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Complex segregation analysis of Carabelli's trait in a melanesian population.

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AMERICAN JOURNAL OF PHYSICAL ANTHROPOLOGY 53:301-308 (1980)
Complex Segregation Analysis of Carabelli's Trait in a
Melanesian Population
DONALD KOLAKOWSKI, EDWARD F. HARRIS, AND HOWARD L. BAILIT
Departments of Behioral Sciences and Community Health (D.K.,
HLB.) and
Orthodortties (EfiH.)),School of Dental Medicine, Uniuersity of Co-ticut
Health
Center, Fartnington, Connecticut 06032
KEY WORDS
Carabelli's trait, Dental genetics, Segregation analysis
ABSTRACT
A complex segregation analysis was performed on Carabelli's
trait on the upper first molar utilizing 358 nuclear families from the Solomon
Islands of Bougainville and Malaita. Simultaneous estimation of three sources
of variation by the method of maximum likelihood demonstrates a significant
effect of shared sibling environment which accounts for over 1Wo of the variance
in liability for the trait. In addition, a statistically significant major gene
influence is discussed and suggestions for quantifying individual liability levels
for this and other dental traits are outlined.
Carabelli's trait is an accessory feature occurring with varying frequency on the mesiolingual surface of upper molars in modern
man, fossil hominids, and anthropoid apes. In
its pronounced form it constitutes a prominent
tubercle with a nearly circular cross section
and a free apex that may occasionally attain
the height of the four main molar cusps. When
the distinct cusp form is absent, diminutive
cusp forms, or one of a range of furrows or
even a pit, can be found at the site. There is
general agreement among investigators that
the intra-individual incidence of these features is such that they are all manifestations
of the same trait (Dietz, '44; Kraus, '51, '59;
Dahlberg, '55; Turner, '67). The etiology, variation in expression, phylogenetic record, and
racial distribution of Carabelli's trait form the
bases of a surprisingly large body of literature,
much of which is reviewed in the excellent
works of Jergensen ('561, Moorrees ('571, and
Korenhof ('60).
The trait is very old phylogenetically (Batujeff, 1896 Windle and Humphreys, 1887;
Frisch, '65; Saheki, '66; Haug, '77). In modern
humans it varies across races in first molar
prevalence from a low of 7-13% in Asian
peoples (Suzuki and Sakai, '57) to a high of
80% in Caucasians (Goose and Lee, '71; Alvesalo e t al., '75). The prevalence is much
reduced in second and third molars (Carabelli,
1842; Hjelmman, '29; Dahlberg, '45, '71; Kirveskari, '741, and the degree of bilateral symmetry of trait expression varies considerably
0002-948318015302-0301$01.70 0 1980 ALAN R. LISS, INC.
across populations ('buji, '58; Kraus, '59 Biggerstaff, '73; Harris, '77). There is no evidence
of sexual dimorphism (Garn et al., '66a; Bang
and Hasund, '72; Scott, '73), and the trait
appears to be independent of molar size and
morphology (Garn et al., '66b).
In 1902, G.V. Black claimed that Carabelli's
cusp "is hereditary, appearing regularly in
teeth of children when present in the teeth of
both parents. It also occurs in a modified form,
when present in but one parent." We know
very little more today about the mode of
inheritance of the trait than did Black. Studies
of concordance rates in twins and in siblings
have produced a range of results, prompting
the authors to draw conflicting conclusions
(Korkhaus, '30; Meredith and Hixon, '54; muji, '58; Garn et al., '66b; Biggerstaff, '73; Alvesalo et al., '75). Several investigators have
erroneously ignored familial relationships and
tried to fit prevalence statistics for the forms
of trait expression to the binomial distribution
(see Sofaer, '70). Kraus ('51) provided the first
systematic attempt a t analyzing the segregation of the trait in families. While his often
cited (Krogman, '60; Goose, '74; McKusick,
'75) conclusion of a possible major locus effect
may yet prove true, more recent data have
exhibited inconsistencies in the assumed mode
of inheritance ("suji, '58; Kraus, '59; Goose
and Lee, '71; Lee and Goose, '72).
W i v e d DeQmber 5,1979aceepted March 31,1980.
30 1
D. KOLAKOWSKI, E.F. HARRIS,AND H.L. BAKIT
302
The interracial, interindividual, and bilateral variability in both the prevalence and
expression of Carabelli's trait suggest that, in
addition to familial relationships, genetic
studies should allow for stochastic fluctuations, developmental events, and environmental sources of variation. Simultaneous estimation of certain effects is also essential if
any intercorrelations or confounding interactions are to be expected. Moreover, the observed pattern of variability (see Discussion)
indicates that Carabelli's trait may well be a
threshold character (Gruneberg, '52; Falconer,
'65), in which case the above influences would
more properly be reflected in an underlying
continuous distribution of "liability" than they
would in the particular manifestation on a
given tooth.
