareholders, management, and prospective bondhold-ers. Various hypotheses have been advanced as to how the relations among these groups affect the capital structure of the firm. These hypotheses are mainly based on the presence of asymmetric information. Asymmetry of information between shareholders and management gives rise to a principal-agent problem that provides the foundation for the managerial entrenchment hypothesis. In this context, Schleifer and Vishny (1989) argue that managers counter disciplinary forces (i.e., the board of directors, labor monitoring, takeover threats) by making themselves valuable and costly to replace. One of the choices available to them is to sign a debt contract that includes a covenant requiring full repayment if the firm is acquired.5
Harris and Raviv (1988), Stulz (1988), Israel (1991), and Hendershott (1991) focus on the managerial entrenchment behavior in connection with the defensive takeover role of financial leverage. Although the analyses differ, all agree that taking on addi-tional debt reduces the probability of takeover. Harris and Raviv (1988) and Stulz (1988) argue that this happens because managers use the proceeds from new debt to repurchase outstanding shares. As a result, the proportion of managerial equity ownership increases, causing the probability of a (hostile) takeover to decrease. Israel (1991) maintains that a takeover by more efficient management increases the value of the firm. The level of debt affects the distribution of this gain among various parties including the target’s shareholders and debtholders and the acquirer’s shareholders. It is shown that the level of debt inversely affects the portion of the gain accumulating to the acquirer. Thus, the probability of takeover decreases as the target’s debt increases. Hendershott (1991) explains the debt-takeover relation in a signaling context. He builds his model based on the notion that an increase in debt increases the probability of bankruptcy which, in turn, reduces the welfare of management. At some level of debt, the expected benefit of control drops to the point at which an inefficient manager will prefer to lose control of the firm rather than to issue a large amount of debt. On the other hand, efficient managers are able to accumulate additional debt without increasing the probability of bankruptcy to an untenable level. This will lead efficient managers to signal their type by accumulating debt to a level not affordable to ineffi-cient managers. Subsequently, acquirers will go after low debt firms. It follows that debt and probability of takeover are negatively related.
bondholders are somewhat aligned through the protection extended to the bondhold-ers. Cook and Easterwood (1994) provide evidence for this “mutual interest hypothe-sis.” They measure the wealth effect of issuing debt with and without poison put covenants on the outstanding debt and equity. Their sample of bonds with poison put covenants consists of 63 bonds issued in the period of 1988 to 1989. They find that the presence of poison puts affects current stockholders negatively and existing bond-holders positively. They conclude that the issuance of bonds with poison put covenants protects managers from hostile takeovers and bondholders from event risk, at the expense of stockholders. Their findings, however, are in contrast to those reached by Bae et al. (1994). These authors use 83 E-rated bonds issued from 1982 to 1990 to examine the effect of poison put provisions on stockholder wealth. Similar to Cook and Easterwood, they utilize an event-study methodology but find positive stock price response to the announcement of an issuance of bonds with event-risk protection. Their regression analysis indicates that the presence of event-risk covenants increases shareholder wealth primarily by reducing the firm’s agency costs of debt. Their findings basically support the argument offered by Kahan and Klausner (1993). Although they maintain that managerial entrenchment is the primary motivation behind the issuance of poison bonds, Kahan and Klausner argue that to the extent that bondholders pay for the protection by accepting a lower interest rate, the increased value accrues to the benefit of shareholders.
In sum, event-risk covenants can be designed to entrench managers and/or provide protection to the bondholders. However, the yield effect, at the time of the issue, depends on the bondholder’s perception of the structure of the covenant. We expect to see a positive reaction in the bond market when the option to put the bond back to the firm is at the discretion of bondholders. Having this attribute, we hypothesize that bonds with super and simple poison puts are priced to yield lower return compared to other comparable bonds without such covenants.
We suggest a different scenario for bonds with poison call provisions. These cove-nants are under the control of managers and are viewed as a mechanism to force the prospective bidders to negotiate directly with them. These covenants, in return for some payoff for management, can be easily negated without bondholders’ recourse. For this, we hypothesize that bonds with poison call provisions convey negative infor-mation to the bond market and tend to increase borrowing costs to the firm.
