Forest Ecology and Management, 61 ( 1993) 17-28
17
Elsevier Science Publishers B.V., Amsterdam
Genetic variation in susceptibility to windthrow
in young Douglas-fir
Roy R. Silen•, Donald L. Olson and John C. Weber
Pacific Northwest Research Station, Forestry Sciences Laboratory, USDA Forest Service,
3200SWJefferson Way, Corvallis, OR 97331, USA
(Accepted 3 May 1993)
Abstract
The genetic component in susceptibility to windthrow was estimated among F2 full-sib families of
Douglas-fir (Pseudotsuga menziesii (Mirb.) Franco) at an experimental forest site in western Oregon
that was severely damaged by a storm on 7 January 1990. Progeny from half-diallel and factorial
crosses had been planted in a randomized block design in two sets ofthree replications, one from seed
sown in 1983 and another from seed sown in 1984. The proportion oftrees within five-tree row family
plots that were wind-thrown (PROP) and the proportion adjusted to remove the effect of mean plot
height at 5 years (ADJPROP) were analyzed, the latter to test whether susceptibility to windthrow
was primarily a function of wind resistance estimated from relative tree size. Trees in the 1983 set
were 7 years old, averaging 2.1 min height, and were produced from 87 crosses among 19 parent trees
from four relatively low elevation seed sources in western Oregon and Washington. Trees in the 1984
set were 6 years old, averaging 1.6 m tall, and were produced from 51 crosses among 13 parent trees
from one relatively high elevation seed source in western Oregon. Susceptibility or resistance to
windthrow among parents was indicated by significant differences in proportion of windthrow in
their progeny. In both sets, there was significant genetic variation in PROP and ADJPROP. The esti­
mated total genetic component in the 1983 set, the sum of general combining ability ( GCA) plus
specific combining ability (SCA), accounted for 21.4% of the total variance in PROP and 18.6% of
the total variance in ADJPROP ( GCA percentage nearly equal to SCA percentage for both PROP
and ADJPROP). In the 1984 set, GCA and SCA together accounted for 33.8% of the total variance
in PROP (GCA percentage nearly equal to SCA percentage) and 22.6% of the total variance in
ADJPROP. Results indicated that susceptibility to windthrow differed by family, and was signifi­
cantly related to family height, but that height accounted for Jess than a third ofthe genetic component
of variation. These data suggest that breeding for resistance to windthrow in Douglas-fir could be
successful but would be expensive and probably reduce genetic gains in growth per generation.
Introduction
Genetic studies of windthrow are rare, unplanned, and unwelcome. On 7
January 1990, a storm of an intensity expected once in 20-25 years centered
*Corresponding author.
18
R.R. Silen eta/. I Forest EcologyandManagement61 (1993) 17-28
on the forests of Oregon. An estimated 350-400 million board feet of trees
were wind-thrown on Oregon National Forests alone (S. Paulson, USFS Re­
gion 6, personal communication, 1991 ) . Local gusts exceeded 161 km ( 100
miles) per hour ( G .H. Taylor, OSU meteorologist, personal communication,
1991 ) . Our genetic test site in the Willamette Valley was in the main path of
the storm.
Coastal Douglas-fir (Pseudotsuga menziesii (Mirb.) Franco var. menzie­
sii) ranks among the most windthrow resistant of commercial conifers (de
Champs et al., 1982). Yet during each rotation, cumulative windthrow causes
substantial losses on most forest ownerships, ranking with pests and fire as a
major disruption to orderly sustained-yield harvest. Windthrow is greatest on
sites with poorly drained soils (Kennedy, 1974 ), in over-dense stands (Faber
and Sissingh, 197 5), or where winds are concentrated by topography. Genetic
variation has been found in susceptibility to windthrow in several coniferous
species (Bell et al., 1983; Liegel, 1984; Telewski and Mordecai, 1986), but we
have found no published reports of family-related variation in this trait for
Douglas-fir.
Information on windthrow has come mainly from studies of natural catas­
trophes, as in this investigation of family genetic variation. Luck is needed.
Statistics become weak if only a few replications of a damaged genetic study
are in the wind-thrown area. Moreover, family differences become undetect­
able wherever windthrow is either too extreme or too light.
