Inbreeding is reduced by female-biased dispersal and mating behavior in Ethiopian wolves

Inbreeding is reduced by female-biaseddispersal and mating behavior in Ethiopianwolves

Deborah A. Randall,a John P. Pollinger,b Robert K. Wayne,b Lucy A. Tallents,a Paul J. Johnson,a andDavid W. MacdonaldaaWildlife Conservation Research Unit, University of Oxford, Tubney House, Abingdon Road, TubneyOX13 5QL, UK and bDepartment of Ecology and Evolutionary Biology, University of California, 621Charles E. Young Drive South, Los Angeles, CA 90095, USA

Molecular tools have enabled wildlife researchers to obtain accurate information on the kinship, mating behavior, and dispersalof individuals. We genotyped 192 Ethiopian wolves (n ¼ 29 packs) in the Bale Mountains for 17 microsatellite loci to 1) elucidatekinship within and between packs, 2) assess parentage of pups, and 3) evaluate whether inbreeding is avoided by dispersal and/ormating behavior. Mean pairwise relatedness within packs (R ¼ 0.39) was significantly greater than that estimated from randomassignment of individuals to packs. However, breeding pairs were most often unrelated, suggesting that female-biased dispersalreduces inbreeding. We assigned maternity to 49 pups and paternity to 47 pups (n ¼ 12 litters) using a combination of exclusion,likelihood analyses (using CERVUS software), and sibship reconstruction. Multiple paternity occurred in 33% of litters; extrapackpaternity accounted for 28% of all resolved paternities, occurring in 50% of litters. We found no evidence that extrapackcopulations reduce inbreeding; however, more detailed analyses may elucidate the effect of recent population declines anddemographic disturbances due to recurring disease outbreaks. The adaptive advantages of female-biased dispersal and theobserved mating system are discussed in relation to Ethiopian wolf sociobiology and ecology. Key words: canids, CERVUS,dispersal, EPC, kinship, microsatellites, paternity. [Behav Ecol 18:579–589 (2007)]

The consequences for fitness of group formation are cen-tral to the evolution of sociality (Macdonald et al. 2004),

and natal philopatry is an important mechanism in the for-mation and maintenance of social groups (Waser 1996).Philopatric young may benefit directly from group living(‘‘benefits of philopatry hypothesis,’’ Waser 1988; Stacey andLigon 1991), or they may be ‘forced’ to remain in groups asa result of the high costs or risks of dispersal and independentbreeding (‘‘ecological constraints hypothesis,’’ Selander 1964;Brown 1969; Woolfenden and Fitzpatrick 1978; Koenig 1981;Emlen 1982). The adaptive advantages of dispersal versus phil-opatry are thought to be related to inbreeding avoidance,access to mates, and competition for resources (Clutton-Brock1989; Johnson and Gaines 1990; Creel and Macdonald 1995;Pusey and Wolf 1996). Sex-biased dispersal is often hypothe-sized as a means of avoiding inbreeding with opposite sexparents (Wolff 1992, 1993). Among social canids, this promptsthe question of which sex should disperse given that sub-ordinates are typically the offspring of the dominant pair(Moehlman 1989). Theoretical models predict that juvenileswill be philopatric only if philopatry maximizes their lifetimereproductive success (Brown 1987). As such, mating systemsplay a key role in determining the degree and direction of biasin dispersal (Greenwood 1980; Dobson 1982). Greenwood’s(1980) ‘‘resource competition hypothesis,’’ predicts that malephilopatry predominates among species that are monoga-mous, require paternal investment in offspring care, and ex-hibit resource (i.e., territory) defense by males. Under such

conditions, male philopatry increases the chances of a maleacquiring a high-quality territory with which to attract afemale.The Ethiopian wolf (Canis simensis) lives in packs of up to 13

individuals, which cooperatively defend an exclusive territory(Sillero-Zubiri and Gottelli 1995b). As in many social canids,well-defined hierarchies exist and the dominant female usu-ally monopolizes breeding, producing a litter of up to 6 pupsannually (Sillero-Zubiri et al. 2004). All group members con-tribute to parental care, but there is little evidence that help-ers accrue inclusive fitness benefits through increased survivalof related offspring (Sillero-Zubiri et al. 2004). Dispersal ap-pears to be extremely costly for Ethiopian wolves due to hab-itat saturation in their Afroalpine ‘‘island’’ habitats and thelack of unoccupied territories on which to breed indepen-dently (Sillero-Zubiri and Gottelli 1995b; Sillero-Zubiri, Gottelli,and Macdonald 1996). Observational data suggest that thestructure of packs and formation of breeding units occur by3 main mechanisms: 1) males are predominantly philopatric,and the few females that remain in the natal pack may even-tually acquire breeding status 2) most females dispersearound 2 years of age, and some become ‘‘floaters’’ that maysuccessfully immigrate into existing packs, 3) subordinate fe-males form alliances with same-pack males to split a natal pack(Sillero-Zubiri 1994; Sillero-Zubiri, Gottelli, and Macdonald1996; Marino 2003; Sillero-Zubiri et al. 2004).Previous studies have explained Ethiopian wolf sociality

in terms of access to resources (Sillero-Zubiri 1994; Marino2003). Ethiopian wolves forage solitarily on the abundant ro-dents of the Afroalpine ecosystem (Sillero-Zubiri and Gottelli1995a). Where resources are particularly rich, Ethiopianwolves appear to tolerate relatives that remain in the pack,thereby enabling group size increase and territory expansion(sensu Kruuk and Macdonald 1985). Under these conditions,

Address correspondence to D.A. Randall. E-mail: [email protected] 5 August 2006; revised 11 January 2007; accepted 3

February 2007.