3. Sample, data, and methodology
provided by the Securities Data Company. Other sources include S&P’s CreditWeek, Moody’s Bond Survey, and Moody’s Industrial and OTC manuals. The straight bonds are collected from the S&P Bond Guide.
Bonds are included in the sample if they satisfy the following criteria: (1) the bond has maturity of seven years or longer; (2) the bond has a fixed coupon rate; and (3) the bond is rated by both S&P and Moody’s. Excluded from the sample are asset-backed securities, deep or zero discount bonds, and bonds issued by a firm’s unit or subsidiary. Issue-specific data (i.e., offer yield, maturity, size, type) are obtained from the S&P Bond Guide and confirmed with Moody’s Bond Survey. Bonds with conflicting data points (except ratings) are excluded from the sample. Other data (i.e., insider ownership, board of directors’ composition, leverage) are collected from proxy state-ments, 10-K reports, and Moody’s Industrial or OTC manuals. The variables used are defined as:
RY: Reoffering yield differential. Calculated as the difference between the bond’s yield and the similar maturity treasury bond at the date of issue.
PRIN: Principal value of the issue.
TR: Average daily yield of ten years and longer treasury bonds on the day of the issue.
VOL: Interest rate volatility. Calculated as the standard deviation of the TR for the 40 business days before the issue.
CALL: Bond callability. A dummy variable that takes a value of one if the bond is callable; zero otherwise.
SF: Sinking fund provision. A dummy variable that takes a value of one if the bond has a sinking fund provision; zero otherwise. MAT: Maturity of the bond in years.
Aaa toCaa: Moody’s ratings. Measured as dummy variables. AAA toCCC2: S&P ratings. Measured as dummy variables.
E1 to E5: S&P event-risk ratings. Measured as dummy variables.
CVT: Convertibility. A dummy variable that takes a value of one if the bond is convertible; zero otherwise.
PCAL: Poison call. A dummy variable that takes a value of one if the bond has a poison call provision; zero otherwise.
PPUT: Poison put. A dummy variable that takes a value of one if the bond has a poison put provision; zero otherwise.
SPP: Super poison put. A dummy variable that takes a value of one if the bond has a super poison put provision; zero otherwise. LEV: The firm’s leverage before the issuance of the new bond. PDIR: Proportion of outside directors in the board.
INOWN: Proportion of insider ownership of the firm.
Table 1
Descriptive statistics for 1015 bonds issued between 1986 and 1990 Panel A:
Variables with
real values Mean values Standard deviation
RY(%) 1.3841 2.1822
Dummy variables No. of observations Proportion of total
Table 1 (Continued) Panel B:
Dummy variables No. of observations Proportion of total
BB2 38 0.0374
B1 64 0.0630
B 87 0.0857
B2 164 0.1616
CCC1 39 0.0384
CCC 22 0.0217
CCC2 6 0.0059
E1 1 0.0010
E2 3 0.0029
E3 32 0.0313
E4 10 0.0098
E5 15 0.0148
PCALL 69 0.0680
PPUT 179 0.1763
SPP 31 0.0305
not directly observable from Table 1, of the total 1015 bonds studied during the 1986 through 1990, 54.98%t received investment grade ratings by Moody’s as compared to 54.68% by S&P. In addition, 78.52% (or 797 issues) of the bonds were callable, 28.67% (or 291 issues) were convertible, while 54.19% of them (or 550 issues) contained sinking fund provisions.
In this study, a self-selection model is employed to deal with the endogeneity of the decision to include a poison covenant. The model uses a two-stage procedure. In the first stage, the covenant inclusion equation is estimated using a probit model. Probit model, in its simplest form, is an equation where the dependent variable, Ci, takes a value of one when the covenant is present and zero otherwise. This categorical variable is essentially a proxy for the benefits of including a covenant (i.e., the depen-dent variable takes a value of one when the benefits are high and the covenant is included and a value of zero when the benefits are low and the covenant is not included). The model is:
Ci5 a Xi1ei (1)
where Xi is the vector of independent variables that determine the benefits of the covenants andeiis assumed to beN(0,se2).