Windthrow from the storm on 7 January 1990 was generally light over the
20 ha of experimental plantings at our genetic test site near Monmouth, OR
(Table 1 ). Moderate damage occurred at a few stand comers that funnelled
Table 1
Studies damaged by the 7 January 1990 storm at a genetic test site near Monmouth, OR 1
Study age
(years)
Design2
No. of
parents
No. of
families
Mean
height
(m)
Trees windthrown/
total
Ratio of
most:least
wind-thrown
trees per parent
19
15
15
10
7
6
PF,WP
D,FS
F,FS
D&F,FS
D&F,FS
D&F,FS
D&F,FS
26
6
6
10
19
18
13
26
30
30
40
64
87
51
18.0
10.2
9.6
5.1
3.3
2.1
1.6
51/936
37/840
36/840
64/1120
54/1032
1088/2376
400/1428
9.0:1
3.3 3 :1
3.34 :1
1.8: I
12.0:1
2.3:1
2.5: I
5
1
The last two lines apply to the 1983 and 1984 data sets reported in this paper. Photograph in Fig. 1 were taken in the first two studies. 2
PF, paired families; D, diallel; F, factorial; FS, full-siblings; WP, wind-pollinated. 3
Count of wind-thrown trees for female plus male families divided by 2. 4
Count ofwind-thrown trees for female parent. R.R. Silen eta/. I Forest Ecology and Management 61 (1993) 17-28
19
winds into limited areas having shallow soils and restricted drainage. Five
studies with trees up to 19 m tall suffered light to moderate windthrow that
appeared to be family related (Fig. 1 ), but the overall percentage damage to
trees in these five studies was too small for a statistically adequate test to be
carried out (Table 1 ) .
Extensive damage occurred on two experimental plantings, however. On a
north-south oriented ridge of the site (elevation approximately 90 m), trees
were wind-thrown in a swathe 200 m long, down a well-drained, gentle lee
slope eastward from where the ridge line dipped slightly, funnelling winds as
described by Gratkowski ( 1956 ). Windthrow, somewhat more dense at the
center of the wide swath, extended beyond the borders of some unusual ex­
perimental material-two different sets of young full-sib families resulting
Fig. 1. Family rows of leaning trees from a 7 January 1990 storm are shown amid undamaged
family rows. Seven studies on this site had such family-related damage (Table 1 ). Photograph
(A) shows a 19-year-old study ( 18.0 m mean height) , and photograph (B) shows a 15-year-old
study (9.6 m mean height), each with three of four trees in the family row selectively wind­
thrown. In only two of the seven studies, with tree ages of 7 and 6 years (referred to as the 1983
and 1984 sets), was the extent of the damage adequate for statistical analysis.
20
R.R. Silen eta/. I Forest Ecology and Management 61 (1993) 17-28
from crosses among F1 open-pollinated progeny from the 1912 Douglas-fir
Heredity Study (Munger and Morris, 1936). These 138 families of the F2
generation had been established to investigate inheritance of growth-related
traits (general and specific combining ability) among top parent trees of the
1912 study. While straightening the 1-3m tall wind-thrown trees (i.e. trees
leaning and prostrate), we noted that entire family-row plots of several crosses
were wind-thrown among adjacent undamaged rows of other parentage. This
situation presented an opportunity to investigate the inheritance of suscepti­
bility to windthrow at the family level in Douglas-fir. However, it also pre­
sented the problems of analyzing an unplanned random event superimposed
on one-quarter of the replications of the original study.
Our objectives were to determine if there was a genetic component to sus­
ceptibility to windthrow among the full-sib families (de Champs et al., 1982),
and if susceptibility to windthrow was related to tree height as an estimate of
relative resistance to wind flow. A strong relation between inherent growth
and windthrow would be an important consideration in Douglas-fir genetic
improvement programs designed to increase height growth on wind-prone
sites. Also discussed are potential gains and problems of breeding for resis­
tance to windthrow in Douglas-fir.