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wolves in larger groups gain the benefit of more prey-richhabitat per individual (Marino 2003). Where resources arescarce, groups are smaller, typically consisting of pairs of in-dividuals (Marino 2003). Thus, the benefits of sociality appearto be primarily acquisition, inheritance, and/or defense of ahigh-quality territory (Marino 2003). Resource-based explan-ations for sociality may also underlie female-biased dispersalin Ethiopian wolves, which contrasts with the predominantlymale-biased dispersal pattern in other mammals (Greenwood1980). Based on the assumptions of the resource competitionhypothesis (Greenwood 1980), males may have a greater stakein resource defense and the prospect of future territoryinheritance. These predictions for Ethiopian wolves aresupported by the pattern of male philopatry and greatermale investment in territorial defense (Sillero-Zubiri andMacdonald 1998).It is well recognized that multiple factors often determine

the relative costs and benefits of dispersal versus philopatry(Macdonald and Carr 1989). Among Ethiopian wolves, in-breeding avoidance may be an additional adaptive advantageof female-biased dispersal and may underlie the observed mat-ing behavior. Male philopatry and the long tenure of breedingfemales increase the potential for incest within groups, whichmay be countered in part by female dispersal (Sillero-Zubiriet al. 2004). However, given the extent of habitat saturation,constraints on dispersal, and occasional female philopatry,alternative behavioral mechanisms may have evolved to avoidmating with close kin. A previous study found that 70%of observed copulations were between dominant femalesand males in neighboring packs (Sillero-Zubiri, Gottelli, andMacdonald 1996), and genetic analyses provided preliminaryevidence for multiple paternity (Gottelli et al. 1994). It hasbeen suggested that such extrapack copulations (EPCs) maybe an inbreeding avoidance strategy for Ethiopian wolves(Sillero-Zubiri, Gottelli, and Macdonald 1996; Sillero-Zubiriet al. 1998), enabling females to gain access to geneticallydissimilar males (see also Jennions and Petrie 2000). However,this remains unsubstantiated, and alternative explanations forEPCs are possible (e.g., deterring infanticide by neighbors,Wolff and Macdonald 2004).We combined demographic and spatial information col-

lected during field observations of focal Ethiopian wolf packsin the Bale Mountains with multilocus microsatellite data to1) determine genetic relatedness within and between socialgroups and 2) assess whether inbreeding avoidance underliesdispersal and/or mating behavior. Based on previous observa-tional data, we expected packs to be predominantly close-kin

groups (Sillero-Zubiri 1994; Marino 2003; Sillero-Zubiri et al.2004). However, if sex-biased dispersal is a mechanism toavoid inbreeding, then we predicted that breeding individualswould be unrelated and incestuous matings would be uncom-mon. We identified mated pairs based on genetic evidence ofparentage, which was also used to quantify the extent of mul-tiple- and extrapack paternity (EPP) in litters. Paternity andrelatedness data were used to test the hypothesis that EPCsenable females to obtain genetically dissimilar mates outsidetheir social group.

METHODS

Study population

We focus predominantly on 2 subpopulations of Ethiopianwolves in the Bale Mountains: the Web Valley (WV) duringthe 2002–2003 breeding season and the Sanetti Plateau (SN)during the 2003–2004 breeding season (Figure 1). Packs wereidentified easily by different group sizes and territory loca-tions. Pack compositions were determined by completeenumeration of individuals observed on multiple occasionsaround the den, during social greetings, and on boundarypatrols (as in Marino et al. 2006). To account for the possibil-ity of missing pack members from a given group sighting, packsize and age/sex structure were deduced from multiple obser-vations of the pack over the course of the breeding seasonuntil no new individuals were recorded for that pack. Individ-uals were classified as adults (.2 years), yearlings (1–2 years),or pups (,1 year). Litter size was defined as the maximumnumber of pups at emergence from dens (Sillero-Zubiri et al.1998). Territories were represented by minimum convex poly-gons after removing 5% of outliers in ARCVIEW version 3.0(Environmental Systems Research Institute, Inc., Redlands,CA) using yearly (April–March) GPS locations (Garmin 12GPS software with map datum WGS 84) pooled for each pack(as in Sillero-Zubiri and Gottelli 1995b).

Sample collection and analysis

Fecal samples were used as our primary source of DNA be-cause we favored a noninvasive approach to DNA sampling.Because individuals without ear tags or other artificial markswere not always individually identifiable, a surplus of fecalsamples was collected from each pack and each litter of pupsso as to maximize the chance that each animal was sampled atleast once. Fecal samples were collected from wolves seen

Figure 1The distribution of Ethiopianwolf packs in (a) the WV and(b) SN. Focal packs are de-picted by black polygons; non-focal packs considered in thepaternity analyses are depictedby white polygons. Inset showswolf range in the Bale Moun-tains and the location of the 2subpopulations.

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defecating so that age, sex, and pack could be determined.Pup feces were collected around the entrance to dens. We alsogenotyped tissue samples from 27 wolves that died duringa rabies outbreak in WV in 2003 (Randall et al. 2004) and85 wolves livetrapped during the subsequent emergency vacci-nation campaign (Knobel D, Fooks AR, Brookes S, Randall DA,Williams SD, Argaw K, Shiferaw F, Tallents LA, Laurenson MK,unpublished data). All samples were preserved in 96%ethanol.We extracted DNA from feces using QIAamp DNA Stool

