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Zero- vs. one-dimensional, parametric vs. non-parametric, and confidence interval vs. hypothesis testing procedures in one-dimensional biomechanical trajectory analysis


Abstract and Figures

Biomechanical processes are often manifested as one-dimensional (1D) trajectories. It has been shown that 1D confidence intervals (CIs) are biased when based on 0D statistical procedures, and the non-parametric 1D bootstrap CI has emerged in the Biomechanics literature as a viable solution. The primary purpose of this paper was to clarify that, for 1D biomechanics datasets, the distinction between 0D and 1D methods is much more important than the distinction between parametric and non-parametric procedures. A secondary purpose was to demonstrate that a parametric equivalent to the 1D bootstrap exists in the form of a random field theory (RFT) correction for multiple comparisons. To emphasize these points we analyzed six datasets consisting of force and kinematic trajectories in one-sample, paired, two-sample and regression designs. Results showed, first, that the 1D bootstrap and other 1D non-parametric CIs were qualitatively identical to RFT CIs, and all were very different from 0D CIs. Second, 1D parametric and 1D non-parametric hypothesis testing results were qualitatively identical for all six datasets. Last, we highlight the limitations of 1D CIs by demonstrating that they are complex, design-dependent, and thus non-generalizable. These results suggest that (i) analyses of 1D data based on 0D models of randomness are generally biased unless one explicitly identifies 0D variables before the experiment, and (ii) parametric and non-parametric 1D hypothesis testing provide an unambiguous framework for analysis when one׳s hypothesis explicitly or implicitly pertains to whole 1D trajectories. Copyright © 2015 Elsevier Ltd. All rights reserved.
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Zero- vs. one-dimensional, parametric vs. non-parametric, and
confidence interval vs. hypothesis testing procedures in
one-dimensional biomechanical trajectory analysis
Todd C. Pata ky1, Jos Vanrenterghem2, and Mark A. Robinson2
1Department of Bioengineering, Shinshu University, Japan
2Research Institute for Sport and Exercise Sciences, Liverpool John Moores University, UK
March 4, 2015
Biomechanical processes are often manifested as one-dimensional (1D) trajectories. It has been shown
that 1D confidence intervals (CIs) are biased when based on 0D statistical procedures, and the non-
parametric 1D bootstrap CI has emerged in the Biomechanics literature as a viable solution. The
primary purpose of this paper was to clarify that, for 1D biomechanics datasets, the distinction between
0D and 1D methods is much more important than the distinction between parametric and non-parametric
procedures. A secondary purpose was to demonstrate that a parametric equivalent to the 1D bootstrap
exists in the form of a random field theory (RFT) correction for multiple comparisons. To emphasize
these points we analyzed six datasets consisting of force and kinematic trajectories in one-sample, paired,
two-sample and regression designs. Results showed, first, that the 1D bootstrap and other 1D non-
parametric CIs were qualitatively identical to RFT CIs, and all were very dierent from 0D CIs. Second,
1D parametric and 1D non-parametric hypothesis testing results were qualitatively identical for all six
datasets. Last, we highlight the limitations of 1D CIs by demonstrating that they are complex, design-
dependent, and thus non-generalizable. These results suggest that (i) analyses of 1D data based on 0D
models of randomness are generally biased unless one has explicitly specified an apriori0D variable,
and (ii) parametric and non-parametric 1D hypothesis testing provide an unambiguous framework for
analysis when one’s hypothesis explicitly or implicitly pertains to whole 1D trajectories.
Corresponding Author:
Todd Pataky,, T.+81-268-21-5609, F.+81-268–21-5318
Keywords: bootstrap confidence interval; kinematics; ground reaction force; statistical parametric map-
ping; random field theory; time series analysis
1 Introduction
Biomechanical processes are often described using one-dimensional (1D) kinematic and force trajectories.
Since these trajectories can be complex, it can be dicult to objectively specify an apriorimethod for
analyzing those trajectories. Many studies therefore adopt an ad hoc approach: visualize the trajectories and
then extract some summary scalar — which was not specified prior to the experiment — to test statistically.
Unfortunately this approach is biased for the following reasons: all statistical analyses require a model of
randomness — from that model one computes the probability that random data would produce the observed
result (i.e. the p value). If one’s apriorihypothesis pertains to zero-dimensional (0D) scalars, then a 0D
model of randomness is appropriate. However, if one’s hypothesis pertains to 1D trajectories, then objectivity
obliges one to employ a 1D model of randomness — one which describes how random 1D trajectories behave.
Since probabilistic conclusions stemming from 0D and 1D models generally dier (Pataky et al., 2013), it is
biased to test a 1D hypothesis using a 0D model.
A randomness model may separately be categorized as either parametric or non-parametric. Parametric
models are constructed by first assuming the nature of the random distribution (usually Gaussian), and then
computing a small number of parameters (usually the mean and standard deviation — SD) which charac-
terize that distribution and thus its random behavior. In contrast, non-parametric models (Good, 2005) are
generally not based on any specific distribution, and are instead constructed using experimental data. When
the data do indeed come from a Gaussian distribution, then the parametric and non-parametric models
converge (Appendix A), and when the data are non-Gaussian parametric procedures are generally not valid.
There are thus four categories of randomness models to consider: 0D parametric, 0D non-parametric, 1D
parametric and 1D non-parametric.
In the Biomechanics literature four relevant approaches have emerged: (i) error clouds — often surround-
ing a mean trajectory (McGinley et al., 2009) (ii) the bootstrap confidence interval (CI) (Olshen et al., 1989;
Lenhoet al., 1999; Peterson et al., 2000; Duhamel et al., 2004), (iii) functional data analysis (Ramsay and
Silverman, 2005), and (iv) random field theory (RFT) (Adler and Taylor, 2007; Pataky et al., 2013). Method
(ii) is non-parametric and the rest are parametric. Since (iii) and (iv) may be regarded as equivalent from
a hypothesis-testing perspective (Appendix B), this paper focusses on only RFT; RFT may be considered
simpler because it requires fewer parameters.
The theoretical inadequacy of (i) error clouds, including the SD cloud, is fortunately easy to address:
they do not stem from a randomness model. While error clouds ob jectively quantify trajectory variability,
experimental design complexities conspire to dissolve the connection between variability and probability
(Schwartz et al., 2004). Since error clouds cannot support probabilistic claims they must be regarded as
descriptive or exploratory in nature.
Unlike error clouds, (ii) CIs do stem from randomness models. Nevertheless it has been shown that 1D
CIs are invalid when based on 0D randomness (Lenhoet al., 1999; Duhamel et al., 2004). The 1D bootstrap
CI (Olshen et al., 1989; Lenhoet al., 1999; Duhamel et al., 2004) is a viable solution because it models
1D randomness in the behavior of the trajectory-wide maximum under random resamplings. The available
literature has explored 0D parametric CIs vs. 1D non-parametric CIs (Lenhoet al., 1999; Duhamel et al.,
2004; Gravel et al., 2010; Dixon et al., 2013; Cutti et al., 2014), and has also explored 0D vs. 1D hypothesis
testing using RFT (Pataky et al., 2013), but to our knowledge there has previously been no systematic
comparison of 0D vs. 1D procedures, parametric vs. non-parametric results, and CIs vs. hypothesis testing.
The primary purpose of this study was to elucidate the theoretical framework of 0D vs. 1D statistical
procedures. Specifically, we sought to clarify that choosing 0D vs. 1D procedures is statistically much more
important than choosing parametric vs. non-parametric procedures because dierences in 0D vs. 1D results
are generally much larger than dierences in parametric vs. non-parametric results. We also sought to clarify
that, in contrast to 1D CIs which are complex and non-generalizable, 1D hypothesis testing results can be
presented consistently across all experimental designs.
2 Methods
2.1 Datasets
Three simulated and three experimental datasets consisting of Jscalar trajectory responses normalized to
Qdiscrete points were analyzed (Table 1). Since the simulated datasets are artificial readers are encouraged
to judge their relevance to real data.
Datasets A and B (Fig.1) mimic a one-sample experiment. These datasets were constructed by adding
ten smoothed, amplified Gaussian noise trajectories (Fig.1a) to two true population means (Fig.1b). Dataset
B’s slightly larger signal at time=80% is evident in both the resulting datasets (Fig.1c,d) and their summary
statistics (Fig.1e,f).
Dataset C (Fig.2) mimics a regression design with one independent variable x. To ten true signals
(Fig.2a), whose maxima were perfectly correlated with x(Fig.2b), we added smooth Gaussian noise (Fig.2c)
to yield the final dataset (Fig.2d). The ten responses were divided into two groups (Fig.2a) to compare
categorical and continuous treatments of x.
Dataset D (Fig.3a) (Neptune et al., 1999) consisted of within-subject mean knee flexion trajectories in
side-shue vs. v-cut tasks during stance. Dataset E (Fig.3b) (Besier et al., 2009) consisted of stance-phase
medial gastrocnemius forces during walking in 16 Controls vs. 27 Patello-Femoral Pain (PFP) patients, as
estimated by Besier et al. from EMG-driven forward-dynamics simulations. Dataset F (Fig.3c) (Dorn et al.,
2012) consisted of left-foot anterior/posterior ground reaction forces (GRF) in one subject running/sprinting
at four dierent speeds.
2.2 General statistical calculations
For simplicity this study focusses on the t statistic. All calculations employed a Type I error rate of
=0.05 and were implemented all in Python 2.7 using Canopy 1.4 (Enthought Inc., Austin, USA) and the
open-source software package ‘spm1d’ (Pataky, 2012).