Methods and materials
The Douglas-fir Heredity Study (described by Munger and Morris, 1936)
included selections of 120 individual parent trees field-selected in 1912 from
13 seed sources in western Oregon and Washington. In 1915 the open-polli­
nated progeny were planted as families at five sites. After growing for 50 years,
the tallest individual progenies of outstanding parents were selected from low
and high elevation seed sources at the two highest elevation sites. Cuttings
from these progeny selections were grafted onto rootstocks, and many were
controlled pollinated in factorials and half-diallel matings as the grafted trees
flowered. We grew the resulting full-sib seedlings in raised beds in our exper­
imental nursery at the Forestry Sciences Laboratory, USDA, Forest Service,
Pacific Northwest Research Station, Corvallis, OR. Owing to the large num­
ber of crosses, families were allocated to one of three different yearly geo­
graphic sets with no parents in common (labeled as 1982, 1983, and 1984,
indicating their year of sowing). Each set was planted at four sites in western
Washington and Oregon.
The storm of 7 January 1990 seriously damaged the 1983 and 1984 sets of
the three contiguous sets planted at our snow-free Willamette Valley experi­
mental site; the other nine sets, three at each of three mountain sites, were
protected under deep snow. Most of the damage at the Willamette Valley site
occurred in the 1983 set, and the denser windthrow may have provided a less
sensitive test offamily differences than the less damaged 1984 set. The 1983
R.R. Silen eta/. I Forest Ecology and Management 61 (1993) 17-28
21
set included progeny from 87 crosses (57 single crosses and 15 pairs of recip­
rocal crosses) involving 19 parent trees selected from two low elevation seed
sources in western Oregon (Gates, elevation approximately 300 m; Palmer,
elevation approximately 600-900 m) and two seed sources in western Wash­
ington (Darington, elevation approximately 180m; Wind River, elevation
approximately 360 m). In the 1984 set, parentage included progeny from 51
crosses ( 49 single and one reciprocal) involving 13 parent trees selected from
one high elevation seed source in western Oregon ( Santiam, elevation ap­
proximately 900-1200 m). Crosses are summarized in Tables 2 and 3.
In each set, the randomized complete block design consisted of three repli­
cations of progeny from each cross planted at 3 mx3 m spacing in five-tree
row plots to allow within-family selection in the next generation ( 1983 set
with 2-1 seedlings, 1984 set with 2-0 seedlings). Each set was surrounded by
a row of buffer trees at the same spacing. Plantings were on tilled agricultural
soil, with good early development and 98% and 88% first year survival of the
two sets, respectively. By January 1990, the 1983 set averaged 2.09 m and the
1984 set 1.63 min height, based on a 99 tree and 65 tree sample, respectively.
These early growth rates are above regional averages of Douglas-fir progeny
tests, and trees in these sets were taller than in the same sets planted at the
other three sites. Excavation of six wind-thrown trees showed normal devel­
opment, with roots extending beyond crown width. Although the swathe of
windthrow was somewhat concentrated at its center, it extended well beyond
the set boundaries. Storm gusts lasted several hours, and wind directions
shifted as it passed. With the 32 parents in the two sets, each represented by
an average of nine crosses randomized in each ofthree replications, the chance
of any parent family having a non-random exposure to wind or soil differ­
ences was very low.
Trees were visually scored as either erect or wind-thrown ( 10° from verti­
cal). We analyzed variation in two variables: the proportion of wind-thrown
trees within plots (PROP) transformed by using the Freeman-Tukey arcsine
transformation of binomial proportions (Mosteller and Youtz, 1961 ) , and
PROP adjusted for mean plot height (ADJPROP). Rather than remeasure
heights after the storm, we used the recently measured heights at age 5
(HEIGHTS) on both sets as the covariate. We assumed that HEIGHT 5 and
PROP family mean heights would be highly related to 6 and 7 year family
means with little effect on inferences. The average (and range) ofHEIGHT5
was 87.2 em (50.5-122.4 em) in the 1983 set (n=261 plot means), and 112.9
em (44.5-162.5 em) in the 1984 set (n=153 plot means). The PROP was
regressed directly on HEIGHTS by using simple linear regression, and the
regression coefficient (b) was used to adjust for height:
ADJPROP=PROP-b(HEIGHT5-set mean HEIGHTS).
