Description Usage Arguments Details Value Note Author(s) References See Also Examples
View source: R/kendallSeasonalTrendTest.R
Perform a nonparametric test for a monotonic trend within each season based on Kendall's tau statistic, and optionally compute a confidence interval for the slope across all seasons.
1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17  kendallSeasonalTrendTest(y, ...)
## S3 method for class 'formula'
kendallSeasonalTrendTest(y, data = NULL, subset,
na.action = na.pass, ...)
## Default S3 method:
kendallSeasonalTrendTest(y, season, year,
alternative = "two.sided", correct = TRUE, ci.slope = TRUE, conf.level = 0.95,
independent.obs = TRUE, data.name = NULL, season.name = NULL, year.name = NULL,
parent.of.data = NULL, subset.expression = NULL, ...)
## S3 method for class 'data.frame'
kendallSeasonalTrendTest(y, ...)
## S3 method for class 'matrix'
kendallSeasonalTrendTest(y, ...)

y 
an object containing data for the trend test. In the default method,
the argument 
data 
specifies an optional data frame, list or environment (or object coercible by

subset 
specifies an optional vector specifying a subset of observations to be used. 
na.action 
specifies a function which indicates what should happen when the data contain 
season 
numeric or character vector or a factor indicating the seasons in which the observations in

year 
numeric vector indicating the years in which the observations in 
alternative 
character string indicating the kind of alternative hypothesis. The
possible values are 
correct 
logical scalar indicating whether to use the correction for continuity in
computing the zstatistic that is based on the test statistic S'.
The default value is 
ci.slope 
logical scalar indicating whether to compute a confidence interval for the
slope. The default value is 
conf.level 
numeric scalar between 0 and 1 indicating the confidence level associated
with the confidence interval for the slope. The default value is

independent.obs 
logical scalar indicating whether to assume the observations in 
data.name 
character string indicating the name of the data used for the trend test.
The default value is 
season.name 
character string indicating the name of the data used for the season.
The default value is 
year.name 
character string indicating the name of the data used for the year.
The default value is 
parent.of.data 
character string indicating the source of the data used for the trend test. 
subset.expression 
character string indicating the expression used to subset the data. 
... 
additional arguments affecting the test for trend. 
Hirsch et al. (1982) introduced a modification of Kendall's test for trend
(see kendallTrendTest
) that allows for seasonality in observations collected over time.
They call this test the seasonal Kendall test. Their test is appropriate for testing for
trend in each season when the trend is always in the same direction across all seasons.
van Belle and Hughes (1984) introduced a heterogeneity test for trend which is appropriate for testing
for trend in any direction in any season. Hirsch and Slack (1984) proposed an extension to the seasonal
Kendall test that allows for serial dependence in the observations. The function
kendallSeasonalTrendTest
includes all of these tests, as well as an extension of the
van BelleHughes heterogeneity test to the case of serial dependence.
Testing for Trend Assuming Serial Independence
The Model
Assume observations are taken over two or more years, and assume a single year
can be divided into two or more seasons. Let p denote the number of seasons.
Let X and Y denote two continuous random variables with some joint
(bivariate) distribution (which may differ from season to season). Let N_j
denote the number of bivariate observations taken in the j'th season (over two
or more years) (j = 1, 2, …, p), so that
(X_{1j}, Y_{1j}), (X_{2j}, Y_{2j}), …, (X_{N_jj}, Y_{N_jj})
denote the N_j bivariate observations from this distribution for season j, assume these bivariate observations are mutually independent, and let
τ_j = \{ 2 Pr[(X_{2j}  X_{1j})(Y_{2j}  Y_{1j}) > 0] \}  1 \;\;\;\;\;\; (1)
denote the value of Kendall's tau for that season (see kendallTrendTest
).
Also, assume all of the Y observations are independent.
The function kendallSeasonalTrendTest
assumes that the X values always
denote the year in which the observation was taken. Note that within any season,
the X values need not be unique. That is, there may be more than one
observation within the same year within the same season. In this case, the
argument y
must be a numeric vector, and you must supply the additional
arguments season
and year
.
