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Correction: Inferring exemplar discriminability in brain representations

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Hamed Nili at University of Cambridge
Hamed Nili
Alexander Walther
Alexander Walther
Alexander Walther
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Arjen Alink
Arjen Alink
Arjen Alink
  • This person is not on ResearchGate, or hasn't claimed this research yet.
Nikolaus Kriegeskorte at Columbia University
Nikolaus Kriegeskorte
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Abstract and Figures

[This corrects the article DOI: 10.1371/journal.pone.0232551.].
Definition of the exemplar discriminability index (EDI) (A) The exemplar discriminability index (EDI) is defined as the difference of the average between-exemplar and average within-exemplar distance. Split-data RDMs (sdRDMs) are shown for two scenarios. In the scenario on the left, the face exemplars are not discriminable and the EDI is insignificant (H0 simulation). In the scenario on the right, the EDI is large and exemplars are discriminable (H1 simulation). (B) Applying multi-dimensional scaling (MDS; Kruskal and Wish, 1978; Torgerson, 1958) allows simultaneous visualisation of the representational geometries from both data sets (H0 on the left, H1 on the right). Exemplars are colour-coded and there are two pattern estimates for each exemplar. MDS gives a low dimensional (2D here) arrangement, wherein the distances between projections approximate the original distances in the high-dimensional pattern space. Here, to obtain MDS plots, we consider a distance matrix composed of comparisons between any two patterns (aggregated from both datasets A and B). When the EDI is significantly positive (right), repetitions of the same exemplar yield patterns that are more similar to each other than presentations of two different exemplars.
Definition of the exemplar discriminability index (EDI) (A) The exemplar discriminability index (EDI) is defined as the difference of the average between-exemplar and average within-exemplar distance. Split-data RDMs (sdRDMs) are shown for two scenarios. In the scenario on the left, the face exemplars are not discriminable and the EDI is insignificant (H0 simulation). In the scenario on the right, the EDI is large and exemplars are discriminable (H1 simulation). (B) Applying multi-dimensional scaling (MDS; Kruskal and Wish, 1978; Torgerson, 1958) allows simultaneous visualisation of the representational geometries from both data sets (H0 on the left, H1 on the right). Exemplars are colour-coded and there are two pattern estimates for each exemplar. MDS gives a low dimensional (2D here) arrangement, wherein the distances between projections approximate the original distances in the high-dimensional pattern space. Here, to obtain MDS plots, we consider a distance matrix composed of comparisons between any two patterns (aggregated from both datasets A and B). When the EDI is significantly positive (right), repetitions of the same exemplar yield patterns that are more similar to each other than presentations of two different exemplars.
… 
The EDI is not strictly symmetrically distributed under H0 EDI is the difference between the average diagonal entries (blue) and the average off-diagonal entries (red). The distributions of these averages can be skewed or non-skewed and have the same or different shapes. Note that under H0, the two distributions have the same expected value (gray vertical line). When the two distributions are both symmetric (right panels) or when they are both of the same shape (lower panels), their difference will be symmetrically distributed about 0. To see this, consider the fact that for any two points (one from each distribution, e.g. blue X and red X), another equally probable two points exist (blue O and red O) such that their difference has the same absolute value and the opposite sign. However, the distribution can be asymmetric (and thus non-Gaussian) when the two distributions are both skewed and have different shapes (upper left panel).
The EDI is not strictly symmetrically distributed under H0 EDI is the difference between the average diagonal entries (blue) and the average off-diagonal entries (red). The distributions of these averages can be skewed or non-skewed and have the same or different shapes. Note that under H0, the two distributions have the same expected value (gray vertical line). When the two distributions are both symmetric (right panels) or when they are both of the same shape (lower panels), their difference will be symmetrically distributed about 0. To see this, consider the fact that for any two points (one from each distribution, e.g. blue X and red X), another equally probable two points exist (blue O and red O) such that their difference has the same absolute value and the opposite sign. However, the distribution can be asymmetric (and thus non-Gaussian) when the two distributions are both skewed and have different shapes (upper left panel).