Variation in PROP and ADJPROP was partitioned into four sources by
ANOVA: general combining ability ( GCA, the additive genetic component);
N
Table2
Summary of controlled crosses 1 and observed percentages2 of wind-thrown trees in the 1983 set
Females
Males
Dl
Dl
D2
D3
D4
D6
G1
G2
G3
G4
G5
G6
G7
Parent
mean
percentage
1
N
D2
26.7
33.3
53.3
22.2
46.7
13.3
13.3
66.7
60.0
40.0
7.1
27.3
46.7
45.9
36.1
D3
D4
D5
46.7
53.3
23.1
22.2
67.7
33.3
26.7
28.6
13.3
6.7
26.7
33.3
50.0
6.7
Gl
60.0
66.7
66.7
G2
G3
G4
G5
G6
80.0
60.0
42.9
33.3
57.1
60.0
60.0
86.7
60.0
93.3
60.0
20.0
100.0
46.7
33.3
85.7
53.3
53.3
73.3
53.3
73.3
53.3
46.7
51.1
47.2
48.4
50.9
63.2
60.0
60.0
G7
PI
WI
W2
W3
W4
W5
50.0
66.7
66.7
20.0
46.7
76.9
40.0
40.0
53.2
41.3
46.7
26.7
53.3
40.0
55.6
20.0
13.3
86.7
--
42.9
40.0
33.3
60.0
6.7
46.7
33.3
6.7
66.7
40.0
45.9
36.1
33.3
27.6
42.4
48.8
51.1
47.2
48.4
50.9
63.2
37.7
35.4
40.0
45.8
86.7
00.0
9.1
20.0
46.7
46.7
Parent
mean
percentage
~
~
~
....
"'~
~
33.3
27.6
27.8
48.8
37.7
43.4
42.6
The 19 parents are from Darrington, WA (D), Gates, OR (G), Palmer, OR(P), and Wind River, WA (W) (Munger and Morris, 1936). The total
number of crosses is 87 (57 single plus 15 reciprocal pairs). Total sample size is 1188 trees (514 wind-thrown). Average sample size for a cross is 13.7
trees (sum of three replications).
2
0bserved percentage of wind-thrown trees of a cross and the female and male parent total percentages are calculated as the sum of wind-thrown trees
across three replications divided by the total number of trees for the appropriate crosses.
'~
~
~
~
c­
~
.,
....
"'
~
o'"'il
Ie:
~
~
,.....
~
~
R.R. Si/en eta/. I Forest Ecology and Management 61 (1993) 17-28
23
Table 3
Summary of controlled crosses 1 and percentage of wind-thrown trees in the 1984 set
Females
Males
Sl
Sl
S2
S3
S4
S5
S6
S7
S8
S9
SIO
Sll
Sl2
Sl3
Parent
mean
percentage
S2
46.7
S3
S4
S5
46.7
57.1
33.3
40.0
33.3
13.3
13.3
26.7
00.0
73.3
8.3
6.7
6.7
00.0
54.3
27.1
28.2
26.7
S7
84.6
46.7
20.0
26.7
S8
53.3
40.0
80.0
66.7
6.7
6.7
7.1
S6
23.2
13.3
80.0
45.5
46.7
22.5
21.6
27.6
SlO
46.2
40.0
20.0
53.3
23.1
20.0
40.0
13.3
33.3
6.7
13.3
6.7
13.3
22.2
S9
40.1
17.8
Sll
14.3
33.3
33.3
21.4
71.4
7.1
29.0
27.1
Parent
mean
percentage
54.3
27.1
28.2
23.2
22.5
21.6
27.6
40.1
17.8
29.0
27.1
6.7
00.0
28.0
'The 13 parents are from Upper Santiam (S) (Munger and Morris, 1936). The total number of crosses
is 51 (49 single and 1 reciprocal). Total sample size is 714 trees (200 wind-thrown). Average sample
size for a cross is 14.0 trees (sum of three replications).