2.2.1 0D and 1D tstatistics
Definitions of 1D tstatistics are trivial extensions of their 0D definitions to a 1D domain q,whereq
represents time in the aforementioned datasets. For example, the 1D one-sample tstatistic is:
t(q)= y(q)
where y,sand Jare the sample mean, sample standard deviation, and sample size, respectively. This 1D
ttrajectory can be assembled simply by computing the tstatistic value separately at each time point q,
thereby approximating the continuous t(q) trajectory just like computing the mean separately at each point
approximates the continuous mean trajectory. Definitions of the tstatistic for other designs are provided as
Supplementary Material (Appendix C).
2.2.2 Parametric 0D and 1D critical thresholds
The critical 0D tstatistic t
0D is given as the solution to:
f0D(x)dx =(2)
where f0D(x) is the usual 0D tstatistic’s probability density function (Appendix D) and P(t>t
probability that the tstatistic will exceed t
0D if the underlying data are 0D Gaussian. In classical hypothesis
testing the null hypothesis is rejected if the observed 0D tvalue exceeds t
The critical 1D test statistic t
1D is given by RFT (Adler and Taylor, 2007) as the solution to:
Pt(q)max >t
1D=1exp Z1
f0D(x)dx ED!=(3)
where t(q)max is the maximum value of the 1D ttrajectory and where ED is the smoothness-dependent Euler
density function (Worsley et al., 2004; Friston et al., 2007). Analogous to the 0D form, Eqn.3 represents the
probability that t(q)max exceeds t
1D when the underlying data are smooth 1D Gaussian, and in classical
hypothesis testing the null hypothesis is rejected if the observed t(q)max value exceeds t
Last, we computed the critical 0D Bonferroni threshold t
0D Bonf as the solution to:
0D Bonf)=Z1
0D Bonf
f0D(x)dx =1(1 )(1/Q)(4)
Note that the Bonferroni threshold assumes Qindependent tests. For smooth 1D data this is clearly a
poor assumption because neighboring values in time are correlated. We nonetheless include t
0D Bonf in our
initial analyses to demonstrate that it is too extreme; provided the 1D trajectories are smooth, the three
critical thresholds are related as follows: t
0D <t
1D <t
0DBonf . Note that although t
0DBonf considers the
entire 1D domain q, it only uses the 0D probability density function and fails to consider 1D smoothness;
therefore only the RFT threshold (Eqn.3) is labeled “1D”.
2.2.3 Non-parametric 0D and 1D critical thresholds
Two non-parametric methods — the bootstrap and the permutation method (Good, 2005) — were used to
estimate both t
0D and t
1D. Descriptions of the 0D bootstrap and permutation methods are provided as Sup-
plementary Material (Appendix E). The 1D bootstrap is described in detail elsewhere (Lenhoet al., 1999).
The 1D permutation method followed Nichols and Holmes (2002) and is summarized in Fig.4a–c. Note that
nD parametric and nD non-parametric methods are conceptually identical in that both describe random
nD behavior to yield t
nD. Moreover, the nD non-parametric results are expected to converge to the nD
parametric results when the underlying data are nD Gaussian (Appendix A).
2.2.4 0D and 1D confidence intervals
Substituting the critical 0D threshold t
0D into Eqn.1 yields the height hof the one-sample 0D CI:
h0D =t
CI heights for two-sample and regression-designs similarly follow from the design-dependent definitions
of the tstatistic (Table 3, Appendix F). Heights of 1D CIs are given simply by substituting t
1D for t
0D in
CI height calculations.
2.3 Specific dataset analyses
For Datasets A and B we sought to compare 0D vs. 1D, parametric vs. non-parametric, bootstrap vs.
permutation and CI vs. hypothesis testing results. We thus computed seven dierent critical one-sample t
values for both datasets: (#1–#3) parametric versions of t
1D and t
0D Bonf, then both bootstrap and
permutation versions of both (#4,#5) t
0D and (#6,#7) t
1D. We then constructed the associated CIs and
qualitatively compared all CIs and hypothesis testing results.
For Dataset C we sought to demonstrate two points: (1) since this is a regression design, neither 0D
nor 1D CIs are suitable when the datum is the mean 1D trajectory, (2) unlike 1D CIs, 1D hypothesis
testing results can be presented identically across designs. We first constructed the narrowest possible CIs
(0D CIs) to emphasize that even these cannot capture probabilistic meaning in regression designs. Next
we qualitatively compared two-sample and regression hypothesis testing results for 0D, 1D parametric, 1D
non-parametric and 0D Bonferroni thresholds.
For Datasets D–F we sought to emphasize both (i) the generalizability of 1D hypothesis testing proce-
dures, and (ii) the similarities between 1D parametric and 1D non-parametric results in real 1D experimental
datasets. Since the bootstrap is unsuitable for arbitrary hypothesis testing (Good, 2005) we conducted only
1D parametric and 1D permutation tests whose results we compared qualitatively.
3 Results
3.1 0D vs. 1D methods (Datasets A and B)
The three 0D CIs were qualitatively identical, and the three 1D CIs were also qualitatively identical,
but the 0D CIs were considerably dierent from both the 1D CIs and the 0D Bonferroni CI (Fig.5a,b).
The cause of these dierences is the underlying randomness model. The 0D parametric model assumes 0D
Gaussian randomness, and the 0D non-parametric procedures discretely approximate the same randomness.
Similarly, the 1D (RFT) parametric model assumes 1D Gaussian randomness and describes the behavior
of the trajectory maximum, and the 1D non-parametric procedures discretely approximate the same 1D
The 0D Bonferroni result assumes 0D randomness, but also corrects for Q=101 independent tests across
the time domain. Since the data are temporally smooth, adjacent time samples are clearly not independent
and thus the Bonferroni correction is overly conservative as has been noted previously (Duhamel et al., 2004).
One-sample hypothesis testing results (Fig.5c,d) mirrored the CI results (Fig.5a,b). In particular, both
the 0D CIs and the 0D hypothesis testing results reached significance at approximate times of 15% and
75%. In contrast, the 1D CIs and 1D hypothesis testing results reached significance only for Dataset B and
only at 75% time (p=0.037). The Bonferroni-corrected results failed to reach significance in any dataset,
emphasizing its overly conservative nature. These results emphasize that CIs are equivalent to one-sample t
tests, and also that the threshold-crossing behavior is somewhat clearer for the hypothesis tests (Fig.5d).
3.2 CIs vs. hypothesis tests (Dataset C)
One-sample 0D CIs failed to separate the groups (Fig.6a). Nevertheless both 0D regression (Fig.6b) and
a 0D two-sample test (Fig.6c – lower threshold) reached significance. This disagreement is explained by
the CI’s complex design-dependence (Table 3). In contrast to CIs, both the two-sample test and regression
results could be presented in an identical, unambiguous format as a t trajectory with critical thresholds
(Fig.6c,d). This result emphasizes that 1D CIs are less generalizable than 1D hypothesis testing.
Note that, even if the 0D CIs in Fig.6a had been constructed more robustly using a 1D two-sample
model, the results would be incorrect because the independent variable (x) is continuous. In this case the
two-sample results fail to reach significance (Fig.6c) but the regression results do (Fig.6d).
3.3 Parametric vs. non-parametric 1D methods (Datsets D–F)
For each of the experimental datasets, 1D parametric and 1D non-parametric results were qualitatively
identical (Fig.7). In particular, (i) the null hypothesis was rejected in all cases, (ii) essentially the same supra-
threshold temporal windows were identified, and (iii) similar probabilities were obtained for suprathreshold
clusters. Unlike CI results, these results are reportable in an identical manner across arbitrary experimental
designs, further emphasizing the generalizability of 1D hypothesis testing.
4 Discussion
4.1 0D vs. 1D methods
This study’s results suggest most broadly that choosing between 0D and 1D methods is likely much more
important than choosing between parametric and non-parametric methods when analyzing 1D biomechan-
ical data. From the discrepancies amongst the 0D and 1D results (Figs.5–6) it is clear that 0D procedures
inaccurately model smooth 1D trajectory variance (Fig.1a, Fig.2c) which characterizes most 1D biomechan-
ical datasets (Duhamel et al., 2004). One may therefore be tempted to ask: “which is the correct method?”
That question is important and easy to answer: both 0D and 1D methods are correct, but they cannot both
be simultaneously correct. Since a method’s validity rests on its assumptions’ justifiability, and since 0D
and 1D methods make dierent assumptions (i.e. 0D randomness vs. 1D randomness), they cannot both be
valid for the same dataset. A 0D procedure is perfectly justifiable if one formulates a specific 0D hypothesis
prior to conducting a 1D experiment, and then analyzes only those specific 0D data (Pataky et al., 2013); in
this case 1D probabilistic methods would be unjustified. On the other hand, if one does not have a specific
0D hypothesis, then by definition one’s hypothesis implicitly pertains to the whole 1D trajectory, in which
case we’d argue that only 1D procedures are justifiable. More simply, one’s apriorihypothesis must drive
one’s analysis and not the other way around.
4.2 Parametric vs. non-parametric procedures
The choice between parametric and non-parametric procedures had negligible eects on the current results
(Fig.5–7) suggesting that RFT’s assumption of 1D Gaussian randomness was a reasonable one. We have
separately observed similar agreement between parametric and non-parametric 1D procedures for a much
greater variety of 1D Biomechanics data, including EMG time series (Robinson et al., 2015), suggesting that
the choice between 0D and 1D models appears to be more important than the choice between parametric
and non-parametric models.