If there is only one observation per season per year (missing values allowed), it is
usually easiest to supply the argument y
as an n \times p matrix or
data frame, where n denotes the number of years. In this case
N_1 = N_2 = \cdots = N_p = n \;\;\;\;\;\; (2)
and
X_{ij} = i \;\;\;\;\;\; (3)
for i = 1, 2, …, n and j = 1, 2, …, p, so if Y denotes the n \times p matrix of observed Y's and X denotes the n \times p matrix of the X's, then
Y_{11}  Y_{12}  \cdots  Y_{1p}  
Y_{21}  Y_{22}  \cdots  Y_{2p}  
\underline{Y} =  .  \;\;\;\;\;\; (4)  
.  
.  
Y_{n1}  Y_{n2}  \cdots  Y_{np} 
1  1  \cdots  1  
2  2  \cdots  2  
\underline{X} =  .  \;\;\;\;\;\; (5)  
.  
.  
n  n  \cdots  n 
The null hypothesis is that within each season the X and Y random variables are independent; that is, within each season there is no trend in the Y observations over time. This null hypothesis can be expressed as:
H_0: τ_1 = τ_2 = \cdots = τ_p = 0 \;\;\;\;\;\; (6)
The Seasonal Kendall Test for Trend
Hirsch et al.'s (1982) extension of Kendall's tau statistic to test the null
hypothesis (6) is based on simply summing together the Kendall Sstatistics
for each season and computing the following statistic:
z = \frac{S'}{√{Var(S')}} \;\;\;\;\;\; (7)
or, using the correction for continuity,
z = \frac{S'  sign(S')}{√{Var(S')}} \;\;\;\;\;\; (8)
where
S' = ∑_{j=1}^p S_j \;\;\;\;\;\; (9)
S_j = ∑_{i=1}^{N_j1} ∑_{k=i+1}^{N_j} sign[(X_{kj}  X_{ij})(Y_{kj}  Y_{ij})] \;\;\;\;\;\; (10)
and sign() denotes the sign
function:
1  x < 0  
sign(x) =  0  x = 0 \;\;\;\;\;\; (11) 
1  x > 0 
Note that the quantity in Equation (10) is simply the Kendall Sstatistic for
season j (j = 1, 2, …, p) (see Equation (3) in the help file for
kendallTrendTest
).
For each season, if the predictor variables (the X's) are strictly increasing (e.g., Equation (3) above), then Equation (10) simplifies to
S_j = ∑_{i=1}^{N_j1} ∑_{k=i+1}^{N_j} sign[(Y_{kj}  Y_{ij})] \;\;\;\;\;\; (12)
Under the null hypothesis (6), the quantity z defined in Equation (7) or (8) is approximately distributed as a standard normal random variable.
Note that there may be missing values in the observations, so let n_j denote the number of (X,Y) pairs without missing values for season j.
The statistic S' in Equation (9) has mean and variance given by:
E(S') = ∑_{j = 1}^p E(S_j) \;\;\;\;\;\; (13)
Var(S') = ∑_{j = 1}^p Var(S_j) + ∑_{g=1}^{p1} ∑_{h=g+1}^{p} 2 Cov(S_g, S_h) \;\;\;\;\;\; (14)
Since all the observations are assumed to be mutually independent,
σ_{gh} = Cov(S_g, S_h) = 0, \;\; g \ne h, \;\; g,h = 1, 2, …, p \;\;\;\;\;\; (15)
Furthermore, under the null hypothesis (6),
E(S_j) = 0, \;\; j = 1, 2, …, p \;\;\;\;\;\; (16)
and, in the case of no tied observations,
Var(S_j) = \frac{n_j(n_j1)(2n_j+5)}{18} \;\;\;\;\;\; (17)
for j=1,2, …, p (see equation (7) in the help file for
kendallTrendTest
).
In the case of tied observations,
Var(S_j) =  \frac{n_j(n_j1)(2n_j+5)}{18}  
\frac{∑_{i=1}^{g} t_i(t_i1)(2t_i+5)}{18}   
\frac{∑_{k=1}^{h} u_k(u_k1)(2u_k+5)}{18} +  
\frac{[∑_{i=1}^{g} t_i(t_i1)(t_i2)][∑_{k=1}^{h} u_k(u_k1)(u_k2)]}{9n_k(n_k1)(n_k2)} +  
\frac{[∑_{i=1}^{g} t_i(t_i1)][∑_{k=1}^{h} u_k(u_k1)]}{2n_k(n_k1)} \;\;\;\;\;\; (18) 
where g is the number of tied groups in the X observations for
season j, t_i is the size of the i'th tied group in the X
observations for season j, h is the number of tied groups in the Y
observations for season j, and u_k is the size of the k'th tied
group in the Y observations for season j
(see Equation (9) in the help file for kendallTrendTest
).