Table 2
Values of the dependent variableCia
Ci PCALL PPUT SPP
0 0 0 0
1 1 0 0
2 0 1 0
3 1 1 0
4 0 1 1
aAnother conceivable combination is (1,1,1) where the same covenant contains both a super poison
put and a poison call. We found a few cases satisfying this combination, but they failed to clearly meet our sample selection criteria. The remaining combinations of (0,0,1) and (1,0,1) are not possible since we cannot have a super poison put without a simple poison put.
priori reason to expect that the utility or profitability ranking of these alternatives would be identical across the firms (i.e., there would be no single ordering). Second, an examination of these three types of covenants reveals that they are not independent. For example, the super poison put is a reinforced version of the simple poison put, and as such it displays all the characteristics, and has all the effects, of the simple put. In other words, a bond cannot include a super poison put covenant without a simple put. This implies that the effects of the three types of covenants cannot be examined in isolation; they have to be estimated simultaneously.
The first complication can be resolved by utilizing either an unordered multinomial probit or a logit model. However, the second complication can only be resolved by the use of unordered multinomial probit, since the logit model requires “independence of the irrelevant alternatives” (see Greene, 1993, p. 671). The unordered multinomial probit, which is employed in the first stage of the estimation process, yields a set of probabilities that correspond to the alternative choices faced by the decision maker. The dependent variableCinow takes values 0 through 4 according to the combination of the covenants included. Table 2 shows how the values of the dependent variable are defined. The multinomial probit model is estimated using a version of the Smooth Simulated Maximum Likelihood Multinomial Probit Model (SSMLLP).6
The estimated values of the probit equation enter the second stage using a multiple regression model, with the reoffering yield differentials (RYi) as the dependent vari-ables as set forth in Eq. (2):
RYi5 bYi1 sv{f(Cˆi)/F(Cˆi)} 1ni (2)
where
Yiis a vector of the independent variables, Cˆiis the fitted value of the probit Eq. (1), f(Cˆi) is the frequency function, and
F(Cˆi) is the cumulative distribution function of Cˆi.
Table 3
Results of the multinomial probit model of the poison covenant inclusion decision
Variables C51 C52 C53 C54
Constant 29.545* 29.315* 28.953* 28.815*
(0.241) (0.312) (0.215) (0.229)
LEV 20.506 0.326 20.525 20.004
(0.556) (0.417) (0.552) (0.616)
PDIR 1.651** 2.070* 2.618* 1.993**
(0.729) (0.631) (0.760) (0.892)
INOWN 0.012** 0.002 0.010** 20.007
(0.005) (0.004) (0.004) (0.011)
LNPRIN 22.893* 23.547* 24.227* 24.199*
(0.727) (0.747) (0.068) (0.081)
Likelihood Ratio Index50.445
The term in parentheses are standard errors. * Significant at the .01 level.
** Significant at the .05 level.
dummy variable) removes the bias of the parameter estimators. In the presence of the selectivity variable, the error termni isN(0,sn2).
4. Results
4.1. The poison covenant inclusion equation
As discussed earlier, firms with high probability of takeover are most likely to include poison covenants in their bond indentures. Therefore, the variables that affect the probability of a takeover are also the variables that are expected to determine the types of covenants to be included. Consequently, the probit model is estimated in conjunction with various combinations of poison covenants specified in Table 2.