The main advantage of (parametric) RFT procedures is that, since they assume an analytical model of
1D randomness, they are very fast. Non-parametric procedures are generally much slower because they build
randomness models iteratively based on experimental data. As examples, for the relatively small Dataset B
our RFT and 1D permutation implementations required an average of 0.020 s and 0.130 s, respectively. For
the larger Dataset E, the durations were 0.023 s and 5.5 s, respectively.
The main disadvantage of RFT procedures is that, like 0D parametric procedures since it assumes a
Gaussian model of randomness, and that assumption may be violated. One should therefore check adherence
to the normality assumption when employing parametric procedures, either explicitly through a test for
normality, or implicitly by checking for agreement between parametric and non-parametric results. However,
such normality checks may be moot: 1D biomechanical trajectories are generally smoothed prior to analysis
(Bisseling and Hof, 2006; Kristianslund et al., 2013), and smoothing, by definition, mitigates outliers and
drives the data toward normality. Non-parametric 1D procedures are generally valid irrespective of the
underlying distribution.
A second disadvantage is that parametric procedures are less flexible than non-parametric procedures. In
particular, it has been shown that SD continuum smoothing can enhance the signal:noise ratio because point-
by-point SD estimations are generally poor, especially for small sample sizes (Nichols and Holmes, 2002).
Such smoothing is valid for non-parametric but not parametric procedures.
4.3 CIs vs. hypothesis tests
The present CI results (Fig.5a,b) agree with previous findings that 1D CIs better model 1D vari-
ance than do 0D CIs (Lenhoet al., 1999; Duhamel et al., 2004; Gravel et al., 2010; Cutti et al., 2014). Al-
though those studies’ 1D methods were limited to the 1D bootstrap, our results suggest that the 1D CI can
also be constructed in at least two additional ways: parametrically using RFT, and non-parametrically using
the permutation procedure of Nichols and Holmes (2002).
Also unlike previous studies, this study’s results (Figs.5–6) suggest that 1D CIs are a poorer choice than
hypothesis tests, primarily because CIs are suitable only for very simple one- and two-sample designs. Even
within those simple designs, CIs have complex design- and datum-dependent interpretations (Table 3), so
when reporting 1D CIs graphically one must explicitly specify both the design and the datum one employed
to construct the CI. We’d argue that this unnecessarily complicates cross-study comparisons. In contrast, 1D
hypothesis testing accommodates arbitrary experimental designs and yet presents 1D results in a much more
consistent manner across studies (Fig.7). It has been shown elsewhere that 1D hypothesis testing results can
be presented identically for multivariate (vector) trajectories (Pataky et al., 2013) and thus most generally
to MANCOVA (Worsley et al., 2004).
The primary advantage of CIs is that they present probabilistic results in the context of the original
data, with identical units (Batterham and Hopkins, 2006). This clearly makes the CI valuable for data
visualization and exploration. However, since CIs embody no unique probabilistic information relative to
hypothesis testing (Table 3), and since CIs are dicult or impossible to interpret in arbitrary experimental
designs (Fig.6), we’d argue that hypothesis testing should preferentially be adopted where possible.
4.4 Limitations of 1D methods
A key assumption of all 1D methods is that trajectories have been appropriately smoothed and regis-
tered (i.e. temporally normalized) (Sadeghi et al., 2003). This may be important considering that smoothing
algorithm particulars can non-trivially aect biomechanical interpretations (Bisseling and Hof, 2006; Kris-
tianslund et al., 2013), and that nonlinear registration procedures can substantially reduce 1D trajectory
variability (Sadeghi et al., 2003). Nevertheless, since these assumptions pertain to data processing and not to
statistical inference, they are not unique to 1D analyses, so should be scrutinized for both 0D and 1D analyses.
As a rule of thumb, if one is confident that one’s mean trajectories are unbiased by smoothing/registration
particulars, then by definition 1D inference procedures are valid.
As an anecdotal exploration of (mis-)registration eects, consider that Dataset F appears to contain
misregistered early-stance posterior GRF extrema (Fig.3c). In this particular case adopting a nonlinear
registration procedure has only moderate quantitative eects on the results and no real qualitative eect
(Appendix G). Nevertheless, registration — and more generally the assumption of data homology — requires
continued scrutiny for both 0D and 1D analyses.
Partially mitigating both smoothing and registration-related eects is RFT’s generalizability to nD con-
tinua (Friston et al., 2007; Pataky, 2010). Since both smoothing and registration are generally parameteriz-
able (e.g. smoothing kernel width) the 1D test statistic continuum can be extended to (K+1) dimensions,
where Kis the number of smoothing/registration parameters (Worsley et al., 1996). Analysis of the re-
sulting (K+1)-dimensional test statistic continuum would constitute a systematic sensitivity analysis of
smoothing/registration assumptions.
Last, a potentially serious limitation of 1D methods exists for routine biomechanical analyses. Many
studies measure a variety of variables including, for example: 3D angles at multiple joints, 3D reaction
forces, and electromyographical time series. While 1D methods can handle multivariate trajectories in
general (Pataky et al., 2013), the main problem is that statistical power reduces as the number of 0D or 1D
variables increases and the sample size remains small. There is no statistical theory of which we are aware
that can maintain statistical power in the face of both small sample sizes and an arbitrarily large barrage
of 1D measurements. Exploratory analyses (e.g. 1D mean and SD interpretations) may be necessary to
formulate specific, feasibly testable hypotheses regarding sub-components of such datasets.
4.5 Summary
This study’s results suggest that 0D methods inaccurately model the behavior of smooth, random 1D
trajectories. Since one’s primary scientific reporting obligation is to specify the probability with which
random data could produce the observed result, these results also suggest that 1D methods should be used
to analyze 1D data except when one has a specific 0D hypothesis prior to conducting an experiment. Finally,
as compared with 1D CIs, 1D hypothesis tests represent a simpler, more generalizable basis for forming
probabilistic conclusions regarding smooth 1D biomechanical trajectories. While parametric 1D (RFT)
procedures may be preferable because of their speed, non-parametric 1D procedures may be necessary when
deviations from normality are non-negligible.
We wish to thank Phil Dixon for helpful discussions pertaining to non-parametric 1D analyses.
Conflict of Interest
The authors report no conflict of interest, financial or otherwise.
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Table 1: Dataset overview. Jand Qare the numbers of responses and time nodes, respectively.
Dataset J Q Model Link
A 10 101 One-sample t test
B 10 101 One-sample t test
C 10 101 Linear regression
D 8 101 Paired t test
E 43 100 Two-sample t test
F 8 100 Linear regression
Table 2: Statistical procedures and randomness models used in this paper. Here µand
are the population mean and standard deviation, respectively. The parameters Qand Ware
the number of trajectory nodes and the trajectory smoothness, respectively (see text). The
Bonferroni procedure assumes Qindependent tests, and the RFT procedure assumes Q/W
independent trajectory processes.
Number Class Procedure Randomness model Parameters
1 Parametric Uncorrected 0D Gaussian µ,
2 Parametric Bonferroni-corrected 0D Gaussian µ,,Q
3 Parametric RFT-corrected 1D Gaussian µ,,Q/W
4 Non-parametric Uncorrected bootstrap 0D empirical None
5 Non-parametric Uncorrected permutation 0D empirical None
6 Non-parametric Corrected bootstrap 1D empirical None
7 Non-parametric Corrected permutation 1D empirical None
Table 3: Significance threshold definitions for confidence intervals (CIs) and hypothesis tests
(see also Appendix F). The Type I error rate defines the critical threshold twhich, in turn,
defines the design-dependent CI height hthat is added to a datum: either one sample’s mean
(yA) or the mean dierence (y). Paired and two-sample t tests assume yAyB. Regression
CIs are possible only when the datum is the regression slope or intercept. The key point is
that, while the CI height is design dependent and the datum ambiguous, the hypothesis testing
threshold is always tand its datum is always zero.
Confidence intervals Hypothesis tests
Datum: yAy0
One-sample t test yAh1>0t1>t
Paired t test
yAhp> yB
2hp> yB+1
Two-sample t test
yAh2> yB
2h2> yB+1
Regression tr>t
Figure 1. Datasets A and B (both simulated). (a) Smooth 1D Gaussian noise (FWHM=25%). (b)
True population mean trajectories. (c,d) Final datasets: sum of true signals and noise. (e,f)
Summary statistics: means with SD clouds.
Figure 2. Dataset C (simulated). (a,b) True simulated signals exhibiting a perfect linear
correlation between the independent variable and signal maxima; data are divided into two
groups for a subsequent comparison between a two-sample t test and regression. (c) Smooth 1D
Gaussian noise (FWHM=25%). (d) Final dataset: sum of true signals and noise.
Figure 3. Experimental datasets. (a) Dataset D (Neptune et al. 1999): cross-subject mean knee
angle trajectories with SD clouds in side-shuffle vs. v-cut maneuvers. (b) Dataset E (Besier et al.
2009): cross-subject mean medial gastrocnemius force trajectory, as estimated from dynamic
simulation, with SD clouds. (c) Dataset F (Dorn et al. 2012): cross-trial horizontal ground
reaction force trajectories in one subject during running/sprinting at various speeds.