Estimating τ, Slope, and Intercept
The function kendall.SeasonalTrendTest
returns estimated values of
Kendall's τ, the slope, and the intercept for each season, as well as a
single estimate for each of these three quantities combined over all seasons.
The overall estimate of τ is the weighted average of the p seasonal
τ's:
\hat{τ} = \frac{∑_{j=1}^p n_j \hat{τ}_j}{∑_{j=1}^p n_j} \;\;\;\;\;\; (19)
where
\hat{τ}_j = \frac{2S_j}{n_j(n_j1)} \;\;\;\;\;\; (20)
(see Equation (2) in the help file for kendallTrendTest
).
We can compute the estimated slope for season j as:
\hat{β}_{1_j} = Median(\frac{Y_{kj}  Y_{ij}}{X_{kj}  X_{ij}}); \;\; i < k; \;\; X_{kj} \ne X_{ik} \;\;\;\;\;\; (21)
for j = 1, 2, …, p. The overall estimate of slope, however, is not the median of these p estimates of slope; instead, following Hirsch et al. (1982, p.117), the overall estimate of slope is the median of all twopoint slopes computed within each season:
\hat{β}_1 = Median(\frac{Y_{kj}  Y_{ij}}{X_{kj}  X_{ij}}); \;\; i < k; \;\; X_{kj} \ne X_{ik}; \;\; j = 1, 2, …, p \;\;\;\;\;\; (22)
(see Equation (15) in the help file for kendallTrendTest
).
The overall estimate of intercept is the median of the p seasonal estimates of intercept:
\hat{β}_0 = Median(\hat{β}_{0_1}, \hat{β}_{0_2}, …, \hat{β}_{0_p}) \;\;\;\;\;\; (23)
where
\hat{β}_{0_j} = Y_{0.5_j}  \hat{β}_{1_j} X_{0.5_j} \;\;\;\;\;\; (24)
and X_{0.5_j} and Y_{0.5_j} denote the sample medians of the X's
and Y's, respectively, for season j
(see Equation (16) in the help file for kendallTrendTest
).
Confidence Interval for the Slope
Gilbert (1987, p.227228) extends his method of computing a confidence interval for
the slope to the case of seasonal observations. Let N' denote the number of
defined twopoint estimated slopes that are used in Equation (22) above and let
\hat{β}_{1_{(1)}}, \hat{β}_{1_{(2)}}, …, \hat{β}_{1_{(N')}}
denote the N' ordered slopes. For Gilbert's (1987) method, a 100(1α)\% twosided confidence interval for the true overall slope across all seasons is given by:
[\hat{β}_{1_{(M1)}}, \hat{β}_{1_{(M2+1)}}] \;\;\;\;\;\; (25)
where
M1 = \frac{N'  C_{α}}{2} \;\;\;\;\;\; (26)
M2 = \frac{N' + C_{α}}{2} \;\;\;\;\;\; (27)
C_α = z_{1  α/2} √{Var(S')} \;\;\;\;\;\; (28)
Var(S') is defined in Equation (14), and z_p denotes the p'th quantile of the standard normal distribution. Onesided confidence intervals may computed in a similar fashion.
Usually the quantities M1 and M2 will not be integers.
Gilbert (1987, p.219) suggests interpolating between adjacent values in this case,
which is what the function kendallSeasonalTrendTest
does.
The Van BelleHughes Heterogeneity Test for Trend
The seasonal Kendall test described above is appropriate for testing the null
hypothesis (6) against the alternative hypothesis of a trend in at least one season.
All of the trends in each season should be in the same direction.
The seasonal Kendall test is not appropriate for testing for trend when there are trends in a positive direction in one or more seasons and also negative trends in one or more seasons. For example, for the following set of observations, the seasonal Kendall statistic S' is 0 with an associated twosided pvalue of 1, even though there is clearly a positive trend in season 1 and a negative trend in season 2.