Mini Kits (QIAGEN, Valencia, CA) and from tissue usingQIAamp DNA Mini Kits (QIAGEN) according to the manu-facturer’s protocol. Negative controls were included in eachset of extractions to monitor for contamination. We selected17 microsatellites from the dog genome (Breen et al. 2001)that were polymorphic in Ethiopian wolves, including 16 tet-ranucleotide repeat loci (FH2001, FH2054, FH2119, FH2137,FH2138, FH2140, FH2159, FH2174, FH2226, FH2293,FH2320, FH2422, FH2472, FH2537, PEZ17, and PEZ19) and1 dinucleotide repeat locus (C05.377). Sex allocation was con-firmed using a Y-chromosome microsatellite locus and theprimer pairs MS34 or MS41 (Olivier et al. 1999). Microsatel-lites were amplified using the QIAGEN Multiplex PCR Kit in10 ll volumes containing 1.5 ll DNA, 1.0 ll primer mix, 0.4 ll10 mg/ml bovine serum albumin, 5.0 ll QIAGEN master-mix,and ddH2O. Amplifications were performed on a programma-ble Peltier Thermal Cycler (MJ Research [Waltham, MA] PTC-200). Genotyping reactions were run on an automated se-quencer with fragment sizes checked manually with referenceto a size standard. Fecal samples were genotyped using amultiple-tubes approach (Navidi et al. 1992; Taberlet et al.1996) in which both alleles at a heterozygous locus were am-plified at least twice and the single allele of a homozygouslocus was amplified at least 3 times. Consensus genotypes werecompleted for a minimum of 14 loci. Complete details of the

genotyping methods and validation techniques are availableelsewhere (Randall 2006; Randall DA, Pollinger JP, TallentsLA, Macdonald DW, Wayne RK, unpublished data).We used CERVUS version 2.0 (Marshall et al. 1998) to

calculate allele frequencies. The Web version of GENEPOP(Raymond and Rousset 1995) was used to test for deviationsfrom Hardy–Weinberg equilibrium per locus within subpopu-lations and overall. We also used GENEPOP (Raymond andRousset 1995) to test for linkage disequilibrium (LD) withinsubpopulations and overall using Fisher’s exact tests. We ex-cluded yearling and pup genotypes to reduce the effect ofdifferent generations and applied adjusted P values at the0.05 nominal level using Bonferroni corrections to accountfor multiple tests (Rice 1989).We calculated the probability that 2 siblings would have

identical genotypes by chance in our population (PIDsib, asin Evett and Weir 1998; Waits et al. 2001) using the programGIMLET version 1.3.1. (Valiere 2002). PIDsib is a conservativeestimate of the power to resolve between individuals wherepopulations exhibit substructure or where comparisons aremade between related individuals (Waits et al. 2001). Theprogram IDENTITY (Allen et al. 1995) was used to checkfor duplicate genotypes in the data set.

Parentage

Maternity and paternity were assigned according to specificrules (Figure 2). Parentage was assigned initially using thelikelihood-based approach in CERVUS version 2.0 (Marshallet al. 1998). This method accounts for the proportion of can-didate parents sampled, allows for genotyping errors, and cal-culates statistical confidence based on the difference in LOD(i.e., the logarithm of the likelihood ratio) scores of candidateparents. However, this method is limited by its inability to

Figure 2Decision-making matrix forparentage assignments. Thenumber of parentage assign-ments made at each stage isshown in parentheses.

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assign parentage to unsampled individuals within or outsidethe study population or to reliably resolve correctly betweenclose relatives (Marshall et al. 1998; Morrissey and Wilson2005). To account for this and further enhance our parentageassignments, we used COLONY version 1.2 (Wang 2004) topredict the most likely number of parents within each sub-population (and social group) by inferring sibling relation-ships among offspring.Because it is unlikely for yearlings to breed (Sillero-Zubiri

1994), we excluded them from the candidate parent pool.Furthermore, long female breeding tenure (Sillero-Zubiri,Gottelli, and Macdonald 1996) meant that yearlings in subpo-pulations were probably full or half sibs of the pups in theirpack and might produce higher likelihood scores than thetrue parents (Marshall et al. 1998; Morrissey and Wilson2005). Litters were defined as one or more pups with the samemother in the same birth year. Maternity was assigned by ex-clusion if only one adult female in a pack had zero mis-matches with one or more pups in her pack (as in Bakeret al. 2004). Otherwise, we used CERVUS to analyze the likeli-hood of paternity for candidate males with respect to all non-excluded mothers.First, we performed an open paternity analysis in which all

males in a given subpopulation and, to account for EPCs,males in packs adjacent to each subpopulation were consid-ered candidate fathers (Figure 1). We also performed a re-stricted analysis in which only within-pack adult males oradult males in neighboring packs were considered candidatefathers. Packs with adjoining territories were consideredneighbors in the restricted analysis (Figure 1). Males (ormother–father pairs where multiple mothers were not ex-cluded) were allocated parentage if they were assigned at95% or higher confidence. Alternatively, males (or mother–father pairs) were allocated parentage if they were assigned at80% or higher confidence if they produced only one or zeroparent–offspring mismatches, which were accepted to allowfor the possibility of genotyping errors and/or mutations.However, assignments with more than one mismatch at thelower confidence level were considered unreliable and re-jected. If the results of the open and the restricted paternityanalyses differed, the father (or mother–father pair) with thehigher confidence was accepted. If the confidences wereequal, the result of the restricted analysis was accepted be-cause within-pack males or males from neighboring packswere more likely to mate with the dominant female thanmales from nonneighboring packs (Sillero-Zubiri, Gottelli,and Macdonald 1996). If more than one mother–father pairwas still not excluded, then we accepted the results with thehigher confidence or with fewer mismatches if the confiden-ces were equal. If the results for different mother–father pairswere equal, then parentage was unresolved using CERVUSand sibship reconstruction using the program COLONY wasused to infer genealogical relationships further.We then used COLONY to assess parentage of pups with yet

unresolved maternity or paternity by inferring full and halfsiblings among pups. Each subpopulation was analyzed sepa-rately with no parents assigned a priori in the first instanceand with mothers assigned a priori where possible in the sec-ond instance (Figure 2). Pups with unresolved mothers (orfathers) were assigned maternity (or paternity) if they wereidentified as full siblings of one or more pups for which ma-ternity (or paternity) was resolved based on exclusion and/orCERVUS analysis.