Figure 4. Non-parametric inference overview. (a) Original simulated data and two other of the 20
total permutations; the 20th permutation is the opposite of the original. (b) Test statistic (t)
trajectories for each permutation. (c) The maximum t value from each t trajectory forms the
primary permutation PDF, from which the critical value t* was computed as the 95th percentile to
ensure that only α=5% of all permutations exceed t*. (d) The original t trajectory exceeds t*,
which provides sufficient evidence to reject the null hypothesis. To qualify the rejection decision,
the maximum suprathreshold cluster integral from each t trajectory was extracted to form a
secondary permutation PDF, from which specific cluster-level p values were computed. (e) Final
hypothesis testing results. Here the original t trajectory was the only one of all 20 permutations to
produce a suprathreshold cluster, so that cluster’s p value is 1/20=0.05. Had RFT-based
parametric inference been conducted on these data the results would have been: t*=5.303,
Figure 5. Results for Datasets A and B. (a,b) Seven different confidence intervals (CIs) are
depicted (see Table 1) as labeled in the two legends. Dark vertical bars highlight key temporal
windows discussed in the text. (c,d) Hypothesis testing results for four different tests; the null
hypothesis is rejected at α=0.05 if the test statistic trajectory (thick black line) traverses the
depicted threshold. In panel (d), the p value is the RFT result, describing the frequency with
which smooth Gaussian trajectories are expected to produce a supra-threshold cluster of that
temporal extent.
Figure 6. Results for Dataset C. (a) Separate one-sample CIs for Groups 1 and 2, using an
uncorrected threshold. (b) Regression results on only y maxima; these results are uncorrected
and therefore invalid if the null hypothesis pertains to the whole trajectory. (c) A two-sample t
test comparing Group 1 and 2 means. (d) Linear regression between x and y(q).
Figure 7. Hypothesis testing results for the experimental data. The top, middle and bottom panel
rows depict results for Datasets D (paired), E (two-sample) and F (regression), respectively (see
Fig.4). The left and right panel columns depict parametric and non-parametric results,
respectively. Critical thresholds (t*) and cluster-level probability values (p) are shown. The main
points are that: (i) parametric and non-parametric results are qualitatively identical, and (ii) unlike
CIs, hypothesis testing results can be presented identically for all experimental designs.
Note to readers:
This Supplementary Material was peer-reviewed along with the main manuscript, but has
not been edited by the journal. Sections appear in the order in which they are cited in the
main manuscript.
Appendix A Parametric vs. non-parametric hypothesis testing
The main dierence between parametric and non-parametric hypothesis testing is that the
former parameterizes probability density functions (PDFs) (Appendix D) and the latter does
not. This distinction exists at two levels:
Experimental data: parametric hypothesis testing assumes that the data are drawn from
a population with a known, parameterizable PDF (usually the Gaussian distribution),
but non-parametric procedures generally makes no such assumption.
Test statistic: parametric procedures base inferences on parameterized test statistic PDFs
which are analytically derived from the population PDF, but non-parametric procedures
generally base inferences on empirically derived test statistic PDFs.
Below we consider these points in detail.
Parametric PDFs
The fundamental PDF upon which most parametric inference is based is the normal (Gaus-
sian) distribution, which is parameterized by the true population mean µand true population
standard deviation (Fig.A1):
f(x)= 1
Figure A1: Gaussian probability density functions.
If we assign numerical values to µand , then we can compute arbitrary probabilities using
Eqns.D.1 and D.2 (Appendix D). For example, if µ=0 and =1, then the survival function
(Eqn.D.2) predicts P(x>0.0)=0.500 and P(x>2.0)=0.023. These probabilities respectively
imply that 50% of random values drawn from this distribution are expected to be greater
than zero, and only 2.3% are expected to be greater than 2.0. To re-emphasize the meaning of
‘parametric’, we note that two simple parameters (µand ) completely specify the probabilistic
behavior of Gaussian data.
The Gaussian PDF (Eqn.A.1) is nevertheless seldom used directly when conducting statis-
tical inference. One reason is that the Gaussian PDF describes a random variable x,whichis
analogous to the raw data we measure experimentally. Most experiments are less interested in
xitself than in averages (one-sample tests), average dierences (two-sample tests), and corre-
lations between xand an independent variable (regression tests). To address these empirical
pursuits, the parametric approach funnels the Gaussian PDF into a particular experimental
design, and generates predictions regarding what Gaussian data would do in that particular
setting, over an infinite number of identical experiments.
Another reason the Gaussian PDF is not used directly for statistical inference is that µand
are true population parameters, but we rarely know these true values because we rarely have
access to the entire population. We instead have to estimate µand using a sample drawn
from that population, but those estimates are imperfect, especially if the data are not sampled
randomly. Even when the data are sampled randomly, estimates of µand worsen as sample
size decreases (Fig.A2), and parametric inference must account for this sample size-dependent
behavior. Student solved this problem in 1908 through use of PDF which depends only on
sample size:
f(x)= +1
Here is the degrees of freedom and is the gamma function. The parameter specifies
the number of values which can vary freely in a particular statistic’s computation. For example,
in the one-sample t test (Table F2) there are Jresponses, but not all response values can vary
freely. In particular, after one estimates the mean, there are only (J1) responses which can
vary freely to produce the same mean, so the SD estimate is normalized using (J1) rather
than J(Table F2).
Equation A.2 is the analytical result obtained when Gaussian data (Eqn.A.1) are funneled
into tstatistic equations (Table F2). In other words, Gaussian data behave in a sample-size
dependent manner (Fig.A3) when the sample is smaller than the population size.
Figure A2: Variability of population parameter estimates as a function of sample size. The
true mean and SD were 0 and 1, respectively. These results were constructed by simulating
106samples of each sample size, computing each sample’s mean and SD, then computing the
SD of each parameter across all 106samples.
Figure A3: Comparison of various tPDFs with the standard normal PDF (µ=0, =1). The
PDFs in panels (a) and (b) are identical, but panel (b) zooms in on one part of the PDF for
The tPDF approaches the standard normal PDF (µ=0, =1) as increases (Fig.A3b).
Equivalently and conversely, large tvalues become increasingly likely as sample size decreases.
Although the eect of may appear small in Fig.A3, consider the following numerical results:
P(x>3.0)=0.020 when =4, but P(x>3.0)=0.00384 when =18. This implies that Gaussian
data are approximately five times more likely to produce tvalues larger than 3.0 for =4 vs.
Last, let us consider a full numerical example, which we shall repeat with non-parametric
analyses below. Imagine that an experiment yields Group A and Group B responses of {1.14,
1.21, 1.25, 1.43, 1.57}and {1.37, 1.52, 1.61, 1.74, 1.54}, respectively. A two-sample independent
ttest (=8) yields t=2.378. From Eqns.D.2 and A.2 we may conclude that Gaussian data are
expected to produce a tvalue this large with a probability of p=0.022 over many random
To summarize, the tstatistic’s PDF (Eqn.A.2) is completely specified by one parameter: ,
and that PDF is derived from the Gaussian PDF (Eqn.A.1), which is also parametric. More
generally, parametric procedures use a small number of parameters to specify both the PDF
from which experimental data are assumed to have been randomly drawn, and the test statistic
PDF upon which statistical inference is based.
Non-parametric PDFs
Non-parametric PDFs are identical to parametric PDFs in the sense that they describe
the behavior of randomly sampled data. The main dierence is that non-parametric PDFs
generally make no assumptions regarding the distribution from which data are drawn, and
instead build PDFs empirically, directly from experimental data. If the underlying data are in
fact Gaussian distributed, then non-parametric PDFs converge to parametric PDFs (Fig.A4)
and non-parametric results converge to parametric results (Appendix E). If experimental data
deviate from Gaussian behavior then parametric approaches based on the Gaussian PDF (like
the tPDF) are generally not valid.
To emphasize these points it is sucient to describe one non-parametric approach to PDF
construction. Below we describe a simple two-sample permutation procedure similar to the one
used in the main manuscript, but somewhat dierent from the one-sample procedure described
in Appendix E . Returning to the numerical example above, the two-sample permutation
approach starts by labeling the original data as follows:
Value 1.14 1.21 1.25 1.43 1.57 1.37 1.52 1.61 1.74 1.54
As we saw before, this particular labeling (AAAAA–BBBBB) yields t=2.378. To build the
permutation PDF, we simply permute these ten labels and recompute the tstatistic for each
permutation. For example, labels of BAAAA–ABBBB and BBAAA–AABBB yield t=1.208
and t=0.154, respectively. Repeating for many or all label permutations builds a permutation
PDF (Fig.A4). In this example there are ten labels, but once we choose positions for the five
A labels, the positions of the five B labels are decided. There are thus 10
5= 10!/(5!5!) =
252 unique permutations. Assembling all or a large number of permutation tvalues forms
a permutation PDF (or empirical PDF), from which probability values can be computed as
P(tu)=Number of permutation values greater than or equal to u
Number of permutations (A.3)
Since this example has 252 permutations, the minimum possible pvalue is 1/252 = 0.004.
Of those 252 permutations, this example yields a total of eight which satisfy tu, includ-
ing a maximum tvalue of 4.804 for a labeling of: AAAAB–ABBBB. Thus the pvalue is
8/252=0.0318, which is similar to the parametric pvalue of 0.022. This indirectly suggests
that the parametric approach’s assumption of normality is a reasonable one.
Which pvalue is correct, the parametric or non-parametric one? Both are correct, but their
meanings are dierent. The interpretation of the parametric pvalue is as follows: if there were
truly no dierence between Groups A and B and if the population data are Gaussian distributed
then a tvalue as large as the observed value would be expected in 2.2% of an infinite number of
identical experiments. The interpretation of the non-parametric pvalue is: if there were truly
no dierence between Groups A and B and the group labels were assigned randomly to the
data then only 3.18% of relabelings would yield a tvalue as large as the observed value. The
primary dierence between the two approaches is thus that the parametric pvalue assumes
that the population distribution is Gaussian but the non-parametric pvalue does not.