Year  Season 1  Season 2 
1  5  8 
2  6  7 
3  7  6 
4  8  5 
Van Belle and Hughes (1984) suggest using the following statistic to test for heterogeneity in trend prior to applying the seasonal Kendall test:
χ_{het}^2 = ∑_{j=1}^p Z_j^2  p \bar{Z}^2 \;\;\;\;\;\; (29)
where
Z_j = \frac{S_j}{Var(S_j)} \;\;\;\;\;\; (30)
\bar{Z} = \frac{1}{p} ∑_{j=1}^p Z_j \;\;\;\;\;\; (31)
Under the null hypothesis (6), the statistic defined in Equation (29) is approximately distributed as a chisquare random variable with p1 degrees of freedom. Note that the continuity correction is not used to compute the Z_j's defined in Equation (30) since using it results in an unacceptably conservative test (van Belle and Hughes, 1984, p.132). Van Belle and Hughes (1984) actually call the statistic in (29) a homogeneous chisquare statistic. Here it is called a heterogeneous chisquare statistic after the alternative hypothesis it is meant to test.
Van Belle and Hughes (1984) imply that the heterogeneity statistic defined in Equation (29) may be used to test the null hypothesis:
H_0: τ_1 = τ_2 = \cdots = τ_p = τ \;\;\;\;\;\; (32)
where τ is some arbitrary number between 1 and 1. For this case, however,
the distribution of the test statistic in Equation (29) is unknown since it depends
on the unknown value of τ (Equations (16)(18) above assume
τ = 0 and are not correct if τ \ne 0). The heterogeneity
chisquare statistic of Equation (29) may be assumed to be approximately
distributed as chisquare with p1 degrees of freedom under the null
hypothesis (32), but further study is needed to determine how well this
approximation works.
Testing for Trend Assuming Serial Dependence
The Model
Assume the same model as for the case of serial independence, except now the
observed Y's are not assumed to be independent of one another, but are
allowed to be serially correlated over time. Furthermore, assume one observation
per season per year (Equations (2)(5) above).
The Seasonal Kendall Test for Trend Modified for Serial Dependence
Hirsch and Slack (1984) introduced a modification of the seasonal Kendall test that
is robust against serial dependence (in terms of Type I error) except when the
observations have a very strong longterm persistence (very large autocorrelation) or
when the sample sizes are small (e.g., 5 years of monthly data). This modification
is based on a multivariate test introduced by Dietz and Killeen (1981).
In the case of serial dependence, Equation (15) is no longer true, so an estimate of the correct value of σ_{gh} must be used to compute Var(S') in Equation (14). Let R denote the n \times p matrix of ranks for the Y observations (Equation (4) above), where the Y's are ranked within season:
R_{11}  R_{12}  \cdots  R_{1p}  
R_{21}  R_{22}  \cdots  R_{2p}  
\underline{R} =  .  \;\;\;\;\;\; (33)  
.  
.  
R_{n1}  R_{n2}  \cdots  R_{np} 
where
R_{ij} = \frac{1}{2} [n_j + 1 ∑_{k=1}^{n_j} sign(Y_{ij}  Y_{kj})] \;\;\;\;\;\; (34)
the sign function is defined in Equation (11) above, and as before n_j denotes the number of (X,Y) pairs without missing values for season j. Note that by this definition, missing values are assigned the midrank of the nonmissing values.
Hirsch and Slack (1984) suggest using the following formula, given by Dietz and Killeen (1981), in the case where there are no missing values:
\hat{σ}_{gh} = \frac{1}{3} [K_{gh} + 4 ∑_{i=1}^n R_{ig}R_{ih}  n(n+1)^2] \;\;\;\;\;\; (35)
where
K_{gh} = ∑_{i=1}^{n1} ∑_{j=i+1}^n sign[(Y_{jg}  Y_{ig})(Y_{jh}  Y_{ih})] \;\;\;\;\;\; (36)
Note that the quantity defined in Equation (36) is Kendall's tau for season g vs. season h.