Simulation parameters

The error rate for the parentage simulations (termed eg inMorrissey and Wilson 2005) was set at the default of 0.01,

which is slightly more conservative than the error rate(0.008) estimated for our data set (Randall 2006; RandallDA, Pollinger JP, Tallents LA, Macdonald DW, Wayne RK, un-published data). The error rate used in our parentage likeli-hood assignments (termed el in Morrissey and Wilson 2005)was set to 0.001. Although the advanced options in CERVUSallow eg and el to vary, most studies set eg ¼ el. Morrissey andWilson (2005) demonstrated that setting 0 , el , eg (i.e.,underestimating the true error rate in the likelihood tests)produced higher assignment success while remaining robustto genotyping errors (i.e., not excluding true parents on thebasis of mismatches). Open and restricted simulations ofpaternity were run separately for each pack using subpopula-tion- and pack-specific parameters, including the number ofcandidate fathers, the proportion of candidate fathers sam-pled, and the effect of relatives in the population (see Sup-plementary Table S1). We simulated the effect of relatives onthe paternity results by including average relatedness amongwithin-pack adult males and the average number of adultmales in each pack as relatives of the true parent. Simulationswere run for 10 000 cycles, and statistical significance was de-termined at 80% and 95% confidence. In COLONY, we usedsubpopulation-specific allele frequencies and genotyping er-ror rates (Randall 2006; Randall DA, Pollinger JP, Tallents LA,Macdonald DW, Wayne RK, unpublished data).

Relatedness, kinship, and inbreeding

We applied rarefaction analysis (by Geffen E, as in Girmanet al. 1997) to determine the number of microsatellitesneeded for accurate estimates of relatedness (Queller andGoodnight 1989). Pairwise relatedness (R) was determinedusing RELATEDNESS 5.0 (Queller and Goodnight 1989) afterincluding a bias correction of population allele frequencies bypack. Standard errors (SEs) were determined by jackknifingover loci. Unrelated pairwise R values were estimated from 15dyads of males in nonneighboring packs in a subpopulation.We also tested whether individuals in the same pack weremore related to each other than to individuals in other packsand whether individuals in neighboring packs were more re-lated to each other than to individuals in nonneighboringpacks. To do this, we compared observed pairwise relatednessvalues for dyads within packs (and for dyads between neigh-boring packs) to permutations that randomly assigned allindividuals to packs with the same size and age/sex structureas those observed (after Girman et al. 1997; de Ruiter andGeffen 1998). Permutation tests were carried out separatelyfor the 2 subpopulations using SAS macros (SAS 1990). Re-sults were considered statistically significant if the observedvalues were in the range of either the lower or upper 2.5%of the simulated values in 1000 permutations (Manly 1994).Likelihood analyses implemented in KINSHIP (Goodnightand Queller 1999) were used to assess kinship between adultswithin packs.To test for inbreeding avoidance, we compared the related-

ness of breeding females to within-pack fathers versus extra-pack fathers using a pairwise t-test (SPSS Inc., Chicago, IL). Inorder to determine whether EPCs might enable females toobtain genetically unrelated mates, we compared the averagerelatedness between breeding females and adult males intheir packs to simulations in which males were randomlyassigned to packs, keeping the same number of males in eachpack as that observed. Because relatedness may be higherbetween neighboring packs, we also compared relatednessbetween breeding females and adult males in their packs tosimulations in which the males randomly assigned to packswere restricted to those in neighboring groups.




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RESULTS

Genetic data set

There were no significant heterozygote deficits for any locusor subpopulation following Bonferroni correction (P . 0.05).Mean FIS values were not significantly different from zero butslightly negative (�0.04 to �0.09, see Supplementary TableS2), suggesting randommating within subpopulations (Dobsonet al. 1997). Seventeen pairs of loci showed significant LDfollowing Bonferroni correction (P , 0.05), but patterns werenot consistent across subpopulations suggesting that LD isprobably due to population structure rather than physicallinkage. Because weak linkage is unlikely to affect the analyses(Meagher 1986), we included all 17 loci in our analyses.The 17 microsatellites in our data set had a low overall prob-

ability of identity among siblings (PIDsib ¼ 3.79 3 10�6). Thisis equivalent to approximately 0.0004% of full siblings sharingthe same genotype by chance, implying that identical geno-types were unlikely in our population. Even if the 3 most in-formative loci (i.e., those with the highest per locus PIDsibvalues) were excluded due to missing data, the overall PIDsibwas 1.55 3 10�5 (approximately 0.002% chance of identicalgenotypes). Our relatedness values also changed little after12 loci were used, implying that the 17 loci used in our analysesprovided high resolution (see Supplementary Figure S1). Lo-cus C05.377 had small, positive null allele frequencies in WVand SN according to CERVUS, thus we ignored homozygous–homozygous parent–offspring mismatches at this locus.

Parentage

Our study population consisted of 179 individuals in 17 focalpacks from which we obtained 148 individual genotypes(Table 1). Pack sizes (adults and yearlings) ranged from 2 to11 with a mean 6SE of 7.2 6 0.7 (n ¼ 17). Seven packs (41%)had multiple adult females (2 adult females in 5 packs; 3 adult

females in 2 packs). Only 13 of 17 (76%) packs reproducedsuccessfully (i.e., pups emerged from the den), producing atotal of 57 pups. Litter sizes at emergence ranged from 1 to 6pups (mean 6 SE ¼ 4.4 6 0.5, n ¼ 13).We analyzed parentage for 52 of 57 (91%) offspring born in

our study population (5 pups were either not sampled or hadincomplete genotypes, Table 1). Maternity was determined for49 pups in 12 litters, of which 48 were assigned to the domi-nant female and 1 to a subordinate female (GAR, Table 2).In all cases, dominant females were confirmed behaviorallyby pregnancy and/or lactation, although the subordinatebreeder showed no such evidence of having bred. Only onepup (MEG) had more than one adult female that could notbe excluded as the true mother, and both adult femalesshowed signs of pregnancy and lactation suggesting that theyboth reared pups that year. Mother–offspring mismatches oc-curred for only 3 pups (2 in DAR, 1 in DOD), but maternityassignments were determined by sibship reconstruction andconfirmed by mother–father–offspring trios assigned at 95%confidence. Given a mutation rate of 10�2 to 10�3 per gener-ation for canid dinucleotide and tetranucleotide markers(Francisco et al. 1996), we could expect approximately 8mother–offspring mismatches in our data set due to mutationalone. Maternity was unresolved for 2 pups in MUL pack be-cause the only adult female in that pack was not genotyped(Tables 1 and 2).Forty-seven of 49 pups with resolved maternity were sired by