Figure A4: Comparison of parametric and non-parametric PDFs for the two-sample ttest
example described in the text. Here =8 completely parameterizes the parametric PDF. The
non-parametric PDF is a histogram of the tvalues computed from all 252 permutations.
Appendix B Functional data analysis and random field theory
Functional data analysis (FDA) (Ramsay and Silverman 2005) emerged in the 1990s as a
tool for statistically analyzing one-dimensional continua or “functions”. By regarding experi-
mentally sampled continua as continuous functions, FDA shows that experimental data can be
well-approximated by a set of mathematically precise basis functions, including for example:
splines and Fourier series. Representing the data in this manner opens up a wide range of
analysis possibilities for describing continua, covariance between continua, etc. Although FDA
was initially developed primarily as an exploratory tool of 1D continuum variance, over the
years it has expanded to a wide array of statistical uses including classical hypothesis testing
in arbitrary experimental designs through a variety of inference techniques.
Random field theory (RFT) (Adler and Taylor 2007) was initially developed in the 1970s
to extend the (0D) Gaussian distribution to n-dimensional continua with arbitrary geometrical
bounds. RFT shows, for example, how smooth 1D Gaussian continua exhibit particular geomet-
ric features (like maximum continuum height) with known probability. Statistical Parametric
Mapping (SPM) emerged in the 1990s to apply RFT to experimentally measured continua
(Friston et al. 2007). In the case of unbroken 1D continua, SPM estimates just one parameter
more than is estimated for common 0D analyses — the ratio of continuum length to smooth-
ness — then uses RFT to make probabilistic conclusions, like the probability that smooth 1D
Gaussian data will yield a tcontinuum which reaches a height of 3.0 in a two-sample experi-
ment. Directly related to classical hypothesis testing, SPM can use RFT to compute the critical
height tabove which only % of tcontinua would reach if those tcontinua were produced by
smooth 1D Gaussian continua in an infinite number of identical experiments.
From a classical hypothesis testing perspective for 1D data, there is thus only one dierence
between FDA and RFT. Whereas FDA inference procedures are widely flexible, with a varying
number of parameters, RFT inference is based on a single parameter: the continuum length-
to-smoothness ratio. Since they can both describe random 1D continua, they may be regarded
as equivalent for the purposes of the present paper. The main manuscript focusses on separate
issues: 0D vs. 1D, parametric vs. non-parametric, and confidence interval vs. hypothesis
testing procedures. While we could have used FDA address these issues, we opted for RFT
simply because we find RFT easier to describe.
Appendix C Extending the t statistic to the time domain
The 1D t statistic is assembled simply by computing the 0D t statistic separately at each
time point q. Since all 1D t statistic definitions are therefore trivial extensions of their 0D
definitions to the 1D domain q, they are listed here only for completeness. The t statistic
continua for the one-sample, paired and two-sample designs are respectively:
t(q)= y(q)
t(q)= y(q)
t(q)= y(q)
For regression against a continuous independent variable x,themodelis:
where 1,0and "are the slope, intercept and prediction error, respectively. Least-squares
estimates of the slope and intercept (denoted ˆ
1and ˆ
0, respectively) produce the following
prediction for the jth response:
ˆyj(q)= ˆ
and the standard error is:
Finally, the regression tstatistic is:
t(q)= ˆ
Appendix D Probability density functions (PDFs)
A PDF is a continuous function f(x) which, when integrated over an interval [x0,x1],
specifies the probability that a random variable xadopts a value in that interval:
f(x)dx (D.1)
The probability that xadopts a specific value ˆxis zero because there are an infinite number
of other values it could adopt. The probability that xlies in the interval [x0,x1] is at least zero
and at most one. All PDFs additionally share the trivial constraint that xlies in the interval
[1,1]. These three constraints can be expressed as follows:
P(1 <x<1)=1
The key probability for classical hypothesis testing is the survival function — the probability
that xexceeds (or ‘survives’) an arbitrary threshold u:
f(x)dx (D.2)
When Eqn.D.2 is set to , then ubecomes a “critical threshold”; an experimentally observed
value ˆxwhich exceeds this threshold leads to null hypothesis rejection.
Random Field Theory (RFT) (Adler and Taylor, 2007) provides the foundation for gener-
alizing Eqn.D.2 to the case of Gaussian nD continua. An important RFT probability is:
P(xmax >u)=Z1
f(x)dx (D.3)
where xmax is the maximum continuum value. For classical hypothesis testing on 1D continua,
setting Eqn.D.3 to and solving for uyields the critical threshold for the null hypothesis
rejection decision.
Appendix E Bootstrap and permutation techniques
The purpose of this appendix is to clarify (a) the similarities and dierences between the
bootstrap and permutation confidence intervals (CIs), and (b) the role of both techniques
in the broader context of parametric and non-parametric hypothesis testing. Note that the
bootstrap has been advocated in the Biomechanics literature for trajectory-level analysis. The
permutation technique is used in the main manuscript because it is more generalizable than the
bootstrap. Interested readers may wish to consult Good (2005) for a more thorough treatment
of these topics for 0D datasets, and to Nichols and Holmes (2002) for a discussion of how these
techniques extend to 1D and higher-dimensional data.
Sections E.1 and E.2 below analyze the following eight-response dataset:
117 104 110 122 119 90 110 97
Section E.1 computes CIs for this dataset using three dierent techniques, and Section
E.2 conducts one-sample hypothesis testing using four dierent techniques. Table E1 below
summarizes the results of those analyses. Considering these results briefly, it is clear that all
techniques produce similar, albeit non-identical CIs and p values. To emphasize why these
results are similar but not identical, Section E.3 repeats the three CI techniques for thousands
of random (Gaussian) datasets to demonstrate why all techniques may be regarded as theoret-
ically equivalent when the data are normally distributed. This Appendix thus shows that it
is sucient in the main manuscript to compare a single parametric technique (which assumes
normality) to a single non-parametric technique (which does not assume normality).
Table E1: Confidence intervals (CI) and one-sample hypothesis tests computed using four
dierent techniques, based on the dataset above.
Class Techni que 95% CI One-sample test
Non-parametric Bootstrap [ 98.4, 117.5 ] p= 0.07559
Non-parametric Permutation [ 98.9, 118.3 ] p= 0.06250
Non-parametric Wlicoxon p= 0.06735
Parametric Student’s t [ 99.3, 117.9 ] p= 0.06411
E.1 Confidence intervals (CIs)
E.1.1 Bootstrap CI
A simple bootstrap CI can be computed as follows:
(a) Compute the sample mean (in this case: 108.625).
(b) Label the responses as follows:
117 104 110 122 119 90 110 97
(c) Resample with replacement: select a random set of labels, allowing labels to repeat, then
compute the mean for the resampled data. For example, a labeling of “AABBBCDE” has
responses: [117, 117, 104, 104, 104, 110, 122, 119], and a sample mean of: 112.125.
(d) Repeat (c) many times and store all sample means. Stop either when (i) all possible
resamplings have been made (i.e. AAAAAAAA through HHHHHHHH), or when (ii) a
specified number of iterations (e.g. 1000) has been completed.
(e) After all sample means have been accumulated, find the value Cupper above which only
2.5% of all estimates traverse, and the value Clower below which only 2.5% of all estimates
traverse. The CI is [Clower,Cupper].
As specified in Table E1 above this procedure yields a CI of [98.4, 117.5].
E.1.2 Permutation CI
A simple permutation CI can be computed as follows:
(a) Compute the sample mean (in this case: 108.625) .
(b) Subtract the sample mean from all responses, then label each observation as “+1”:
+1 +1 +1 +1 +1 +1 +1 +1
8.375 –4.625 1.375 13.375 10.375 –18.625 1.375 –11.625
(c) Resample without replacement: permute using either a “+1” or a “–1” label for each
response, then multiply each response by each label. For example, a labeling of “+1 +1
+1 –1 –1 –1 +1 –1” produces the new sample “8.375 –4.625 1.375 –13.375 –10.375 18.625
1.375 11.625”. For each new sample compute the one-sample t statistic. If there are n
responses, there are 2npossible labelings (256 in this case).
(d) Repeat (c) many times and store all t statistic values for all resamplings. Stop either when
(i) all possible resamplings have been made (i.e. “+1 +1 +1 +1 +1 +1 +1 +1” through
“–1 –1 –1 –1 –1 –1 –1 –1”), or when (ii) a specified number of iterations (e.g. 1000) has
been completed.
(e) After all t statistic values have been accumulated, find the critical height above which only
2.5% of t statistic values traverse, then compute the CI according to Appendix F.
This results in a CI of [98.9, 118.3] (Table E1).
E.1.3 Parametric CI
The parametric CI can be computed using the critical height h, which is defined via the
one-sample t statistic distribution (see Appendix F, and in particular the “One-sample” row
of Table F3). This procedure yields a CI of [99.3, 117.9], which is very similar to both the
bootstrap and permutation results.
E.1.4 Comparison of CI results
All three techniques yield similar, but non-identical results. Since the parametric technique
assumes that the data come from a normal (Gaussian) distribution, all CI techniques should, by
definition, converge to the identical value when (a) the data are normal and (b) the sample size
is large. The dierent techniques will only produce precisely the same result when the sample
size is infinitely large, as we will see in Section E.3 below. Investigators must therefore judge
whether the discrepancies amongst the techniques is negligible or non-negligible. Evidence
of departure from normality, for example, would be a good reason to choose one of the non-
parametric techniques. For the results above (Table E1), the discrepancies amongst the dierent
CI techniques are likely negligible for most applications. The main point is that the three CI
techniques are theoretically equivalent when the data are normally distributed.