For the case of missing values, Hirsch and Slack (1984) derive the following modification of Equation (35):
\hat{σ}_{gh} = \frac{1}{3} [K_{gh} + 4 ∑_{i=1}^n R_{ig}R_{ih}  n(n_g + 1)(n_h + 1)] \;\;\;\;\;\; (37)
Technically, the estimates in Equations (35) and (37) are not correct estimators of
covariance, and Equations (17) and (18) are not correct estimators of variance,
because the model Dietz and Killeen (1981) use assumes that observations within the
rows of Y (Equation (4) above) may be correlated, but observations between
rows are independent. Serial dependence induces correlation between all of the
Y's. In most cases, however, the serial dependence shows an exponential decay
in correlation across time and so these estimates work fairly well (see more
discussion in the BACKGROUND section below).
Estimates and Confidence Intervals
The seasonal and overall estimates of τ, slope, and intercept are computed
using the same methods as in the case of serial independence. Also, the method for
computing the confidence interval for the slope is the same as in the case of serial
independence. Note that the serial dependence is accounted for in the term
Var(S') in Equation (28).
The Van BelleHughes Heterogeneity Test for Trend Modified for Serial Dependence
Like its counterpart in the case of serial independence, the seasonal Kendall test
modified for serial dependence described above is appropriate for testing the null
hypothesis (6) against the alternative hypothesis of a trend in at least one season.
All of the trends in each season should be in the same direction.
The modified seasonal Kendall test is not appropriate for testing for trend when there are trends in a positive direction in one or more seasons and also negative trends in one or more seasons. This section describes a modification of the van BelleHughes heterogeneity test for trend in the presence of serial dependence.
Let \underline{S} denote the p \times 1 vector of Kendall Sstatistics for each season:
S_1  
S_2  
\underline{S} =  .  \;\;\;\;\;\; (38) 
.  
.  
S_p 
The distribution of \underline{S} is approximately multivariate normal with
μ_1  
μ_2  
E(\underline{S}) = \underline{μ} =  .  \;\;\;\;\;\; (39) 
.  
.  
μ_p 
σ_1^2  σ_{12}  \cdots  σ_{1p}  
σ_{21}  σ_2^2  \cdots  σ_{2p}  
Var(\underline{S}) = Σ =  .  \;\;\;\;\;\; (40)  
.  
.  
σ_{n1}  σ_{n2}  \cdots  σ_n^2 
where
μ_j = \frac{n_j(n_j  1)}{2} τ_j, \;\; j = 1, 2, …, p \;\;\;\;\;\; (41)
Define the p \times p matrix \underline{m} as
\frac{2}{n_1(n_1  1)}  0  \cdots  0  
0  \frac{2}{n_2(n_2  1)}  \cdots  0  
\underline{m} =  .  \;\;\;\;\;\; (42)  
.  
.  
0  0  \cdots  \frac{2}{n_p(n_p  1)} 
Then the vector of the seasonal estimates of τ can be written as:
\hat{τ}_1  2S_1/[n_1(n_11)]  
\hat{τ}_2  2S_2/[n_2(n_21)]  
\underline{\hat{τ}} =  .  =  .  = \underline{m} \; \underline{S} \;\;\;\;\; (43) 
.  .  
.  .  
\hat{τ}_p  2S_p/[n_p(n_p1)] 
so the distribution of the vector in Equation (43) is approximately multivariate normal with
τ_1  
τ_2  
E(\underline{\hat{τ}}) =  E(\underline{m} \underline{S}) =  \underline{m} \underline{μ} =  \underline{τ} =  .  \;\;\;\;\;\; (44) 
.  
.  
τ_p 
Var(\underline{\hat{τ}}) = Var(\underline{m} \; \underline{S}) = \underline{m} Σ \underline{m}^T = Σ_{\hat{τ}} \;\;\;\;\;\; (45)
where ^T denotes the transpose operator. Let \underline{C} denote the (p1) \times p contrast matrix
\underline{C} = [\; \underline{1} \;  \; I_p] \;\;\;\;\;\; (46)
where I_p denotes the p \times p identity matrix. That is,
1  1  0  \cdots  0  
1  0  1  \cdots  0  
\underline{C} =  .  .  
.  .  
.  .  