Table4
Expected mean squares for analysis ofvariance ofwindthrow 1
Source
Expected mean square
General combining ability ( GCA)
Specific combining ability (SCA)
Replications (REPS)
Residuals (RES)
aks +ca§c:A +d~ocA
afu:s + ba§c:A
afu:s + aaiEPs
aks
'Coefficients a, b, c and d, respectively, are: 87, 3.58, 3.75 and 24.94 in 1984 set; and 51, 3.05, 3.08
and 20.90 in 1985 set.
specific combining ability (SCA, a non-additive genetic component arising
from specific families); replications (REPS); residual (RES) (Griffing,
1956). Variation was analyzed oy using the diallel computer program ( Schaf­
fer and U sanis, 1969). Expected mean squares for construction ofF-ratios
and estimation of variance components are given in Table 4. Degrees of free­
dom for testing GCA were approximate (Satterthwaite, 1946 ). Variance
components were estimated for each source of variation and expressed as a
percentage of total variance (i.e. sum of variance components). Although we
found significant variation (P<O.Ol) due to reciprocal crosses in the 1983
set, we did not include reciprocal genetic effects in the above model because
24
R.R. Si/en eta/. I Forest Ecology and Management 61 (1993) I 7-28
the unbalanced distribution of reciprocal crosses made the error term for such
a test questionable. Although racial variation may also be present in the data,
valid tests are not possible with these data sets.
To assist in ranking parents for susceptibility or resistance to windthrow,
we divided the number of wind-thrown trees by the number of trees sampled
summed over all crosses and replications for each parent (parent mean per­
cent figures in Table 2). Simple linear regressions and correlations of per­
centage of windthrow versus height were used for inferences concerning
breeding potential and problems.
Results and discussion
Inheritance
Nearly half the trees in the 1983 set and over a quarter in the 1984 set were
wind-thrown ( 45.8% and 28.0%, respectively). With so many random vari­
ables affecting the data, the percentages ofwind-thrown trees per family ranged
from 0 to 100% (Tables 1 and 2). Despite large random environmental vari­
ation and use of only a quarter of the replications of the original study design,
a pattern of genetic variation in susceptibility to windthrow on the two sets is
confirmed in an ANOVA. Also, striking examples are readily observed in Ta­
bles 2 and 3.
In the 1983 set (Table 2), the percentage ofwindthrow in crosses involving
Parent G6 is higher than the average parent performance (parent mean per­
centage) in 21 of the 22 crosses, whether G6 was the male or the female par­
ent in the cross. For Parent 04, progeny from five of its six crosses incurred
lower percentages of windthrow than the mean of its crosses. In the 1984 set
(Table 3 ), crosses involving Parents S1 and S8 incurred more than average
windthrow in all but one of their 17 crosses, and Parents S5 and S9 incurred
less windthrow in all but five oftheir 19 crosses.
The ANOVA (Tables 4 and 5) for the two independent sets indicated highly
significant genetic variation in PROP due to general combining ability
(GCA). In both sets, variation due to specific combining ability ( SCA) was
nearly equal to GCA and attained a 10% level of significance. To estimate
total genetic variation (Griffing, 1956) in the 1983 set, GCA and SCA to­
gether accounted for 21.3% of the total variance in PROP. In the 1984 set,
GCA and SCA together accounted for 33.6% of the total variance. These per­
centages are substantial under circumstances of such a chance event, and in­
heritance of windthrow at the family level is confirmed for the two different
sets at this site.
We suspected that at least part of the variation in susceptibility to windth­
row was related to tree height. Crosses involving Parents G6 and S1, two 'sus­
ceptible' examples cited above, had both averaged 2% taller than the mean
25
R.R. Silen eta/. I Forest Ecology and Management 61 (1993) 17-28
Table 5
Analysis of variance ofwindthrow (PROP and ADJPROP) in 1983 and 1984 sets
Source
d.f. 1
F-ratio 2•3 Mean square
Variance component as
percentage of sum of
variance components
PROP
ADJPROP
PROP
ADJPROP
PROP
ADJPROP
18
53
2
187
260
0.317026
0.116275
0.154063
0.087590
0.275992
0.112780
0.159301
0.087257
2.695....