E.2 Hypothesis tests
Thorough descriptions of one-sample hypothesis tests using the bootstrap, permutation,
Wilcoxon and parametric (one-sample t test) techniques can be found in many statistics text-
books so in interest of brevity are not repeated here. Additionally, as will be shown in Appendix
F, CIs are equivalent to one-sample hypothesis tests, so re-describing the techniques here would
be redundant. This section therefore just focusses on the results in Table E1 above.
Like the CI results, all four hypothesis test procedures produce similar, but non-identical
pvalues. For classical hypothesis testing, the null hypothesis would not be rejected for any of
the four tests at =0.05. The next section explores why these four approaches are theoretically
equivalent (when the data are normal) even when the results are not precisely equivalent.
E.3 Convergence of CIs
Repeating the bootstrap, permutation and parametric CI procedures on thousands of ran-
dom datasets (drawn from the Gaussian distribution) of increasingly larger sample sizes yields
the results in Fig.E1. The two non-parametric CIs clearly converge to the parametric CI as
sample size increases, implying theoretical equivalence amongst the three procedures (when the
data are normally distributed).
Figure E1: Convergence of the CI for three estimation procedures. These results were obtained
by: (i) producing a random sample from the Gaussian distribution of the given sample size,
with a mean of 100 and a variance of 10, (ii) estimating the CI using the three procedures
indicated (Bootstrap, Permutation, Parametric), and (iii) repeating 500 times for each sample
size. Single results depict the mean values across the 500 repetitions.
This Appendix has shown that there is fundamentally little dierence between the bootstrap
and permutation approaches, but that they might produce non-negligibly dierent numerical
results in certain situations, like when sample sizes are very small. The larger point is that the
bootstrap procedure is not particularly unique, as has been implied in the literature. Instead
the bootstrap procedure yields a solution which can also be obtained using other techniques,
and its scope is also relatively limited in the broader context of generalized hypothesis testing
Parametric Non-parametric
Student’s t Permutation
Student’s t Permutation
Student’s t Permutation
Student’s t Permutation
Snedecor F
(No CI)
Figure E2: Context of the bootstrap (for both 0D and 1D tests). Light grey, white and
dark grey boxes respectively depict: experimental designs, statistical inference procedures, and
confidence intervals (CI).
Appendix F Confidence interval design dependence
Confidence intervals (CIs) are defined as:
CI =y0±h(F.1)
where y0is a datum and his the design-dependent critical height. More specifically, his
given by a critical t value and simple algebraic manipulation of the design-dependent t statistic
To clarify, first consider that design-dependent mean and SD definitions (Table F1) yield
design-dependent t statistic definitions (Table F2). Next, given tone may compute the design-
dependent h(Table F2). Last, after choosing a datum y0, there are various acceptable null
hypothesis rejection criteria (Table F3).
The main point is that hypothesis testing employs a single unambiguous criterion: (t>t
irrespective of the particular design, making it easy to compare results across experiments. In
contrast, CIs are both design- and datum-dependent.
Clearly his valuable for data visualization because it represents the null hypothesis rejec-
tion criterion in the same units as the original data. However, it is also clear that hmust be
computed with careful attention to both the datum and the design, and can only be interpreted
by readers if the precise datum and design are made explicit. The main manuscript therefore
argues that hypothesis testing is simpler.
Table F1: Mean and standard deviation (SD) definitions for one-sample, paired and two-sample
designs. For simplicity equal variance is assumed in the two-sample case.
Design Mean SD
One-sample y=1
Paired y=1
JP(yAj yBj)sp=r1
J1P(yAj yBj)y2
Two-sample y=yAyBs2=s(JA1)s2
Table F2: Design-dependence of the CI’s critical height h.
Design tMean h
One-sample t1=y
Paired tp=y
Two-sample t2=y
Table F3: Design- and datum-dependence of h-based null hypothesis rejection criteria. All
criteria assume yAyB.
Design Datum (y0) Criterion: zero Criterion: mean Criterion: tail
One-sample y y h
1>0— —
Paired yyh
p>0— —
p> yByAh
2> yB+h
Two-sample yyh
2>0— —
2> yByAh
2> yB+h
Appendix G Dataset F reanalyses
Here we reanalyze the Dataset F dataset using (i) two-tailed inference, and (ii) nonlinear
registration followed by two-tailed inference. Regarding two-tailed inference: the results in
the main manuscript (Fig.7De,f) are based on one-tailed inference at =0.05. One-tailed
inference has only a single positive critical threshold (i.e. t>0). Two-tailed inference has
two thresholds: +tand t, and excursion beyond either threshold warrants null hypothesis
rejection. Moreover, +tis higher than in a one-sample test; +tfor two-tailed inference at
is equivalent to tfor one-tailed inference at /2. Results suggest that, although two-tailed
found highlighted a temporal of significant negative correlation between GRF and walking speed
(Fig.G2a), two-tailed inference did not aect the main manuscript’s null-hypothesis rejection
Regarding nonlinear registration: we used a simple piecewise linear registration approach
(Kneip et al. 2000) to align the first two local extrema in Datasets F, which occurred be-
tween approximately 10% and 25% stance (Fig.G1a), resulting in reduced temporal variability
of those extrema (Fig.G1b). Non-linear registration also produced slightly amplified supra-
threshold tsignals with respect to the original data. However, this aected neither the null
hypothesis rejection decision nor the general biomechanical interpretation. In particular, both
original and registered results suggest negative correlation between posterior GRF and running
speed in early stance, and positive correlation in late stance. Misregistration eects may be
non-negligible in other datasets (Sadeghi et al.2003).
Kneip A, Li X, MacGibbon KB (2000). Curve registration by local regression, Canadian Journal of Statistics
28(1): 19–29.
Sadeghi H, Mathieu PA, Sadeghi S, Labelle H (2003). Continuous curve registration as an intertrial gait
variability reduction technique, IEEE Transactions on Neural Systems and Rehabilitation Engineering 11(1):
Figure G1: Dataset F, (a) original and (b) registered trajectories.
Figure G2: Dataset F, two-tailed results for (a) original and (b) registered data.
... package in Matlab. SPM allows the generalization of classical statistical tests to time series data, so that regions of signi cant difference in trajectories instead of singular points of interest can be investigated [41]. For transition analysis, paired t-tests across the ten subjects were conducted between LW and TR, and then between TR and SW, with α = 0.05 and a Bonferroni correction for multiple comparisons. ...
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Human ambulation is typically characterized during steady-state isolated tasks (e.g., walking, running, stair ambulation). However, general human locomotion comprises continuous adaptation to the varied terrains encountered during activities of daily life. To fill an important gap in knowledge that may lead to improved therapeutic and device interventions for mobility-impaired individuals, it is vital to identify how the mechanics of individuals change as they transition between different ambulatory tasks, and as they encounter terrains of differing severity. In this work, we study lower-limb joint kinematics during the transitions between level walking and stair ascent and descent over a range of stair inclination angles. Using statistical parametric mapping, we identify where and when the kinematics of transitions are unique from the adjacent steady-state tasks. Results show unique transition kinematics primarily in the swing phase, which are sensitive to stair inclination. We also train Gaussian process regression models for each joint to predict joint angles given the gait phase, stair inclination, and ambulation context (transition type, ascent/descent), demonstrating a mathematical modeling approach that successfully incorporates terrain transitions and severity. The results of this work further our understanding of transitory human biomechanics and motivate the incorporation of transition-specific control models into mobility-assistive technology.
... The D' Agostino-Pearson K2 test was used to assess the time series data normality. Data were not normally distributed; therefore, the non-parametric version of vector field analysis, statistical non-parametric mapping (SnPM) was used [24]. SnPM paired t-tests (p < 0.05), with the number multi-segment foot model and the Oxford Foot Model (OFM) applied simultaneously. ...
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Background Different multi-segment foot models have been used to explore the effect of foot orthoses. Previous studies have compared the kinematic output of different multi-segment foot models, however, no study has explored if different multi-segment foot models detect similar kinematic changes when wearing a foot orthoses. The aim of this study was to compare the ability of two different multi-segment foot models to detect kinematic changes at the hindfoot and forefoot during the single and double support phases of gait when wearing a foot orthosis. Methods Foot kinematics were collected during walking from a sample of 32 individuals with and without a foot orthosis with a medial heel bar using an eight-camera motion capture system. The Oxford Foot Model (OFM) and a multi-segment foot model using the Calibrated Anatomical System Technique (CAST) were applied simultaneously. Vector field statistical analysis was used to explore the kinematic effects of a medial heel bar using the two models, and the ability of the models to detect any changes in kinematics was compared. Results For the hindfoot, both models showed very good agreement of the effect of the foot orthosis across all three anatomical planes during the single and double support phases. However, for the forefoot, the level of agreement between the models varied with both models showing good agreement of the effect in the coronal plane but poorer agreement in the transverse and sagittal planes. Conclusions This study showed that while consistency exists across both models for the hindfoot and forefoot in the coronal plane, the forefoot in the transverse and sagittal planes showed inconsistent responses to the foot orthoses. This should be considered when interpreting the efficacy of different interventions which aim to change foot biomechanics.