1  0  0  \cdots  1 
Then the null hypothesis (32) is equivalent to the null hypothesis:
H_0: \underline{C} \underline{τ} = 0 \;\;\;\;\;\; (47)
Based on theory for samples from a multivariate normal distribution (Johnson and Wichern, 2007), under the null hypothesis (47) the quantity
χ_{het}^2 = (\underline{C} \; \underline{\hat{τ}})^T (\underline{C} \hat{Σ}_{\hat{τ}} \underline{C}^T)^{1} (\underline{C} \; \underline{\hat{τ}}) \;\;\;\;\;\; (48)
has approximately a chisquare distribution with p1 degrees of freedom for “large” values of seasonal sample sizes, where
\hat{Σ}_{\hat{τ}} = \underline{m} \hat{Σ} \underline{m}^T \;\;\;\;\;\; (49)
The estimate of Σ in Equation (49) can be computed using the same formulas that are used for the modified seasonal Kendall test (i.e., Equation (35) or (37) for the offdiagonal elements and Equation (17) or (18) for the diagonal elements). As previously noted, the formulas for the variances are actually only valid if t = 0 and there is no correlation between the rows of Y. The same is true of the formulas for the covariances. More work is needed to determine the goodness of the chisquare approximation for the test statistic in (48). The pseudoheterogeneity test statistic of Equation (48), however, should provide some guidance as to whether the null hypothesis (32) (or equivalently (47)) appears to be true.
A list of class "htest"
containing the results of the hypothesis
test. See the help file for htest.object
for details.
In addition, the following components are part of the list returned by
kendallSeasonalTrendTest
:
seasonal.S 
numeric vector. The value of the Kendall Sstatistic for each season. 
var.seasonal.S 
numeric vector. The variance of the Kendall Sstatistic for each season.
This component only appears when 
var.cov.seasonal.S 
numeric matrix. The estimated variancecovariance matrix of the Kendall
Sstatistics for each season. This component only appears when 
seasonal.estimates 
numeric matrix. The estimated Kendall's tau, slope, and intercept for each season. 
Kendall's test for independence or trend is a nonparametric test. No assumptions are made about the distribution of the X and Y variables. Hirsch et al. (1982) introduced the seasonal Kendall test to test for trend within each season. They note that Kendall's test for trend is easy to compute, even in the presence of missing values, and can also be used with censored values.
van Belle and Hughes (1984) note that the seasonal Kendall test introduced by Hirsch et al. (1982) is similar to a multivariate extension of the sign test proposed by Jonckheere (1954). Jonckheeere's test statistic is based on the unweighted sum of the seasonal tau statistics, while Hirsch et al.'s test is based on the weighted sum (weighted by number of observations within a season) of the seasonal tau statistics.
van Belle and Hughes (1984) also note that Kendall's test for trend is slightly less powerful than the test based on Spearman's rho, but it converges to normality faster. Also, Bradley (1968, p.288) shows that for the case of a linear model with normal (Gaussian) errors, the asymptotic relative efficiency of Kendall's test for trend versus the parametric test for a zero slope is 0.98.
Based on the work of Dietz and Killeen (1981), Hirsch and Slack (1984) describe a modified version of the seasonal Kendall test that allows for serial dependence in the observations. They performed a Monte Carlo study to determine the empirical significance level and power of this modified test vs. the test that assumes independent observations and found a tradeoff between power and the correct significance level. For p = 12 seasons, they found the modified test gave correct significance levels for n ≥q 10 as long as the lagone autocorrelation was 0.6 or less, while the original test that assumes independent observations yielded highly inflated significance levels. On the other hand, if in fact the observations are serially independent, the original test is more powerful than the modified test.
Hirsch and Slack (1984) also looked at the performance of the test for trend introduced by Dietz and Killeen (1981), which is a weighted sums of squares of the seasonal Kendall Sstatistics, where the matrix of weights is the inverse of the covariance matrix. The DietzKilleen test statistic, unlike the one proposed by Hirsh and Slack (1984), tests for trend in either direction in any season, and is asymptotically distributed as a chisquare random variable with p (number of seasons) degrees of freedom. Hirsch and Slack (1984), however, found that the test based on this statistic is quite conservative (i.e., the significance level is much smaller than the assumed significance level) and has poor power even for moderate sample sizes. The chisquare approximation becomes reasonably close only when n > 40 if p = 12, n > 30 if p = 4, and n > 20 if p = 2.