According to Harris and Raviv (1988), Stulz (1988), Israel (1991), Hendershott (1991), among others, the main variables that seem to affect the probability of a firm receiving a takeover bid are: a lower level of the existing debt; a larger percentage of the outsiders in the board of directors; and a smaller block of the common shares held by the insiders (i.e., management). The probit model estimated is:
Ci5 a01a1LEVi1 a2PDIRi1a3INOWNi1 a4LNPRINi1ei (3)
The dependent variables, Ci, take values from 1 through 4, with C 50 (the case of straight bond) serving as default. VariablesLEV,PDIR, andINOWNare defined in the previous section, andLNPRINis the logarithm of the principal of the new bond issue. Although the size of the issue may not have any bearing on the probability of a takeover, its interrelationship with the inclusion of poison covenants is of interest. That is, firms issuing large amounts of debt might be required to include some types of poison covenants in their bond indenture irrespective of their status of being takeover candidates.
likelihood ratio index, a measure approximately equivalent to theR2, shows that the
model explains about 45% of the variation in the dependent variables.
The proportion of outside directors and the logarithm of the new bond’s principal turn out to be the variables that display the most consistent statistical significance. The results indicate that investors assign higher probability of takeover to firms with higher proportion of outside directors and, hence, demand some kind of protective covenants. In contrast to our expectation, the logarithm of the new bond’s principal has a consistently negative effect, indicating that poison covenants are encountered more frequently in the smaller issues. The existing level of leverage, however, has no statistically significant effect on the inclusion of poison covenants. Interestingly, the effect of the proportion of insider ownership is positive and statistically significant when the inclusion of a poison call is considered. This result suggests that when insiders control a large block of common shares, they intend to issue bonds with poison calls in an attempt to force the prospective bidders to negotiate directly with them. This is consistent with the Stulz’s (1988) argument of management entrenchment at higher levels of managerial share ownership.
4.2. The reoffering yield equation
In the second stage of the estimation process a linear regression model is used to determine the effects of various poison covenants upon the cost of the firm’s bond issue. The cost is measured as the reoffering yield differential between the bond’s yield to maturity with the yield on a treasury bond having the same time to maturity. The following Eq. (4) is estimated using the Ordinary Least Squares method:
RYi5d01d1LNPRINi1d2TRi1d3VOLi1d4CALLi1 d5SFi
1d6MATi1d7RATGi1 d8CVTi1d9Zi1ni (4)
where
RATGi: a vector of dummy variables describing the ratings assigned to the bonds by either Moody’s or S&P. The highest rating (triple A) is the default variable.
Zi: a vector of the four selectivity variables calculated in Eq. (3). ni: the error term assumed to have zero mean and constant variance.
Tables 4 and 5 present the coefficient estimates of theZ-variables using Moody’s and S&P ratings, respectively. In each table, two sets of estimates are reported: one using the full sample and the other excluding the convertible bonds, reducing the sample to 724 bonds. Regression diagnostics (variance inflation factors) show that the data do not suffer from the multicollinearity problem. The reported standard errors of the coefficients are corrected for heteroskedasticity using White’s (1980) method to yield consistent estimates.
Table 4
Reoffering yield estimation results: Moody’s ratings
Full sample Convertibles excluded
Coefficient Standard Coefficient Standard
Variables # of observations51015 error # of observations5724 error
Constant 3.530* 0.592 2.150* 0.626
LNPRIN 0.141* 0.053 0.054 0.061
TR 20.378* 0.043 20.334* 0.044
VOL 0.122 0.375 20.012 0.380
CALL 0.054 0.078 20.080 0.065
CVT 24.701* 0.122 — —
SF 0.330* 0.111 0.128 0.112
MAT 20.011** 0.005 0.010** 0.005
AA1 0.164 0.131 0.015 0.162
Aa2 0.475* 0.134 0.433* 0.118
Aa3 0.441* 0.153 0.434* 0.111
A1 0.524* 0.124 0.571* 0.107
A2 0.677* 0.119 0.690* 0.104
A3 0.929* 0.130 0.846* 0.113
Baa1 1.286* 0.158 1.069* 0.117
Baa1 1.261* 0.142 1.175* 0.122
Baa3 1.748* 0.235 1.014* 0.248
Ba1 2.377* 0.477 2.282* 0.800
Ba2 2.909* 0.264 3.351* 0.568
Ba3 3.105* 0.176 3.531* 0.188
B1 3.304* 0.191 3.808* 0.238
B2 3.980* 0.171 4.721* 0.166
B3 4.305* 0.246 5.105* 0.308
Caa 4.624* 0.691 3.990* 1.241
Z1 0.230 0.212 0.813** 0.323
Z2 20.434* 0.171 0.177 0.217
* Significant at the .01 level. ** Significant at the .05 level. *** Significant at the .10 level.