In this paper we attempt a rather sophisticated analysis of familial data on Carabelli's
trait. We utilize one of a number of quantitative genetic models and their associated computer programs, which have become available
only in the last decade and which constitute
an impressive conceptual and technological
advance in human genetic analysis. It simultaneously takes into account information on
trait incidence in first-degree relatives, various assumptions about genetic transmission,
the distribution of liability in the population,
a specific environmental influence, and random variation (Morton and MacLean, '74).
METHODS
al., '71); d) These groups exhibit a very low
prevalence of caries and tooth loss (Bailit et
al., '68).
Trait evaluation
Carabelli's trait was classified on all scorable upper first permanent molars (UM1) using
the eight-grade scale of Scott ('731, all observations being made by one person (EFH).
Provenience, sex, and familial relationship
were unknown to the observer, and the examination sequence was randomized for these
variables. Intra-observer reliability, assessed
as percent concordance for presence and grade
from two unrelated scoring sessions, is 97%
(n = 168 teeth). This agrees favorably with
other reported estimates (Sofaer et al., '72;
Scott, '73). Summary data by island, village,
sex, tooth, and grade are given in Harris ('77).
The classes were subsequently collapsed into
three groups because of the constraints of the
chosen analysis. The categories are: 0) Absence of the trait. 1) The variety of negative
features constituting a pit, a furrow, two furrows, or a pit in combination with a furrow;
and slight tubercles demarcated by a lingual
bulge outlined by a nearly continuous semicircular groove but lacking a free apex. 2) A
pronounced tubercle, distinguished by a free
apex and a well defined furrow setting off the
cusp of Carabelli from the protocone.
Shapiro ('49)and others (e.g., Meredith and
Hixon, '54) have argued for comparability in
the criteria used to grade Carabelli's trait.
The above trichotomy conforms to a grouping
of the often used standards of Dahlberg ('55,
'63) and to the hypothetical "genotypes" employed by Kraus ('51). Data for the six language groups were pooled, and sexes were not
distinguished in the analysis, since the trait
is not dimorphic. To avert questions of how to
treat asymmetry, only the score for the left
maxillary first molar was used.
Data base
The sample consists of dental casts for 1,208
individuals collected from six language groups
on the islands of Bougainville and Malaita in
the Solomon chain (Damon, '73; Friedlaender,
'75). Several characteristics of these groups
make them particularly suitable for the present investigation: a) The usual problems associated with defining the breeding populaAnalytic model
tion, its boundaries, and a random sample are
eliminated by their isolation on the basis of
The complex segregation analysis of Morton
spoken language and by virtually total popu- and MacLean ('74) is particularly well suited
lation sampling; b) The high level of endoga- to the problem a t hand because 1) it allows
my, cultural homogeneity, and similar stan- for the familial analysis of a trichotomized
dards of living should reduce environmental trait, which provides more information than
covariation between relatives to minimal lev- the usual presence-absence dichotomy; 2) it
els for human groups; c) Virtually all individ- simultaneously estimates by the method of
uals in the sample are members of identified maximum likelihood the effects of both a mafamily units, the pedigrees having been con- jor locus and of polygenic heritability, which
structed and verified in previous expeditions provides a conservative test of the former
by cultural anthropologists (Damon, '73). Af- (MacLean et al., '75); and 3) it also (simultafirmation of biologic parentage was effected neously) estimates the effects of common sibusing six serologic systems (Friedlaender et ling environment, which, in the form of nutri-
SEGREGATION ANALYSIS OF CARABELLI'S TRAIT
tion and pre- and postnatal maternal effects,
is likely to be the most significant environmental influence on tooth morphogenesis
(Bader, '65; Sciulli et al., '79).
The genetic model assumes an underlying
continuous distribution linearly related to liability for the trait in question, which we
assume without loss of generality to have
mean u = 0 and variance V = 1.0. It also
assumes that there are three types of individuals in the population corresponding to the
three genotypes aa, Aa, and AA associated
with the two alleles, a and A, a t one autosomal
locus. 'Jkait liability of each individual comes
from one of three normal distributions with a
common variance and means paa,pAa,and
pAA,
in order of increasing levels, respectively.