1.327
1.759
2.421­
1.293
1.826
7.66
7.69
0.73
83.92
6.39
7.01
0.81
85.79
12
37
2
0.301897
0.087863
0.028884
0.058712
0.232712
0.068316
0.093002
0.061287
3.425­
1.497
0.492
3.403**
1.115
1.517
13.03
12.16
0.00 4
74.81
10.91
3.20
0.86
85.03
1984 set
GCA
SCA
REPS
RES
1985 set
GCA
SCA
REPS
RES
.!Q!
152
1
Degrees of freedom. For RES the degrees of freedom for ADJPROP were one less than shown be­
cause of the height adjustment. See text for abbreviations. 2
Asterisks indicate significance ofF-ratios: **0.00 1 < P < O.Ql; no asterisk, P> 0.05. 3
F-ratios for testing GCA are approximate (Satterthwaite, 1946). Approximate degrees offreedom for numerator and denominator, respectively, are 18 and 49 in the 1983 set, and 12 and 36 in the 1984 set. 4
Negative variance component set to 0. plot height at ageS, and crosses involving Parents D4 and S9, which seemed
resistant to windthrow, both averaged 17% shorter than mean HEIGHTS. In
Fig. 2, in which both family HEIGHTS and observed percentage of windth­
row of the two sets are scaled in percentage of the set means, a 10% height
superiority would suggest an approximate 19.6% increase in susceptibility to
windthrow. However, there is also much scatter of data points. Simple linear
regressions of PROP on HEIGHTS accounted for approximately 3.7% of the
variation in PROP in the 1983 set, (P<0.01, n=261 plots; Pearson r=0.193,
P<0.01) and approximately 11.6% ofthe variation in PROP in the 1984 set
(P<0.01, n= 1S3 plots; Pearson r=0.33S, P<0.01 ). Thus, HEIGHTS, as an
estimate of crown size and resistance to wind flow, accounts for a highly sig­
nificant but relatively small portion of the variation in PROP.
A quantitative comparison of family height contribution to windthrow is
the reduction in total genetic variation after PROP is adjusted for mean plot
height (ADJPROP). GCA and SCA together account for 18.6% of the total
variance of ADJPROP in the 1983 set and 22.So/o in the 1984 set ( GCA per­
centage nearly equal to SCA percentage for 1983, but mainly due to GCA for
1984 set). The reduction, approximately 3% and approximately 11 o/o respec­
26
R.R. Silen eta/. I Forest Ecology and Management 61 (1993) 17-28
300
~(J)
0
250
o
•1984SET
~~
200
~~
~
o
~
~!.!1
~~
o
o 1985 SET
::!l
0
•••
150
0
"
z~
oz
.. :
• • •o.
• •o
0
.~.a
... •
0
__
0
•
•
~
0
0
-------------
o ~......•;............
!>b____
0.
~~
~ffi 100t----~~~~~~~~~======~==~
• • .• . . .•,..o-...; o o
Og:o..
*'. . . . . -~. . . g.. c.
50
• "'•
.............
~
_ ......... --
o
0
ooo
o •
•
0 0 ••
o
•
o••
llo
o
o
o
GSxW2
• D4xD2
• 03xD1 0
•o
o~lx~
~
•
j m~o:D:;~m
•o6xD5 o$6xS5
0
80
90
100
110
120
130
HEIGHT AS PERCENT OF SET MEANS
Fig. 2. The proportion of wind-thrown trees tends to increase with mean family height for the
138 crosses in the two sets. Proportions and height are scaled as percentage of set means. The
seven highlighted crosses are the tallest and least susceptible to windthrow (top 10%). Note
common parentage among them.
tively, suggests that one-sixth and one-third of the total genetic variation in
windtl:uow was related to tree height. Thus, the influence of unmeasured ge­
netic factors seem more important in our data sets than height alone, but height
contributions appear to be real. Many questions, such as relevance of tree
size, seed origin, and unbalanced study design cannot be statistically ad­
dressed with our two data sets. Still, any relation with family height clouds
growth benefits of tree improvement, and every opportunity should be taken
to gain clarifying data from future studies, hopefully with larger trees.