Table 5
Reoffering yield estimation results: S&P ratings
Full sample Convertibles excluded
Coefficient Standard Coefficient Standard
Variables # of observations51015 error # of observations5724 error
Constant 3.746* 0.622 2.654* 0.676
LNPRIN 0.103** 0.052 0.027 0.062
TR 20.309* 0.043 20.297* 0.044
VOL 0.339 0.382 20.058 0.351
CALL 0.039 0.079 20.095 0.068
CVT 24.647* 0.114 — —
SF 0.428* 0.110 0.195*** 0.115
MAT 20.016* 0.005 0.006 0.005
Aa1 0.164 0.154 0.004 0.204
AA 0.152 0.107 0.231** 0.101
AA2 0.451* 0.119 0.409* 0.098
A1 0.309* 0.111 0.400* 0.105
A 0.498* 0.094 0.538* 0.093
A2 0.682* 0.108 0.672* 0.095
BBB1 1.012* 0.132 0.934* 0.136
BBB 0.987* 0.147 0.843* 0.144
BBB2 1.730* 0.185 1.151* 0.156
BB1 2.635* 0.189 2.886* 0.273
BB 3.718* 0.665 4.269* 0.776
BB2 2.933* 0.240 3.474* 0.411
B1 3.111* 0.195 3.609* 0.300
B 3.115* 0.179 4.044* 0.266
B2 3.583* 0.154 4.381* 0.151
CCC1 4.173* 0.212 4.991* 0.192
CCC 4.176* 0.501 4.882* 0.849
CCC2 5.314* 0.303 4.778* 0.109
E1 21.712* 0.219 20.562* 0.219
E2 0.587* 0.218 0.331* 0.118
E3 20.357 0.262 20.169 0.273
E4 0.076 0.288 20.007 0.265
E5 20.185 0.221 20.228 0.265
Z1 0.363*** 0.212 0.662** 0.301
Z2 20.578 0.178 0.112 0.240
Table 6
The effects of poison provisions on the reoffering yields of corporate bonds in terms of basis points, 1986 through 1990
Moody’s ratings S&P ratings
Convertibles Convertibles Convertibles Convertibles
Poison covenants included excluded included excluded
Z1: PCALL 0.267 0.542** 0.421*** 0.359***
Z2: PPUT 20.586* 0.230 20.780* 0.146
Z3: PCALL, PPUT 20.391* 20.245*** 20.582* 20.347*
Z4: PPUT, SPP 20.173 20.116 20.416*** 20.361
* Significant at the .01 level. ** Significant at the .05 level. *** Significant at the .10 level.
treasury rate implies that the yield differential shrinks by about 31 to 38 basis points every time the treasury rate’s yield increases by 1%. The coefficients of the ratings variables can be interpreted as the yield differential, in basis points, between a bond and AAA or Aaa bond. For example, the convertibility feature reduces the yield differential by about 470 basis points as reported in Table 4. Interestingly, the results indicate that issuers of the speculative bonds could enjoy the yield differential of almost an AAA bond by making their bonds convertible.
It should be noted that the E-rating bonds are included in the regressions using the S&P ratings, while they are excluded under Moody’s. As pointed out before, the E1 rating indicates the strongest event-risk protection while the E5 indicates the weakest. The signs, the size, and the statistical significance of the coefficients of these ratings are erratic at best. While the rating of E1 has a negative sign and is statistically significant, the rating of E2 has a counterintuitive positive and significant effect on the yield differential.7 Similarly, the rating of E4 in the full sample has a positive,
albeit statistically insignificant, effect. The remaining ratings have the expected nega-tive signs, but they are not statistically significant.