The variability within genotypes is assumed
to arise from additive polygenic sources,
shared sibling environment, and random environmental effects. The parameters included
in the likelihood equation are listed in Table
1, with their definitions and additional assumptions (Morton and MacLean, '74; also see
Elston, '79). The general model therefore requires the estimation of the latter five parameters. Certain parameters may then be fixed
to create subhypotheses under which only the
remainder are estimated. For each hypothesis
we calculate - 2 In L + c, where In L is the
log-likelihood of the sample and c is a (computer programming) constant. If - 2 In L,
c is the value when m + k parameters are
estimated, and - 2 In L, + c when only m of
the m k parameters are estimated, then (2 In L, + c) - (- 2 In L, + c) = 2 In (L,/L,) is
asymptotically distributed as chi-square on k
degrees of freedom, testing a null hypothesis
on these k parameters (the likelihood ratio
test).
303
ronment, polygenic heritability being sufficient). All restrictions on the general model
were rejected. The data are consistent with
the hypothesis that the expression of Carabelli's trait is influenced by both a major gene
(~'3
= 1485.20 - 1469.38 = 15.82; p<.OOl)
and common sibling environment (x2, =
1481.06 - 1469.38 = 11.68; p<.OOl), virtually
all of the genetic variation being attributable
to the major locus (H = .018,whereas variance
of the major locus = .339; Morton and MacLean, '74). Moreover, if we were t o ignore the
major locus, sibling environmental resemblance would still be significant (xzl = 1489.38
- 1485.20 = 4.18; p<.05).
As an additional check on the validity of
the results, we tested for homogeneity over
mating types. Due to limited numbers of pronounced-cusp (2) individuals in the parental
generation, and the relatively low power to be
expected in the case of a trichotomized dependent variable, i t was not possible to analyze matings in any interpretable configuration. Rather, we had to pool the matings
among the four phenotypes into the two
groups specified in Table 3, with sample sizes
524 in 161 families and 628 in 197 families,
respectively. Therefore, while homogeneity between groups would tend to strengthen our
results, heterogeneity would be fairly uninformative. Separate analyses for the two groups
of matings are presented in Table 3. In the
first group the major locus hypothesis falls
somewhat short of significance (xZ3= 676.99
- 670.40 = 6.59; .05<p< .1) and B = 0 is
borderline (xZ1 = 674.13 - 670.40 = 3.73;
pc.055); whereas, in the second group both
hypotheses are confirmed (x23 = 807.87 798.68 = 9.19; pC.03 and x21 = 805.12 798.68 = 6.44; p<.02, respectively). For all
models, the two groups of matings are homoRESULTS
geneous with respect to the estimated paramThe ratings of trait expressivity were tri- eters (Table 3). Given the smaller sample size
chotomized as described above, yielding the in the former group, we conclude that the
following population proportions: Pronounced environmental effect is valid but that the
(2) = .1104; Intermediate (1) = .2857; Absent major locus hypothesis needs further verifi(0) = .6039. In addition, siblings with one or cation.
both parents unsampled generated a fourth
DLSCUSSIOM
phenotype, Unknown (U). Sample size was
Previous
results
relating environmental ef1,152 rated subjects in 358 families. The results of the complex segregation analysis are fects to the dentition have been lacking in
presented in Table 2. Given that significant specificity (Biggerstaff, '79). While environgenetic variation has been amply demonstrat- mental stress has been discussed in relation
ed in the literature, the subhypotheses consid- to random asymmetries in tooth size, no defiered were: q=t=d=O (no major locus); B=O nite relationships have been found (Bailit et
(no sibling environmental resemblance); and al., '70; Doyle and Johnston, '77; DiBennardo
B=q=t=d=O (neither major locus nor envi- and Bailit, '78). Maternal health status during
+
+
D.KOLAKOWSKI, E.F. HARRIS,AND H.L. BAILIT
304
TABLE 1. Complex segregatdon analysis: Parameters in the likelihood
expression.
Parameter
Description
Mean value of the liability distribution in the popuhtion:
standardized at 0 for this analysis.
Population variance in liability; standardized at 1.0 for
this analysis.
Polygenic heritability: the proportion of the variance, V,
attributable to the additive effects of an indefinite
number of loci.
Proportion of the population variance, V, attributable to
the effects of environment common to siblings.
Gene frequency of allele A at the major locus. Allele a
has frequency (1-3.
The displacement in mean liability levels between
homozygotes at the major locus: i.e., the difference pAApa, (measured in standard deviation units of 1.0).
The degree of dominance of the major locus; i.e., the
proportionate distance (pAa- pas) / (pa*- pas).