males we genotyped in this study, and a minimum of 14 fa-thers were identified for the 12 litters with resolved maternity(Table 2). Thirty-seven paternities were assigned at 95% con-fidence, 6 paternities were assigned at 80% confidence, and 4paternities were assigned by sibship reconstruction. Paternitywas unresolved for 3 pups with unresolved maternity (1 inMEG and 2 in MUL) and 2 pups for which no genotyped malecould be assigned (BBC). A single father sired all the pups in7 litters for which all pups were genotyped, and a single father

Table 1

Composition of packs in WV and SN subpopulations

Adults Yearlings Pups

Pack Female Male Female Male Female Male Unknown Total

WV

ALA 1 (1) 3 (3) 2DAR 2 (1) 3 (3) 2 (2) 2 (1) 4 2DOD 1 (1) 2 (2) 1 (1) 3 (1) 2 2KOT 1 (1) 4 (4) 2 (0) 3 (3) 2 4MEG 2 (2) 3 (2) 2 (1) 1 (1) 1 2MUL 1 (0) 4 (2) 1 (1) 3 (1) 1 1 2SOD 1 (0) 1 (1)TAR 1 (1) 5 (3) 2 (0) 2 (2) 3 2WOL 1 (1) 1 (1) 1Total 11 (8) 26 (21) 10 (5) 14 (9) 15 12 4 92 (70)

SN

BAD 1 (1) 3 (3) 4 (4) 2 (2) 2 4BAT 2 (1) 3 (3) 2 (2) 1 (1) 3 2 1BBC 1 (1) 4 (3) 3 (3) 1 (1) 2 2GAR 3 (3) 3 (3) 2 (1) 3 (3) 3 1LEN 2 (1) 4 (3)NYA 3 (3) 3 (2)QUA 2 (2) 4 (4) 2 4SUL 1 (0) 4 (3)Total 15 (12) 28 (24) 11 (10) 7 (7) 12 13 1 87 (78)

Values in parentheses indicate the number of individuals genotyped in each sex/age class. Only pupswith a designated sex were genotyped.




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was assumed in an eighth litter (BAT) with one unsampledpup. Two fathers were confirmed in 3 litters in which all pupshad resolved paternity and suspected in a fourth litter (BBC)in which 2 pups had unresolved paternity. Thus, the numberof sires per female was 1–2, and the rate of multiple paternitywas 33% (4 of 12 litters). Overall, 13 pups with resolved pa-ternity were sired by extrapack males (Table 2); in all cases thefather was from a neighboring pack (Figures 1 and 5). Threebreeding females (DOD, KOT, BAD) had both within packsires and extra-pack sires. A fourth breeding female (BBC)is assumed to also have both a within pack and extra-pack sire(if the 2 pups with unresolved paternity were sired by theunsampled within pack male). Six females had only withinpack sires, and 2 females (both in GAR) had only extra-packsires. Our paternity results suggest an EPP rate of 28% (13 of47) of pups and 50% (6 of 12) of litters.Overall, adult males sired 0 to 6 pups in a breeding season,

but reproductive success among males was highly skewed with67% and 75% of males analyzed siring zero offspring in WVand SN, respectively. Dominance appeared to be an importantpredictor of reproductive success. In 4 of 6 packs for whichsocial status within packs was deduced, the dominant male wasassigned paternity of all pups (DAR, n ¼ 6; TAR, n ¼ 5, WOL,n ¼ 1), and in the other 2 the dominant male was assignedpaternity for the majority of pups (KOT, n ¼ 5 of 6; DOD, n ¼3 of 4).

Relatedness

Mean (6SE) pairwise R values calculated for dyads withknown relationships were similar to the theoretical values of0.5, 0.5, and 0 for parent–offspring, full sibling, and unrelateddyads, respectively (mother–offspring, n ¼ 49, R ¼ 0.494 6

0.029; father–offspring, n ¼ 47, R ¼ 0.511 6 0.032; fullsiblings, n ¼ 78, R ¼ 0.522 6 0.025; unrelated, n ¼ 15,R ¼ �0.031 6 0.042), but half siblings had a higher mean Rthan the expected value of 0.25 (n ¼24, R ¼ 0.313 6 0.050,Figure 3).Mean (6SE) pairwise R values within packs were 0.298 6

0.016 for adult and yearling dyads (n ¼ 16), 0.392 6 0.065 foradult females (n ¼ 4), and 0.333 6 0.023 for adult males (n ¼15) (Figure 3). All values were significantly higher than re-latedness determined by random distribution of individuals(P , 0.001, Table 3), except for adult female comparisonsin WV for which we had only one dyad. Mean relatednessbetween individuals in neighboring packs was also lower thanthat between individuals in the same pack for all age/sexcategories but did not consistently differ significantly fromrelatedness determined by randomly allocating wolves toneighboring packs (Table 3). In WV, relatedness was signifi-cantly higher between neighbors than a random distributionfor adult and yearling comparisons (R ¼ 0.069, P , 0.0001)and male-only comparisons (R ¼ 0.104, P , 0.0001) but not

Table 2

Parentage assignments for Ethiopian wolf pups in WV and SN subpopulations

Paternities resolved

Maternities resolved Likelihood

PackNo. ofpupsa Exclusion

FullSibship

No. ofmothers 80% 95%

Fullsibship

No. offathers

No. ofEPP

WV

ALA 2 (2) 2 1 2 1DAR 6 (6) 4 2 1 6 1DOD 4 (4) 3 1 1 1 3 2 1KOT 6 (6) 6 1 2 4 2 1MEG 3 (1) 0b 0 — — — — — —MUL 4 (2) —c — — — — — — —TAR 5 (5) 5 1 2 2 1 1WOL 1 (1) 1 1 1 1