... SPM1D uses random field theory to make statistical inferences at the continuum level regarding sets of one-dimensional measurements. It is based on the idea to quantify the probability that smooth, random one-dimensional continua would produce a test statistic continuum whose maximum exceeds a particular test statistic value and has been previously used to analyze one-dimensional kinematic, biomechanical or force trajectories [77,78]. In case of significant effects, post-hoc tests implemented in the SPM1D software were used which are corrected for multiple comparisons using Bonferroni corrections. ...
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The brain's state of arousal influences cognitive functioning and mental well-being. It is controlled by several neuromodulatory nuclei in the brainstem and, particularly, by the locus coeruleus (LC). The LC is the main source of noradrenaline (NA) in the central nervous system where it exerts powerful effects on neural processing and autonomic function. Here, we investigate whether human participants can gain volitional control of their brain's arousal state using a new neurofeedback approach which exploits the mechanism that the eye's pupil diameter provides an indirect readout of arousal if light conditions are controlled. We show that pupil-based neurofeedback training is essential for learning how to self-regulate pupil size. Once acquired, pupil self-regulation significantly modulates neuromodulatory brainstem centers involved in arousal control and particularly the LC-NA system when carefully measured with functional magnetic resonance imaging. Further, it modulates heart rate, a cardiovascular marker of autonomic function, and it has a significant effect on behavior and specific psychophysiological responses during an oddball task, an attention task that has been shown to be evoke stimulus-dependent LC-NA activity. Considering the modulatory effects of the LC-NA system and other arousal-regulating centers on cognitive functioning and various behaviors including stress-related responses, pupil-based neurofeedback has a tremendous potential to be translated to behavioral and clinical applications across various domains.
... If the normal distribution was satisfied, a one-way repeated measures ANOVA with one-dimensional statistical parametric mapping (SPM1d) was performed. If the distribution is not normal, a one-way repeated-measures ANOVA with one-dimensional statistical nonparametric mapping (SnPM1d) is performed [62]. In the case of significant main effects (directions), Bonferroni adjustment was used to post hoc paired comparisons of significant main effects (directions). ...
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There are still few portable methods for monitoring lower limb joint coordination during the cut-ting movements (CM). This study aims to obtain the relevant motion biomechanical parameters of the lower limb joints at 90°, 135°, and 180° CM by collecting IMU data of the human lower limbs, and utilizing the Long Short Term Memory (LSTM) deep neural network framework to predict the coordination variability of selected lower extremity couplings at the three CM directions. There was a significant(p<0.001)difference between the three couplings during the swing, espe-cially at 90° vs the other directions. At 135° and 180°, the coordination variability of cou-plings was significantly greater than at 90° (p<0.001). It is important to note that the coordination variability of Hip rotation /Knee flexion-extension was significantly higher at 90° than at 180° (p<0.001). By the LSTM, the CM coordination variability for 90° (CMC = 0.99063, RMSE = 0.02358), 135° (CMC = 0.99018, RMSE = 0.02465) and 180° (CMC = 0.99485, RMSE = 0.01771) were accurately predicted. The predictive model could be used as a reliable tool for predicting the coordination variability of different CM directions in patients or athletes and real-world open scenarios using inertial sensors.
... Whilst mean and peak kinetic and kinematic variables have been extensively reported, a more sophisticated and detailed analysis of the force-time data may provide additional insight into where the differences occur between loading conditions, and how practitioners can appropriately implement these exercises. It is recommended that when testing non-directed hypotheses involving biomechanical vector fields, researchers should implement statistical parametric mapping analysis (SPM) as it is generally biased to test one-dimensional data (1D) using zero-dimensional methods, and SPM may reduce such bias (Pataky et al., 2013(Pataky et al., , 2015(Pataky et al., , 2016. Researchers have utilized time-normalized curve analysis (sometimes termed waveform or temporal phase analysis) to assess force-, velocity-, power-and displacement-time data during weightlifting derivatives (Kipp et al., 2021;Suchomel & Sole, 2017a, 2017b and jumps (Cormie et al., 2008(Cormie et al., , 2009McMahon, Murphy et al., 2017;. ...
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The effect of load on time-series data has yet to be investigated during weightlifting derivatives. This study compared the effect of load on the force-time and velocity-time curves during the countermovement shrug (CMS). Twenty-nine males performed the CMS at relative loads of 40%, 60%, 80%, 100%, 120%, and 140% one repetition maximum (1RM) power clean (PC). A force plate measured the vertical ground reaction force (VGRF), which was used to calculate the barbell-lifter system velocity. Time-series data were normalized to 100% of the movement duration and assessed via statistical parametric mapping (SPM). SPM analysis showed greater negative velocity at heavier loads early in the unweighting phase (12-38% of the movement), and greater positive velocity at lower loads during the last 16% of the movement. Relative loads of 40% 1RM PC maximised propulsion velocity, whilst 140% 1RM maximized force. At higher loads, the braking and propulsive phases commence at an earlier percentage of the time-normalized movement, and the total absolute durations increase with load. It may be more appropriate to prescribe the CMS during a maximal strength mesocycle given the ability to use supramaximal loads. Future research should assess training at different loads on the effects of performance.
... The last approach, using i.e. a statistical technique called Statistical Parametrical Mapping (SPM), seems to be more sensitive. Pataky et al. (2015) stated that using one value may be biased. In fact it can sometimes show statistical significance when there is no 1-D effect (Pataky et al., 2016). ...
The aim of the current study was to compare the arm-stroke kinematics during maximal and sub-maximal breaststroke swimming using both discrete and continuous data analysis. Nine male breaststrokers swam 2 x 25 m with maximal and sub-maximal intensity and their full body 3-D kinematics were obtained using eight video cameras. The arm-stroke was divided into five phases: recovery, glide, out-sweep, in & down-sweep and in & up-sweep. The statistical treatment of selected discrete variables was conducted using t-test, while the analysis of their equivalent time series, when applicable, was conducted using Statistical Parametric Mapping. Sub-maximal trial, compared to maximal, presented lower swimming velocity, greater stroke length and less stroke rate. Moreover, the absolute and relative duration of the glide phase was longer, while the relative duration of all the other phases was shorter. The resultant hand velocity during the arm recovery was slower, as well as the hand velocity time series in the transverse and longitudinal axis which were slower from ∼45% to ∼60% and from ∼5% to ∼15% of the stroke cycle, respectively. Both discrete and continuous data analysis revealed that the main discriminating factor between the two conditions concerns to the adjustment of the glide and the recovery phase and consequently the continuation of the propulsive movements.
Investigating of locomotor disturbances are relevant in human injury and performance. Therefore, lower extremity kinematics were analysed in response to decelerative perturbations during running using statistical parametric mapping (SPM). 13 asymptomatic individuals (8 females & 5 males, 28±3 years, 171±9 cm, 68±10 kg) completed an 8-minute running protocol with 30 one-sided perturbations (15 each side) to generate decelerative disturbances. A 3D-motion capture system was employed to record kinematic data. Joint angles of the ankle, knee, and hip in addition to stride duration, stride length and step width were calculated for leading and trailing strides. Results were analysed descriptively, followed by SPM of paired t-tests (P<0.025). Reactively (after perturbation), perturbations caused decreased hip adduction and stride duration of the leading leg. The trailing leg reacted with ankle inversion, knee and hip flexion, hip abduction, as well as an increase in stride duration and step width (P<0.025). In preparation for perturbation, the trailing leg reduced ankle dorsiflexion, knee flexion, hip flexion, and adduction. In summary, applied perturbations produced substantial reactive (feedback) and predictive (feedforward) responses of the lower limbs, most apparent in the trailing leg.
Vibration has the potential to compromise performance in cycling. This study aimed to investigate the effects of vibration on full-body kinematics and muscle activation time series. Nineteen male amateur cyclists (mass 74.9 ± 5.9 kg, body height 1.82 ± 0.05 m, Vo2max 57 ± 9 ml/kg/min, age 27 ± 7 years) cycled (216 ± 16 W) with (Vib) and without (NoVib) vibration. Full-body kinematics and muscle activation time series were analysed. Vibration did not affect lower extremity joint kinematics significantly. The pelvic rotated with vibration towards the posterior direction (NoVib: 22.2 ± 4.8°, Vib: 23.1 ± 4.7°, p = 0.016, d = 0.20), upper body lean (NoVib: 157.8 ± 3.0°, Vib: 158.9 ± 3.4°, p = 0.001, d = 0.35) and elbow flexion (NoVib: 27.0 ± 8.2°, Vib: 29.4 ± 9.0°, p = 0.010, d = 0.28) increased significantly with vibration. The activation of lower extremity muscles (soleus, gastrocnemius lat., tibialis ant., vastus med., rectus fem., biceps fem.) increased significantly during varying phases of the crank cycle due to vibration. Vibration increased arm and shoulder muscle (triceps brachii, deltoideus pars scapularis) activation significantly over almost the entire crank cycle. The co-contraction of knee and ankle flexors and extensors (vastus med. – gastrocnemius lat., vastus med. – biceps fem., soleus – tibialis ant.) increased significantly with vibration. In conclusion vibrations influence main tasks such as propulsion and upper body stabilization on the bicycle to a different extent. The effect of vibration on the task of propulsion is limited due to unchanged lower body kinematics and only phase-specific increases of muscular activation during the crank cycle. Additional demands on upper body stabilization are indicated by adjusted upper body kinematics and increased muscle activation of the arm and shoulder muscles during major parts of the cranking cycle.