U
V
H
B
9
t
d
The model also assumes Mendelian transmission probabilities between generations;
genotypic frequencies in Hardy-Weinberg proportions; mutual independence of the
major locus, polygenic and sibling environmental effects; and the absence of all other
nonrandom sources of variation, such as environment common to parents and children
and polygenic dominance.
T U L E 2. Complex seeregation a d y s i s of Cambelli's tmit on UMl.
-2 I h L
Hypothesis
General
q=t=d=O
B=O
B
=
q
=
t
=
d
=
0
+C
1,469.38
1,485.20
1,481.06
1,489.38
Hi
B
9
,018 -+ .08
,321
,194 f .05
,151
.191
0
0
,031 f .02
0
,074
.449
t
3.99
?
d
.86
,598 e .16
0
0
2.85
,584
0
0
0
Symbols are d e h e d in Table 1.
TABLE 3 . Complex segregation annlysis of mating types for Carabelli's tmit on U M l .
Mating types*
Hypothesis
2 x 2, 1 x 2, 2 x U
0 x 2,1 x 1,0 x 1
UXl,UxO,OxU
general
q =t =d
2 x 1, U x 2, 2 x 0
1 x 0 , l x U, 0 x 0
u x u
general
q =t =d
B=O
=
0
B=O
=
Heterogeneity over mating types:
0
+c
H
B
9
t
d
670.40
676.99
674.13
.004 f .I9
,378
,220
.197 -+ .12
.lo7
0
.029 e .02
3.51 -c 1.1
0
3.63
.644 e .32
798.68
807.87
805.12
,030 f .ll
.271
,158
,147 e .34
.193
,027
3.58 2 .42
0
3.57
,837 e .14
0
,523
- 2 In L
0
0
,026
f .01
0
,046
for general, e5= 1469.38 - (670.40 + 798.68) = 0.30; n.s.
for B = 0, x24 = 1481.06 - (674.13 + 805.12) = 1.81; n.s.
for q = t = d = 0, xz2 = 1485.20 - (676.99 + 807.87) = 0.34; n.8.
*Order of mating is paternal x maternal; symbols are defined in the text.
0
,637
SEGmGATION ANALYSIS OF CARABELLI'S TRAm
pregnancy has also exhibited a relationship to
tooth size (Garn et al., '79), but this effect
necessarily confounds environment with maternal and offspringgenotypes; also, no dental
data were available for the mothers. %in
studies of heritability and/or concordance for
dental traits have as their only alternative a
residual random component attributed to environment by default. Moreover, the developmental resilience exhibited in dental morphogenesis (e.g., Garn et al., '65a, b) has tended
to generate an image of teeth as essentially
genetic entities, which is not realistic.
Therefore, the finding of significant environmental effects on Carabelli's trait is a welcome
addition to the known characteristics of this
trait and of tooth morphology in general. In
terms of the complex segregation model, siblings are significantly more similar to each
other in degree of trait expression than they
are to their parents, and this effect alone
accounts for 19%0 of the variance in liability.
This result is much stronger than the residual
random component of a heritability study,
interpretation here being limited to maternal
and familial effects.
With regard to the possibility of a major
locus, we note that, due to the low frequency
of individuals with pronounced cusps in the
parent generation, six language groups from
the islands of Bougainville and Malaita have
been pooled for the present analyses. This
poses a substantial risk of violation of the
Hardy-Weinberg equilibrium assumption, even
to the point of fixation in certain groups. It is
possible that the present 6% incidence attributable to the major locus overall corresponds
to a much larger effect in a subset of the
groups, with absence of the relevant allele in
the remainder. Given that the analysis is
conservative with respect to the detection of
a major locus, the ability to obtain significance
with half the sample and to discriminate between the two halves may indicate adequate
statistical power and the possibility that
something is indeed there. Inasmuch as majorgene influence on tooth morphology would
have substantial genetic, anthropologic, and
phylogenetic significance, this hypothesis is
being investigated further.
Finally, a much more powerful segregation
analysis of a morphological trait could be done
if one could identify a continuous variable
that might be functionally related to liability
for the trait. In a developmental context, we
know that the Carabelli trait has a dentinal
precursor and is not just an enamel feature
305
(Korenhof, '60).If present, the trait is mineralized near the end of amelogenesis, after the
occlusal morphology is finalized but before
cessation of enamel formation at the dentinoenamel junction (Butler, '56 Kraus and Jordan, '65).Thus, it is a temporal phenomenon.