The coefficients of the analysis variablesZ1toZ4are somewhat difficult to interpret
in their present forms. A convenient way is to transform the coefficients into familiar basis points and evaluate the effects of the analysis variables at their corresponding means. Table 6 reports the results.
The relative structures of the effects of poison covenants remain the same under both rating systems. As expected, the presence of a single poison call provision has a positive effect on the yield differential. This positive effect is about 27 to 42 basis points in the full sample and becomes larger, about 36 to 54 basis points, and strongly significant when the convertible bonds are excluded. The effect of a single simple poison put in the full sample is negative and significant, reducing the yield differential by 58 to 78 basis points. When the convertible bonds are excluded, the effect becomes positive, but insignificant.
on the yield differential will emerge. The effect ranges from225 to258 basis points and remains significant even after purging the effects of convertible bonds. The effect of super poison put is negative and insignificant under Moody’s ratings. The effect, however, becomes marginally significant under the S&P rating system, reducing the yield by 42 basis points.
5. Summary and conclusion
This study investigates the effects of three types of poison covenants on the reoffer-ing yields of corporate bonds issued from 1986 through 1990. The covenants include poison call, poison put, and super poison put. Given the intense takeover threats prevailing during this period, an emphasis is placed on the managerial role in deciding to include poison provisions in the bond indenture. It is proposed that managers include poison covenants either to entrench themselves and/or extend protection to bondholders. Furthermore, in this article, it is proposed that the decisions to include poison covenants relate directly to the probability of the firm being a takeover target. Accordingly, a self-selection model is designed to deal with the endogeneity of the decision, and to obtain consistent estimates of the effects of poison covenants on the reoffering yields.
The results show that investors negatively receive the presence of a poison call in the bond indenture. Specifically, we find that bonds with poison call covenants are priced to yield up to 54 basis points higher compared to bonds without such provisions. A poison call allows management to call the bond when a designated event, such as a change in control, occurs. Consistent with the managerial entrenchment, this finding supports our hypothesis that poison call covenants are mainly designed to protect the interests of management rather than bondholders. On the other hand, we find that bondholders generally react positively to bonds containing poison puts. Bonds with single puts are priced to yield up to 78 basis points lower compared to bonds without poison put covenants. Interestingly, our findings indicate that the imposition of an additional stringent provision to convert a simple poison put into a super poison put does not significantly decrease the reoffering yields further. We conclude that placing the redemption right under the control of bondholders can reduce the costs of debt.
Appendix
Examples of event-risk covenants
Poison put
Super poison put
Anheuser-Bush Companies, Inc.: $350 million, 9%, debenture, offered in December, 1989, due December, 2009. Moody’s rating: A1; S&P rating: AA2. Super poison put feature: Upon a designated risk event and downgrading of the debentures to below investment grade, the holders of the debentures may require Company to repurchase the debentures at 100% plus accrued interest.
Poison call
Mesa, LP: $300 million, 13.5% subordinated note, offered in May 1989, due May, 1999; Moody’s rating: Ba3; S&P rating: BB1. Call feature: Callable, as a whole or in part, at any time on or after May 4, 1994 at prices ranging from 105.063% in 1995 to 101.688% in 1997, and 100% thereafter. Poison call feature: If a change in control occurs, the notes may be redeemed at the option of the Obligors, as a whole but not in part, on a redemption date on or before the 120th day following the change in control at prices ranging from 113.500% in 1990 to 101.688% in 1997, and 100% thereafter.
Combination of poison call and put
Mitchell Energy & Development Corp.: $200 million, 11.25%, senior note, offered in February 1989, due February, 1999. Moody’s rating: Ba1; S&P rating: BBB1. Call feature: Callable, as a whole or in part at any time not prior to February 15, 1994, at the option of Company, at prices ranging from 103.21% in 1995 to 101.61% in 1996, and 100% thereafter. Poison call/put feature: The notes are redeemable at the option of the holders, at 100% plus accrued interest to the redemption date, and at the option of Company at the prices set forth herein, plus accrued interest to the redemption date upon (1) the sale of all or substantially all of the assets of Company, or (2) the merger of Company that results from or in, or is attributable to, a change in control.