Lettenmaier (1988) notes the poor power of the test proposed by Dietz and Killeen (1981) and states the poor power apparently results from an upward bias in the estimated variance of the statistic, which can be traced to the inversion of the estimated covariance matrix. He suggests an alternative test statistic (to test trend in either direction in any season) that is the sum of the squares of the scaled seasonal Kendall Sstatistics (scaled by their standard deviations). Note that this test statistic ignores information about the covariance between the seasonal Kendall Sstatistics, although its distribution depends on these covariances. In the case of no serial dependence, Lettenmaier's test statistic is exactly the same as the DietzKilleen test statistic. In the case of serial dependence, Lettenmaier (1988) notes his test statistic is a quadratic form of a multivariate normal random variable and therefore all the moments of this random variable are easily computed. Lettenmaier (1988) approximates the distribution of his test statistic as a scaled noncentral chisquare distribution (with fractional degrees of freedom). Based on extensive Monte Carlo studies, Lettenmaier (1988) shows that for the case when the trend is the same in all seasons, the seasonal Kendall's test of Hirsch and Slack (1984) is superior to his test and far superior to the DietzKilleen test. The power of Lettenmaier's test approached that of the seasonal Kendall test for large trend magnitudes.
Steven P. Millard (EnvStats@ProbStatInfo.com)
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Hirsch, R.M. and J.R. Slack. (1984). A Nonparametric Trend Test for Seasonal Data with Serial Dependence. Water Resources Research 20(6), 727732.
Hirsch, R.M., R.B. Alexander, and R.A. Smith. (1991). Selection of Methods for the Detection and Estimation of Trends in Water Quality. Water Resources Research 27(5), 803813.
Hollander, M., and D.A. Wolfe. (1999). Nonparametric Statistical Methods, Second Edition. John Wiley and Sons, New York.
Johnson, R.A., and D.W. Wichern. (2007). Applied Multivariate Statistical Analysis, Sixth Edition. Pearson Prentice Hall, Upper Saddle River, NJ.
Kendall, M.G. (1938). A New Measure of Rank Correlation. Biometrika 30, 8193.
Kendall, M.G. (1975). Rank Correlation Methods. Charles Griffin, London.
Mann, H.B. (1945). Nonparametric Tests Against Trend. Econometrica 13, 245259.
Millard, S.P., and Neerchal, N.K. (2001). Environmental Statistics with SPLUS. CRC Press, Boca Raton, Florida.
Sen, P.K. (1968). Estimates of the Regression Coefficient Based on Kendall's Tau. Journal of the American Statistical Association 63, 13791389.
Theil, H. (1950). A RankInvariant Method of Linear and Polynomial Regression Analysis, IIII. Proc. Kon. Ned. Akad. v. Wetensch. A. 53, 386392, 521525, 13971412.
USEPA. (2009). Statistical Analysis of Groundwater Monitoring Data at RCRA Facilities, Unified Guidance. EPA 530/R09007, March 2009. Office of Resource Conservation and Recovery Program Implementation and Information Division. U.S. Environmental Protection Agency, Washington, D.C.
USEPA. (2010). Errata Sheet  March 2009 Unified Guidance. EPA 530/R09007a, August 9, 2010. Office of Resource Conservation and Recovery, Program Information and Implementation Division. U.S. Environmental Protection Agency, Washington, D.C.
van Belle, G., and J.P. Hughes. (1984). Nonparametric Tests for Trend in Water Quality. Water Resources Research 20(1), 127136.
kendallTrendTest
, htest.object
, cor.test
.
1 2 3 4 5 6 7 8 9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24 25 26 27 28 29 30 31 32 33 34 35 36 37 38 39 40 41 42 43 44 45 46 47 48 49 50 51 52 53 54 55 56 57 58 59 60 61 62 63 64 65 66 67 68 69 70 71 72 73 74 75 76 77 78 79 80 81 82 83 84 85 86 87 88 89 90 91 92 93 94 95 96 97 98 99 100 101 102 103 104 105 106 107 108 109 110 111 112 113 114 115 116 117 118 119 120 121 122 123 124 125 126 127 128 129 130  # Reproduce Example 1410 on page 1438 of USEPA (2009). This example