Summarizing these analyses, genetic variation in susceptibility to windth­
row was confirmed in the affected fraction of this test of young Douglas-fir
families. The genetic components ( GCA and SCA) accounted for a fifth to a
third ofthe total variation in windthrow. Of the genetic component, a third
or less appeared associated with crown size expressed as tree height.
Although windthrow in older studies on our experimental site was occa­
sionally confined to family rows (Fig. 1 ), the applicability ofthese results for
environments different to those of our data sets is undetermined. Genetic
analyses of other material under broader study conditions are necessary to
determine the true extent of genetic variation in susceptibility to windthrow
and its possible relation to growth. Nevertheless, these results suggest that
breeding for resistance to windthrow could be successful.
Breeding
Candidate trees to breed for resistance to windthrow would most likely be
chosen from wind-damaged genetic studies such as ours. Hence, our data pro­
R.R. Silen eta/. I Forest Ecology and Management 61 (1993) 17-28
27
vide opportunities to observe practical problems with selection, testing, and
environmental factors. Breeding for resistance to windthrow would probably
be complex, expensive, and slow. If slower height growth were acceptable,
selecting resistant parentage would seem straightforward. Parents displaying
both greater resistance to windthrow and fast growth were rare in our test,
however. Only seven crosses ofthe 138 in the two sets exceeded 10% superi­
ority in both mean height and resistance (highlighted in Fig. 2). Only three
of the 12 parents involved in these seven crosses produced families that av­
eraged more than 2% above set mean height; families of six parents averaged
below set mean height. In our data sets, rapid improvement in resistance to
windthrow would use parents that could reduce gains in height growth rate.
Development of artificial means to quantify and test potential susceptibil­
ity to windthrow in young material may be costly but would seem imperative
to a breeding program for resistance to windthrow. Decades could pass with­
out encountering enough natural windthrow at a study site. In addition, ef­
fects of crown size and shape, stem and wood characteristics, root structure
and penetration, and several environmental conditions including landform,
soils, and stand structure on windthrow would be needed (Constantinescu,
1965; Kennedy, 1974; Faber and Sissingh, 1975; de Champs et al., 1982). In
our test, for example, we noted large differences among family rows in the
force needed to restraighten trees, suggesting that root anchorage differed
among families. The ultimate test ofthe product, a wind-resistant population,
might wait decades for near-record winds. Nonetheless, breeding of more ge­
netically windfirm Douglas-fir would be of interest, particularly in the light
of new forestry proposals for Douglas-fir that would expose individuals for­
merly growing in stands (Gillis, 1990).
In contrast to breeding, silvicultural techniques to minimize windthrow may
be simpler and less expensive. Extensive studies of wind damage have pro­
vided a reasonable understanding of contributing factors and many sugges­
tions for reducing potential damage (Kennedy, 1974; Faber and Sissingh,
1975; Bell et al., 1983; Liegel, 1984 ). Because Douglas-fir ranks among the
more wind-resistant species (de Champs et al., 1982), most trees on average,
well-managed sites may be adequately resistant to such near-record storms.
Much windthrow could probably be avoided from recognition and proper
treatment of a small proportion offorest areas that are potential problem sites
(e.g. wet soils, exposed stand edges and comers, and 'wind-funnelling'
topography).
The most immediately useful genetic information from this study may be
that some resistance to windthrow in seed-orchard seed could result from ro­
guing susceptible parentage based on windthrow damage recorded in the
hundreds of Douglas-fir progeny tests now regularly examined. In our test,
the elimination of crosses involving the one parent most susceptible to
windthrow might have reduced the population mean windthrow from 45.8 to
28
R.R. Silen eta/. I Forest Ecology and Management 61 (1993) 17-28
40.8% in the 1984 set, and from 28.0 to 23.7% in the 1985 set: a 10% and 15%
reduction, respectively. These simpler and more immediate measures may
serve reasonably well until populations resistant to windthrow are bred.
Acknowledgments
We thank Nancy Mandel, Rich Sneizko, and Doug Neeley for assistance in
statistical analysis. We also thank Robert Campbell, Lauren Fins, Donald Mi­
nore, and Dean DeBell for critical reviews of the manuscript.
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