SN

BAD 6 (6) 6 1 1 4 1 2 5BAT 6 (5) 5 1d 5 1e

BBC 4 (4) 4 1 1 1 2f 2GAR 4 (4) 311g 2 3 1 111 311QUA 6 (6) 5h 1h 1 6 1

TOTAL 45 4 12 6 37 4 14i 13

a Figures in parentheses indicate the number of pups genotyped.b Two females had zero mismatches and neither could be assigned.c The only female in the pack was not genotyped.d Maternity not resolved for pup not genotyped, but singular breeding is assumed.e The minimum number of father assuming pup not genotyped has the same father.f Two pups with unresolved paternity assumed to be fathered by unsampled within-pack male.g Three females had zero mismatches, 3 pups assigned to dominant female, and 1 pup assigned tosubordinate female by CERVUS.

h Two females had zero mismatches, 5 pups assigned to dominant female by CERVUS, and 1 pup assignedto dominant female by sibship reconstruction.

i Two males fathered pups in 2 separate litters (KOT dominant male: 5 pups KOT and 1 pup DOD; BADmale of unknown social status: 1 pup BAD and 2 pups BBC).




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female-only comparisons (R ¼ �0.084, P ¼ 0.90). In SN therewas no significant difference between neighbor relatednesscompared with a random distribution for adult and yearlingcomparisons (R ¼ 0.014, P ¼ 0.90) and male-only compari-sons (R ¼ �0.026, P ¼ 0.090) but significantly higher related-ness for female comparisons (R ¼ 0.054, P , 0.0001).

We used likelihood tests (Goodnight and Queller 1999)to determine if we could classify kinship among adults withinpacks. For the full sibling/parent–offspring (primary hypoth-esis) versus half sibling (null hypothesis), the estimated pro-portion of tests resulting in type II errors (falsely accepting thenull hypothesis) at P ¼ 0.05 was 0.348 and 0.323 in WV and

Figure 3Mean pairwise relatedness(6SE) for (a) individuals withknown kinship and (b) individ-uals with unknown kinship.Numbers above bars indi-cate (a) number of dyads or(b) number of packs from whichmean values were calculated.

Table 3

Mean pairwise relatedness (R) between individuals in the same pack and between individuals inneighboring packs in WV and SN

Within packs Between neighboring packs

nObservedR value

RandomizedR value P value n

ObservedR value

RandomizedR value P value

WV

Adults and yearlings 8 0.224 0.000 ,0.0001 9 0.069 0.000 ,0.0001Adult males 7 0.322 0.000 ,0.0001 9 0.104 0.000 ,0.0001Adult females 1 0.134 0.030 0.13 9 �0.084 �0.083 0.90

SN

Adults and yearlings 8 0.314 0.000 ,0.0001 8 0.014 0.028 0.90Adult males 8 0.297 0.040 ,0.0001 8 �0.026 0.022 0.90Adult females 3 0.459 0.010 ,0.0001 8 0.054 0.000 ,0.0001

Observed values are compared to those obtained from permutations that randomly assignedindividuals of the appropriate age/sex to packs, keeping pack age/sex structure consistent with thatobserved. Number of packs for which we had pairwise comparisons (n) are shown.




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SN, respectively, even assuming R ¼ 0.25 for half siblings.Thus, substantial overlap in the distribution of simulated Rvalues makes the results of such a test equivocal (Figure 4).Predicted type II error rates were similarly high (0.426 and0.450) for the significance test of half siblings versus unrelatedindividuals. Therefore, we tested whether adult dyads withinpacks were significantly more likely to be full siblings/parent–offspring or unrelated as this produced lower type II errorrates (0.038 and 0.033). Overall, 100% of female dyads (n ¼8) and 77% (36 of 47) male dyads were significantly morelikely to be full siblings/parent–offspring than unrelated.On the other hand, only 43% (24 of 56) female–male dyadswere significantly more likely to be full siblings/parent–off-spring than unrelated. Within packs, mean relatedness be-tween opposite sex adults (n ¼ 14, R ¼ 0.103) was alsolower than relatedness for same-sex comparisons (Figure 3),and the pattern was fairly consistent across packs (data notshown).

Adult kinship patterns within packs varied (Figure 5).Three packs (ALA, DOD, BAT) consisted of 2 or more relatedmales (R ¼ 0.23–0.72; KINSHIP full sibling/parent–offspringtest, all P, 0.05) paired to an unrelated female (R ¼ �0.49 to–0.08, all P . 0.05). Four packs (DAR, KOT, TAR, BAD) con-sisted of only an unrelated dominant pair (R ¼ �0.20 to 0.06,all P . 0.05) and philopatric male offspring related to boththe dominant individuals (R ¼ 0.27–0.53, all P , 0.05) or tothe dominant female only (one pup in DAR, R ¼ 0.27, P ,0.05). In BBC and GAR, adults had mixed kinship: BBC had 2related males (R ¼ 0.54, P , 0.05) unrelated to the breedingfemale (R ¼ �0.27, both P . 0.05) and 1 adult male that mayhave been the offspring of the dominant female (R ¼ 0.34,P , 0.05); GAR had 1 adult male unrelated to the dominantfemale (R ¼ �0.16, P . 0.05), 1 adult male that may havebeen the offspring of the dominant female (R ¼ 0.16, P ,0.05), and one adult male that was unrelated to the dominantfemale (R ¼ 0.04, P. 0.05) and unrelated to the other 2 adult

Figure 5Kinship among adults withinfocal packs in (a) WV and (b)SN. Where known, dominantfemales and their within-packmates are shaded in gray; un-sampled individuals are shadedin black. Solid lines indicate in-dividuals that are significantlymore likely to be full siblingsor parent–offspring pairs thanunrelated (KINSHIP, P, 0.05).Dotted arrows indicate extrapackpaternities obtained by neigh-boring males with the numberofpaternities indicated.Thecon-figuration of packs approximatesthe spatial distribution of terri-tories (depicted in Figure 1).