Acknowledgments
The authors would like to thank an anonymous referee for the insightful comments and Andrea Amelinckx and Linda Janz for their editorial assistance.
Notes
1. A large number of top managers also suffered as a result of the takeover surge of 1980s. Martin and McConnell (1991) report that about 42% of top managers of the target firms were replaced following the year of a takeover. The “change in control” was reported to account for the most frequently reasons cited for the replacement.
2. Some put covenants are triggered only if the takeover is declared ‘hostile.’ 3. See Hessol and Samson (1989) and Hessol (1989) for a description of the
criteria for the E-rated system.
1989 include event-risk covenants in their indentures. They find a significantly higher percentage (about 50%) for unsecured bonds issued by firms subject to takeover attempts or takeover rumors. They conclude that the more likely the firm is to be a takeover target, the more likely the firm is to have explicit protection covenant.
5. Malatesta and Walkling (1988) and Ryngaert (1988) provide evidence of mana-gerial entrenchment in cases involving poison pill securities. Both studies find significant losses to shareholders of firms adopting poison pill defenses. 6. We are heavily indebted to Professor Vassilios Hadjivassiliou of the London
School of Economics for providing us with a copy of the SSMLMP program. 7. The small number of E-rated bonds may be responsible for the inconsistency
of the results.
References
Asquith, P., & Wizman, T. A. (1990). Event risk, covenants and bondholder returns in leveraged buyouts.
Journal of Financial Economics 27, 195–213.
Bae, S. C., Klein, D. P., & Padmaraj, R. (1994). Event risk bond covenants, agency costs of debt and equity, and stockholder wealth.Financial Management 23, 28–41.
Cook, D. O., & Easterwood, J. C. (1994). Poison put bonds: an analysis of their economic role.Journal of Finance 49, 1905–1920.
Crabbe, L. (1991). Event risk: an analysis of losses to bondholders and ’super poison put’ bond covenants.
Journal of Finance 46, 689–706.
Greene, W. H. (1993).Econometric Analysis.New York: Macmillan Publishing Co.
Harris, M., & Raviv, A. (1988). Corporate control contests and capital structure.Journal of Financial Economics 20, 55–86.
Hendershott, R. J. (1991). Debt signaling, managerial ability, and corporate control.Mimeo, Ohio State University.
Hessol, G. (1989). Event risk criteria update.S&P CreditWeek(September 18), 19–20.
Hessol, G., & Samson, S. (1989). Event risk covenant rankings.S&P CreditWeek(July 24), 17–18. Israel, R. (1991). Capital structure and the market for corporate control: the defensive role of debt
financing.Journal of Finance 46, 1391–1409.
Kahan, M., & Klausner, M. (1993). Antitakeover provisions in bonds: bondholder protection or manage-ment entrenchmanage-ment.UCLA Law Review 40, 931–982.
Lehn, K., & Poulsen, A. (1991). Contractual resolution of bondholder-stockholder conflicts in leveraged buyouts.Journal of Law and Economics 34, 645–673.
Maddala, G. S. (1983).Limited-Dependent Variables and Qualitative Variables in Econometrics.Cambridge, MA: Cambridge University Press.
Malatesta, P. H., & Walkling, R. A. (1988). Poison pill securities: stockholder wealth, profitability, and ownership structure.Journal of Financial Economics 20, 347–376.
Martin, K. J., & McConnell, J. J. (1991). Corporate performance, corporate takeovers, and management turnover.Journal of Finance 46, 671–687.
Ryngaert, M. (1988). The effect of poison pill securities on shareholder wealth. Journal of Financial Economics 20, 377–417.
Schleifer, A., & Vishny, R. W. (1989). The case of manager-specific investments.Journal of Financial Economics 25, 123–139.
Stulz, R. M. (1988). Managerial control of voting rights: financing policies and the market for corporate control.Journal of Financial Economics 20, 25–54.
Warga, A., & Welch, I. (1993). Bondholder losses in leveraged buyouts.Review of Financial Studies 6, 959–982.