Proximal femur fractures in the elderly are associated with significant loss of independence, mobility and quality of life. This prospective study aimed to: (1) investigate gait biomechanics in intertrochanteric fracture (ITF) patients (A1 and A2 AO/OTA) managed via femoral nailing at six weeks and six months post‐operative and how these compared with similarly aged elderly controls; and (2) investigate whether femoral offset shortening (FOS) and lateral lag screw protrusion (LSP) were associated with changes in gait biomechanics at post‐operative timepoints. Hip radiographs and gait data were collected for thirty‐four patients at six weeks and six months post‐operatively. Gait data were also collected from similarly aged controls. FOS and LSP were measured from radiographs. Joint angles, external moments and powers were calculated for the hip, knee and ankle and compared between timepoints in ITF patients and healthy controls using statistical parametric mapping. The relationship between radiographic measures with gait speed, step length, peak hip abduction and maximum hip abduction moment were assessed using a Pearson correlation. External hip adduction moments and hip power generation improved in the first six months post‐operative, but differed significantly from healthy controls during single limb stance. LSP showed a moderate correlation with maximum hip abduction moment at six weeks post‐operative (r=‐0.469, p=0.048). This article is protected by copyright. All rights reserved.
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Multi-muscle EMG time-series are highly correlated and time dependent yet traditional statistical analysis of scalars from an EMG time-series fails to account for such dependencies. This paper promotes the use of SPM vector-field analysis for the generalised analysis of EMG time-series. We reanalysed a publicly available dataset of Young versus Adult EMG gait data to contrast scalar and SPM vector-field analysis. Independent scalar analyses of EMG data between 35% and 45% stance phase showed no statistical differences between the Young and Adult groups. SPM vector-field analysis did however identify statistical differences within this time period. As scalar analysis failed to consider the multi-muscle and time dependence of the EMG time-series it exhibited Type II error. SPM vector-field analysis on the other hand accounts for both dependencies whilst tightly controlling for Type I and Type II error making it highly applicable to EMG data analysis. Additionally SPM vector-field analysis is generalizable to linear and non-linear parametric and non-parametric statistical models, allowing its use under constraints that are common to electromyography and kinesiology. Copyright © 2014 Elsevier Ltd. All rights reserved.
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When investigating the dynamics of three-dimensional multi-body biomechanical systems it is often difficult to derive spatiotemporally directed predictions regarding experimentally induced effects. A paradigm of ‘non-directed’ hypothesis testing has emerged in the literature as a result. Non-directed analyses typically consist of ad hoc scalar extraction, an approach which substantially simplifies the original, highly multivariate datasets (many time points, many vector components). This paper describes a commensurately multivariate method as an alternative to scalar extraction. The method, called ‘statistical parametric mapping’ (SPM), uses random field theory to objectively identify field regions which co-vary significantly with the experimental design. We compared SPM to scalar extraction by re-analyzing three publicly available datasets: 3D knee kinematics, a ten-muscle force system, and 3D ground reaction forces. Scalar extraction was found to bias the analyses of all three datasets by failing to consider sufficient portions of the dataset, and/or by failing to consider covariance amongst vector components. SPM overcame both problems by conducting hypothesis testing at the (massively multivariate) vector trajectory level, with random field corrections simultaneously accounting for temporal correlation and vector covariance. While SPM has been widely demonstrated to be effective for analyzing 3D scalar fields, the current results are the first to demonstrate its effectiveness for 1D vector field analysis. It was concluded that SPM offers a generalized, statistically comprehensive solution to scalar extraction’s over-simplification of vector trajectories, thereby making it useful for objectively guiding analyses of complex biomechanical systems.
In an age where the amount of data collected from brain imaging is increasing constantly, it is of critical importance to analyse those data within an accepted framework to ensure proper integration and comparison of the information collected. This book describes the ideas and procedures that underlie the analysis of signals produced by the brain. The aim is to understand how the brain works, in terms of its functional architecture and dynamics. This book provides the background and methodology for the analysis of all types of brain imaging data, from functional magnetic resonance imaging to magnetoencephalography. Critically,Statistical Parametric Mappingprovides a widely accepted conceptual framework which allows treatment of all these different modalities. This rests on an understanding of the brain's functional anatomy and the way that measured signals are caused experimentally. The book takes the reader from the basic concepts underlying the analysis of neuroimaging data to cutting edge approaches that would be difficult to find in any other source. Critically, the material is presented in an incremental way so that the reader can understand the precedents for each new development. This book will be particularly useful to neuroscientists engaged in any form of brain mapping; who have to contend with the real-world problems of data analysis and understanding the techniques they are using. It is primarily a scientific treatment and a didactic introduction to the analysis of brain imaging data. It can be used as both a textbook for students and scientists starting to use the techniques, as well as a reference for practicing neuroscientists. The book also serves as a companion to the software packages that have been developed for brain imaging data analysis. * An essential reference and companion for users of the SPM software * Provides a complete description of the concepts and procedures entailed by the analysis of brain images * Offers full didactic treatment of the basic mathematics behind the analysis of brain imaging data * Stands as a compendium of all the advances in neuroimaging data analysis over the past decade * Adopts an easy to understand and incremental approach that takes the reader from basic statistics to state of the art approaches such as Variational Bayes * Structured treatment of data analysis issues that links different modalities and models * Includes a series of appendices and tutorial-style chapters that makes even the most sophisticated approaches accessible.
Quantitative motion analysis protocols have been developed to assess the coordination between scapula and humerus. However, the application of these protocols to test whether a subject's scapula resting position or pattern of coordination is "normal", is precluded by the unavailability of reference prediction intervals and bands, respectively. The aim of this study was to present such references for the "ISEO" protocol, by using the non-parametric Bootstrap approach and two parametric Gaussian methods (based on Student's T and Normal distributions). One hundred and eleven asymptomatic subjects were divided into three groups based on their age (18-30, 31-50, and 51-70). For each group, "monolateral" prediction bands and intervals were computed for the scapulo-humeral patterns and the scapula resting orientation, respectively. A fourth group included the 36 subjects (42±13 year-old) for whom the scapulo-humeral coordination was measured bilaterally, and "differential" prediction bands and intervals were computed, which describe right-to-left side differences. Bootstrap and Gaussian methods were compared using cross-validation analyses, by evaluating the coverage probability in comparison to a 90% target. Results showed a mean coverage for Bootstrap from 86% to 90%, compared to 67-70% for parametric bands and 87-88% for parametric intervals. Bootstrap prediction bands showed a distinctive change in amplitude and mean pattern related to age, with an increase toward scapula retraction, lateral rotation and posterior tilt. In conclusion, Bootstrap ensures an optimal coverage and should be preferred over parametric methods. Moreover, the stratification of "monolateral" prediction bands and intervals by age appears relevant for the correct classification of patients.
Turning is a requirement for most locomotor tasks; however, knowledge of the biomechanical requirements of successful turning is limited. Therefore, the aims of this study were to investigate the spatio-temporal and lower-limb kinematics of 90° turning. Seventeen typically developing children, fitted with full body and multi-segment foot marker sets, having performed both step (outside leg) and spin (inside leg) turning strategies at self-selected velocity, were included in the study. Three turning phases were identified: approach, turn, and depart. Stride velocity and stride length were reduced for both turning strategies for all turning phases (p<0.03 and p<0.01, respectively), while stance time and stride width were increased during only select phases (p<0.05 and p<0.01, respectively) for both turn conditions compared to straight gait. Many spatio-temporal differences between turn conditions and phases were also found (p<0.03). Lower-limb kinematics revealed numerous significant differences mainly in the coronal and transverse planes for the hip, knee, ankle, midfoot, and hallux between conditions (p<0.05). The findings summarized in this study help explain how typically developing children successfully execute turns and provide greater insight into the biomechanics of turning. This knowledge may be applied to a clinical setting to help improve the management of gait disorders in pathological populations, such as children with cerebral palsy.
Current approaches to detecting significantly activated regions of cerebral tissue use statistical parametric maps, which are thresholded to render the probability of one or more activated regions of one voxel, or larger, suitably small (e. g., 0.05). We present an approximate analysis giving the probability that one or more activated regions of a specified volume, or larger, could have occurred by chance. These results mean that detecting significant activations no longer depends on a fixed (and high) threshold, but can be effected at any (lower) threshold, in terms of the spatial extent of the activated region. The substantial improvement in sensitivity that ensues is illustrated using a power analysis and a simulated phantom activation study. © 1994 Wiley-Liss, Inc.
Functional data analysis involves the extension of familiar statistical procedures such as principal-components analysis, linear modelling and canonical correlation analysis to data where the raw observation is a function x, (t). An essential preliminary to a functional data analysis is often the registration or alignment of salient curve features by suitable monotone transformations hi(t). In effect, this conceptualizes variation among functions as being composed of two aspects: phase and amplitude. Registration aims to remove phase variation as a preliminary to statistical analyses of amplitude variation. A local nonlinear regression technique is described for identifying the smooth monotone transformations hi, and is illustrated by analyses of simulated and actual data.L'analyse de données se présentant sous la forme de fonctions x,(t) repose sur la généralisation d'outils statistiques familiers tels que l'analyse en composantes principales, les modèles linéaires et l'analyse des corrélations canoniques. L'étalonnage des caractéristiques saillantes des courbes à l'aide de transformations monotones hi(t) constitue souvent un préiequis essentiel au traitement statistique de telles données. II découle d'une décomposition en deux parties de la variation entre les fonctions observées: une phase et une amplitude. L'étalonnage vise à éliminer la première de ces deux sources de variation, ce qui permet de concentrer ensuite l'analyse sur la seconde. Les auteurs décrivent ici une technique de régression non linéaire locale facilitant l'identification de transformations monotones lisses hi appropriées. Leur propos est illustré à l'aide de données réelles et simulées.