Figure 4Distribution of R values for un-related, half-sibling, and full-sibling dyads based on 1000simulated individuals in WVand SN using KINSHIP (Good-night and Queller 1999).




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males (R ¼ �0.11 to 0.22, both P . 0.05) but related as a fullsibling/parent–offspring to a subordinate female (R ¼ 0.31,P, 0.05). Two subordinate adult females in GAR were relatedas full sibling/parent–offspring to the dominant female (R ¼0.50 and 0.54, both P , 0.05) and one or more males (R ¼0.31–0.72, all P , 0.05), implying female philopatry. We alsoobserved strong kinship between 2 adult females in both MEG(R ¼ 0.13, P , 0.05) and QUA (R ¼ 0.50, P , 0.05), althoughwe had insufficient evidence to infer sister or mother–daugh-ter relationships. WOL and QUA appeared inbred with Rvalues consistent with full sibling/parent–offspring kinshipin almost all pairwise comparisons (R ¼ 0.27–0.71, all P ,0.05). Only 2 adult males were unrelated to all other adultssuggesting possible male immigration (BAD, R ¼ �0.10 to0.05, all P . 0.05; NYA, R ¼ �0.28 to 0.08, all P . 0.05). Allyearlings were related to the dominant female in their pack asparent–offspring (R ¼ 0.31–0.67, all P , 0.05).

Inbreeding

Kinship analysis showed that 11 of 15 (73%) of successfulmatings (7 of 9 within pack and 4 of 6 extrapack) were be-tween unrelated individuals (R ¼ �0.26 to 0.22, P . 0.05).Four apparently incestuous matings resulted in 9 of 47 (19%)pups being potentially inbred. In 2 packs (WOL and QUA),females may have been related to the within-pack sires of theirpups (R ¼ 0.18 and 0.44, respectively; KINSHIP full sibling/parent–offspring test, both P , 0.05). In KOT and GAR (sub-ordinate female), females mated with extrapack males thatwere significantly more likely to be related to them as fullsiblings/parent–offspring than unrelated (R ¼ 0.41 and0.44, P , 0.01). On average, breeding females were more re-lated to adult males in their pack than to adult males in otherpacks in the same subpopulation. For WV, the mean R valuewas 0.093 (mean of randomizations ¼ �0.017, permutationtest, P ¼ 0.02); for SN, the mean R value was 0.034 (mean ofrandomizations ¼ �0.123, P , 0.01). However, breeding fe-males were similarly related to males in their pack and tomales in neighboring packs in WV (mean of randomizations¼0.077, P ¼ 0.65) and SN (mean of randomizations ¼ 0.020,P ¼ 0.60). Actual mean R values between breeding femalesand neighboring males were 0.048 and �0.039 in WV and SN,respectively. In 2 of 3 packs where breeding females matedboth within and outside their pack, females were more (notless) related to extrapack sires of their offspring than within-pack sires, but the difference was not significant (pairwiset-test: t ¼ �0.944, df ¼ 2, P ¼ 0.445), but our sample sizewas small.

DISCUSSION

Our results demonstrate higher relatedness within Ethiopianwolf packs than between packs, implying that social cohesionand cooperation are associated with kinship ties. Values weresimilar for adults (females R ¼ 0.39, males R ¼ 0.33) andyearlings (R ¼ 0.30), all of which approximate relatednessamong half siblings (R ¼ 0.31). These results have also beenfound in other social canids, such as African wild dogs (Lycaonpictus, R ¼ 0.27 overall within packs, R ¼ 0.28 for adult fe-males, and R ¼ 0.35 for adult males, Girman et al. 1997).Although adults of the same sex are typically related, breedingpairs were typically unrelated, including both within-packmates (7 of 9 breeding pairs) and extrapack mates (4 of 6breeding pairs). However, like most canids, Ethiopian wolvesalso show large variation in mean kinship within packs, whichis probably due to variation in the number of EPPs amonglitters and degree of natal philopatry among offspring. Obser-vational data suggest that packs and breeding units are usually

formed by 1) alliances between one or more males and animmigrant floater female, 2) pack fission by a subordinatefemale and one or more same-pack males that together splitfrom their natal pack, and 3) subordinate female ascendanceto breeding status after the death of the incumbent (Sillero-Zubiri 1994; Marino 2003; Sillero-Zubiri et al. 2004). Our ge-netic data on kinship within packs lend support to the occur-rence of all 3 mechanisms, although the latter 2 are difficult todiscern from genetic evidence alone.In packs for which we had adequate data on kinship (n ¼

11), breeding females were either unrelated to their within-pack mates (n ¼ 6) or unrelated to at least one male in theirpack (n ¼ 3). In the remaining packs (n ¼ 2), we can infer thefollowing. In QUA pack, the breeding female was related to allother individuals in her pack, except one adult male but in-cluding a subordinate female. Thus, she ascended to breedingstatus in her natal pack (in which case the subordinate femaleis her sister or daughter), split from her natal pack with otherindividuals, including the subordinate female (alternativelythe subordinate female may be her daughter from a subse-quent litter), or mated with her son (in which case the un-related male may be her mate from a previous breedingseason and the other individuals their offspring). In WOLpack, the 2 related individuals are probably siblings that splitfrom their natal pack together. These results provide geneticevidence that breeding females are most likely to be immi-grants, resulting in low relatedness between mates and reduc-ing the potential for inbreeding. On the other hand, femalephilopatry appears to increase the likelihood of inbreedingwithin packs (e.g., QUA, WOL) or between packs (as a resultof EPCs, as in the case of the GAR subordinate female).Our genetic results also support behavioral observations