# tests for trend in analyte concentrations (ppm) collected monthly
# between 1983 and 1985.
head(EPA.09.Ex.14.8.df)
# Month Year Unadj.Conc Adj.Conc
#1 January 1983 1.99 2.11
#2 February 1983 2.10 2.14
#3 March 1983 2.12 2.10
#4 April 1983 2.12 2.13
#5 May 1983 2.11 2.12
#6 June 1983 2.15 2.12
tail(EPA.09.Ex.14.8.df)
# Month Year Unadj.Conc Adj.Conc
#31 July 1985 2.31 2.23
#32 August 1985 2.32 2.24
#33 September 1985 2.28 2.23
#34 October 1985 2.22 2.24
#35 November 1985 2.19 2.25
#36 December 1985 2.22 2.23
# Plot the data
#
Unadj.Conc < EPA.09.Ex.14.8.df$Unadj.Conc
Adj.Conc < EPA.09.Ex.14.8.df$Adj.Conc
Month < EPA.09.Ex.14.8.df$Month
Year < EPA.09.Ex.14.8.df$Year
Time < paste(substring(Month, 1, 3), Year  1900, sep = "")
n < length(Unadj.Conc)
Three.Yr.Mean < mean(Unadj.Conc)
dev.new()
par(mar = c(7, 4, 3, 1) + 0.1, cex.lab = 1.25)
plot(1:n, Unadj.Conc, type = "n", xaxt = "n",
xlab = "Time (Month)",
ylab = "ANALYTE CONCENTRATION (mg/L)",
main = "Figure 1415. Seasonal Time Series Over a Three Year Period",
cex.main = 1.1)
axis(1, at = 1:n, labels = rep("", n))
at < rep(c(1, 5, 9), 3) + rep(c(0, 12, 24), each = 3)
axis(1, at = at, labels = Time[at])
points(1:n, Unadj.Conc, pch = 0, type = "o", lwd = 2)
points(1:n, Adj.Conc, pch = 3, type = "o", col = 8, lwd = 2)
abline(h = Three.Yr.Mean, lwd = 2)
legend("topleft", c("Unadjusted", "Adjusted", "3Year Mean"), bty = "n",
pch = c(0, 3, 1), lty = c(1, 1, 1), lwd = 2, col = c(1, 8, 1),
inset = c(0.05, 0.01))
# Perform the seasonal Kendall trend test
#
kendallSeasonalTrendTest(Unadj.Conc ~ Month + Year,
data = EPA.09.Ex.14.8.df)
#Results of Hypothesis Test
#
#
#Null Hypothesis: All 12 values of tau = 0
#
#Alternative Hypothesis: The seasonal taus are not all equal
# (ChiSquare Heterogeneity Test)
# At least one seasonal tau != 0
# and all nonzero tau's have the
# same sign (z Trend Test)
#
#Test Name: Seasonal Kendall Test for Trend
# (with continuity correction)
#
#Estimated Parameter(s): tau = 0.9722222
# slope = 0.0600000
# intercept = 131.7350000
#
#Estimation Method: tau: Weighted Average of
# Seasonal Estimates
# slope: Hirsch et al.'s
# Modification of
# Thiel/Sen Estimator
# intercept: Median of
# Seasonal Estimates
#
#Data: y = Unadj.Conc
# season = Month
# year = Year
#
#Data Source: EPA.09.Ex.14.8.df
#
#Sample Sizes: January = 3
# February = 3
# March = 3
# April = 3
# May = 3
# June = 3
# July = 3
# August = 3
# September = 3
# October = 3
# November = 3
# December = 3
# Total = 36
#
#Test Statistics: ChiSquare (Het) = 0.1071882
# z (Trend) = 5.1849514
#
#Test Statistic Parameter: df = 11
#
#Pvalues: ChiSquare (Het) = 1.000000e+00
# z (Trend) = 2.160712e07
#
#Confidence Interval for: slope
#
#Confidence Interval Method: Gilbert's Modification of
# Theil/Sen Method
#
#Confidence Interval Type: twosided
#
#Confidence Level: 95%
#
#Confidence Interval: LCL = 0.05786914
# UCL = 0.07213086
#==========
# Clean up
#
rm(Unadj.Conc, Adj.Conc, Month, Year, Time, n, Three.Yr.Mean, at)
graphics.off()

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