However, initiation of Carabelli's trait cannot
be mediated by a global regulator such as
growth hormone levels because it occurs on
both deciduous (Jergensen, '56; Tsuji, '58; Butler, '71; Joshi, '75; Hanihara, '76) and permanent molars, which have a mean developmental spacing of over nine years (Garn and
Polachek, '59). Nor can it simply be the result
of a timely perturbation in the availability of
structural components of the extra-cellular
matrix, because it is site-specifk and its occurrence is correlated across antimeres and
within the molar field, as well as with the
occurrence of its phylogenetic analog on the
mandibular molars: the protostylid (Scott,
'73). Further, there is no element of necessity
in its pattern of occurrence either bilaterally
or in the mesial direction: We have found a
present-absent side difference in 38% of Melanesians expressing the trait, and Tsuji ('58)
reports that 11.m of his Japanese sample
exhibit the trait on UM', while absent on dm2.
The initiation of crown formation is now
believed to be neurotrophic, resulting from the
innervation of the mesenchyme by the trigeminal fifth ganglion (Kollar and Lumsden,
'79). We may therefore hypothesize that a
Carabelli (Iprotostylid) -initiating message
originates very early under neural sensory
control in a temporally ordered relationship to
other instructions for crown formation, and
that such a message will be better synchronized bilaterally (and between arcades) than
are local developmentalprocesses in non-nervous tissue (Garn et al., '67; Van Valen, '70).
Physiologic variation in local growth rates
and consequent shifts in tissue maturation
status relative to the timing of the neural
message could, therefore, affect trait penetrance and expression and give rise to the bilateral variation (and low association with the
protostylid) that we observe. Similarly, the
great temporal spacing in molar field fonnation may often result in tissue maturation
getting out of synchrony with timing of the
neural message, with a subsequent loss of
trait expression and the reduced incidence,
increased variability, and decreased bilateral
symmetry we observe in the posterior teeth.
Such shifts of local maturation status are
observable radiographically (Moorrees et al.,
306
D. KOLAKOWSKI, E.F. HARRIS, AND H.L. BAILIT
’63; Demirjian et al., ’73; Anderson et al., ’76)
and can be assessed relative to an antimere,
to earlier-forming teeth, or to overall mineralization. These shifts might also be reflected
in the eruption time of the relevant teeth,
albeit to a lesser extent (Garn et al., ’57).
Ultimately, these ontogenetic timing lapses
might be associated with early or late overall
maturation status, as reflected in total eruption patterns or skeletal age, although, again,
correlations between tooth mineralization and
eruption patterns or bone age range no higher
than .46 (Cattell, ’28a, b; Green, ’61; Anderson
et al., ’75).
Therefore, we would propose “crown mineralization age,” dental age, and skeletal age as
univariate continua potentially related to liability for Carabelli’s trait (and/or the protostylid). Such inferred or latent variables might
most effectively be estimated on a n interval
scale by multivariate probit analysis from
ordinal-scale ratings of mineral deposition
made on dental or skeletal radiographs (Murray et al., ’711, from longitudinal tooth eruption time data, or from cross-sectional data on
patterns of tooth eruption (Kolakowski and
Bock, ’80). The validity of the proposed relationships to liability could be tested across
subjects utilizing data on early and late molar
crown formatioderuption, and within subjects
by examining the antimeric symmetry of
amelogenesis/eruption time in conjunction
with ratings of trait expression. Positive results for any single paradigm would yield a
temporal index functionally related to liability. Such results for multiple paradigms would
permit construction of a composite index for
which the relative discriminatory power of
individual dentakkeletal ratings would be a
byproduct of the probit analysis. The liability
index could then be used to perform more
powerful segregation analyses, to estimate a n
optimal value for the hypothesized neural
message and/or make a determination as to
whether or not there is but a single message
(i.e., present-absent as suggested by Alvesalo
et al., ’751, to determine whether or not current ratings for degree of trait expression
constitute a biologically meaningful ordinal
scale, and to investigate the possible neurotrophic identity of Carabelli’s trait and the
protostylid.
ACKNOWI3DGMEWIS
This research was supported in part by the
National Institute of Dental Research, NIH,
awards No. R01-DE04082, K04-DE00015, and
T32-DE07047; and by a grant to the Population Genetics Laboratory, University of Hawaii, a World Health Organization Collaborating Center for Reference in Processing of
Human Genetics Data. We are also indebted
to the staff of the Peabody Museum, Harvard
University, and especially to Dr. Jonathan S.
Friedlaender for making all the Solomon Island data available to us, and to Dr. Newton
E. Morton for his sponsorship and helpful
comments. Useful discussions with Dr. Edward J. Kollar and the programming assistance of Gerald Takase is gratefully acknowledged.
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