that only one female usually breeds per pack (Sillero-Zubiriet al. 2004). Only 1 of 7 multifemale packs in our study hada subordinate female breeder (GAR pack); she reared a singlepup in the same den as 3 pups produced by the dominantfemale. Behavioral observations suggest that the subordinatefemale in MEG pack at least attempted to raise pups in thesame den as the dominant female, although maternity of thelone pup analyzed in this study was unresolved. Polyandry andEPCs were not uncommon, with multiple paternity and EPPoccurring in 33% and 50% of litters, respectively, and EPPaccounted for 28% of pups. Similar results were found ina study of red foxes (Vulpes vulpes), which demonstrated highlevels of polygynous and polyandrous matings within andamong social groups (Baker et al. 2004). However, the levelsof multiple paternity and EPP in Ethiopian wolves and redfoxes are higher than those reported in other canids (Girmanet al. 1997; Roemer et al. 2001). As expected from behavioralobservations of EPCs in Ethiopian wolves (Sillero-Zubiri,Gottelli, and Macdonald 1996), all EPPs were obtained bymales in neighboring packs in this study, including at leastone EPP obtained by a subordinate male.We found no evidence that EPCs reduce inbreeding in this

population. There was a trend toward higher (not lower) re-latedness with extrapack sires, although sample sizes were toosmall for reliable statistical tests. However, these results at leastprovide evidence that females cannot or do not consistentlychoose unrelated males during EPCs. Furthermore, althoughbreeding females were significantly more related to maleswithin their pack than to a random distribution of males,there was no difference in breeding female relatedness tomales in her pack and males in neighboring packs. BecauseEPPs are restricted to males in neighboring packs, EPCs donot appear to increase the probability of producing outbredoffspring. Thus, females may seek EPCs for other reasons suchas infanticide avoidance (Wolff and Macdonald 2004, but seebelow).




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Near-zero FIS values imply that inbreeding within subpopu-lations is neither promoted nor minimized by dispersal andmating behavior (i.e., mating is random). Similar findingshave been reported in other social species such as alpinemarmots (Marmota marmota, Goossens et al. 2001) and black-tailed prairie dogs (Cynomys ludovicianus, Foltz and Hoogland1983). Under such conditions, the number of incestuous mat-ings observed (4 of 15 mated pairs in this study) should re-flect that predicted by chance in a population of this size.However, inbreeding may be higher than it was historicallybecause population sizes have decreased from historic levels(Gottelli et al. 2004). Furthermore, outbreaks of infectiousdisease (Sillero-Zubiri, King, and Macdonald 1996; Randallet al. 2004), which disrupt social cohesion and appear tomediate male dispersal, may increase the chances of inbreed-ing in a population, for instance, if females are more phil-opatric or disperse shorter distances, males disperse moreoften, or both. For example, Sugg et al. (1996) demonstratedthat inbreeding increased in black-tailed prairie dogs whennatal dispersal of the normally philopatric sex was increasedas this increased the probability of mating between close kin.Given the frequency of disturbances in this population asa result of recurring disease outbreaks (reviewed in Randallet al. 2006), it is also conceivable that EPCs may have evolvedas an inbreeding avoidance strategy but are no longer func-tioning as such. In particular, demographic events andchanges in dispersal behavior may have resulted in neighbor-ing packs and breeding pairs being more closely related thanthey would have been historically. This may explain the pat-tern of high relatedness between neighboring packs in WVbut not SN as WV has had higher number of pack disruptionsand/or extinctions in 16 years of monitoring (1988–2004),primarily due to disease outbreaks (Sillero-Zubiri 1994;Randall et al. 2004; Marino et al. 2006). In summary, due torecent demographic changes, the level of inbreeding re-corded in this study may be higher than is typical for thispopulation historically.A fair amount of research has been devoted to understand-

ing variation in mating systems and social structure amongcanids (Bekoff et al. 1981; Moehlman 1989; Geffen et al.1996). Often empirical data demonstrate that social structureresults from a number of selective pressures operating simul-taneously and teasing apart the various processes can be dif-ficult. Previous studies suggested that group formation andgroup augmentation in Ethiopian wolves were mediated by1) the low costs of tolerating relatives when resources areabundant and 2) retention or acquisition of high-quality ter-ritories (Sillero-Zubiri and Gottelli 1995b; Sillero-Zubiri andMacdonald 1998; Marino 2003). We show that a reduction ininbreeding may be an adaptive advantage of female-biaseddispersal and mating behavior. However, inbreeding occursby chance when dispersal and mating are random as we ob-served among Ethiopian wolves in this study. Genetic datafrom additional breeding seasons and populations are beingcollected as part of the long-term monitoring plan for Ethio-pian wolves in the Bale Mountains and elsewhere in Ethiopiato enable a more accurate assessment of the level of inbreed-ing in this species. Given its potential deleterious effectson fitness and long-term population viability (Hedrick andKalinowski 2000), inbreeding among Ethiopian wolves shouldbe monitored regularly, especially if population sizes fluctuateor are further reduced overall.

SUPPLEMENTARY MATERIAL

Supplementary Figure S1 and Tables S1 and S2 can be foundat http://www.beheco.oxfordjournals.org/.

We thank the Wildlife Conservation Department, the Oromiya RuralLand and Natural Resources Administration Authority, and the BaleMountains National Park for permission to undertake research inEthiopia. We are grateful for the support of the Ethiopian Wolf Con-servation Programme for their assistance in the field and to JorgelinaMarino, Claudio Sillero-Zubiri, and Hannah Dugdale for commentson an earlier draft of the manuscript. Thanks also to Toby Dunn forassistance with the SAS code for the randomizations. Fieldwork wasfunded by the Iris Darnton Trust, Jesus College, and the Sophie Dan-forth Conservation Biology Fund (Rhode Island Park Zoo). D.A.R. wassupported by an Overseas Student Award (UK) and a Clarendon FundAward from the University of Oxford.

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