… 
T test and Wilcoxon signed-rank tests examine different null hypotheses about the data distribution The graph depicts four different sets of simulated EDIs (A to D) sampled from Gaussian distributions with different means and variances. Gray points indicate the sample data. The superimposed black point depicts the mean of each sample. The sample variance around its mean is depicted by black lines emanating from the mean (i.e., the black dot). Each set consists of 12 values, representing 12 subjects, which are submitted to both a t test and a Wilcoxon signed-ranked test. Both tests are right-tailed (meaning the tail extending into the positive direction) and test the null hypothesis that the data come from a distribution with zero mean (meaning no exemplar information across participants). p-values were pronounced significant (Y) if they passed the conventional threshold of p<0.05. Otherwise, they were unsignificant (N). (A) The sample mean is positive and the sample variance is small, hence both tests pass the significance threshold. (B) The sample mean is positive. However, an extreme value in the sample inflates the variance. Therefore the t test does not return a significant p-value anymore. (C) The mean is positive, but the sample contains a considerable number of negative EDIs. Accounting for the weight of the ranks of the data points, the Wilcoxon-sigend rank test penalizes this presence of negative data points, therefore its p-value does not pass the significance threshold. (D) The sample mean is close to zero, hence neither test indicates significant exemplar information.
T test and Wilcoxon signed-rank tests examine different null hypotheses about the data distribution The graph depicts four different sets of simulated EDIs (A to D) sampled from Gaussian distributions with different means and variances. Gray points indicate the sample data. The superimposed black point depicts the mean of each sample. The sample variance around its mean is depicted by black lines emanating from the mean (i.e., the black dot). Each set consists of 12 values, representing 12 subjects, which are submitted to both a t test and a Wilcoxon signed-ranked test. Both tests are right-tailed (meaning the tail extending into the positive direction) and test the null hypothesis that the data come from a distribution with zero mean (meaning no exemplar information across participants). p-values were pronounced significant (Y) if they passed the conventional threshold of p<0.05. Otherwise, they were unsignificant (N). (A) The sample mean is positive and the sample variance is small, hence both tests pass the significance threshold. (B) The sample mean is positive. However, an extreme value in the sample inflates the variance. Therefore the t test does not return a significant p-value anymore. (C) The mean is positive, but the sample contains a considerable number of negative EDIs. Accounting for the weight of the ranks of the data points, the Wilcoxon-sigend rank test penalizes this presence of negative data points, therefore its p-value does not pass the significance threshold. (D) The sample mean is close to zero, hence neither test indicates significant exemplar information.
… 
H0 simulation settings We use simulations to test the distribution of EDIs under the null hypothesis, H0 (here illustrated for only two response channels). For both datasets 1 and 2, patterns of exemplars were simulated in a multi-dimensional response space. The number of dimensions of the response space is the same as the number of response channels (e.g. voxels in a region of interest in fMRI data). Response patterns were simulated by first moving along the all-ones vector to reach the average activation, whose strength was determined by the activation component. A Gaussian pattern with a pattern component variance was added to the activation component, resulting in the response pattern centroid. Individual exemplars of each data split were then generated by adding random Gaussian noise with a noise component variance to the centroid. At each iteration of the simulation, the same centroid underlay all exemplars in both data splits (as exemplars are indiscriminable under H0). Exemplar labels were randomly assigned to experimental conditions and their repetitions. For details of the simulations see section 2.4.1.
H0 simulation settings We use simulations to test the distribution of EDIs under the null hypothesis, H0 (here illustrated for only two response channels). For both datasets 1 and 2, patterns of exemplars were simulated in a multi-dimensional response space. The number of dimensions of the response space is the same as the number of response channels (e.g. voxels in a region of interest in fMRI data). Response patterns were simulated by first moving along the all-ones vector to reach the average activation, whose strength was determined by the activation component. A Gaussian pattern with a pattern component variance was added to the activation component, resulting in the response pattern centroid. Individual exemplars of each data split were then generated by adding random Gaussian noise with a noise component variance to the centroid. At each iteration of the simulation, the same centroid underlay all exemplars in both data splits (as exemplars are indiscriminable under H0). Exemplar labels were randomly assigned to experimental conditions and their repetitions. For details of the simulations see section 2.4.1.
… 
Different exemplar sets (A) and ROIs (B) considered for the analysis of real fMRI data In order to evaluate the different tests, we apply the tests to real fMRI data. Having data from different regions of interest (V1, V2, V3, LOC, FFA and PPA, displayed for an example subject on the right) and testing the discriminability for different exemplar sets (i.e. different sets of experimental conditions, displayed on the left) confronts the tests with a wide range of situations. To obtain an estimate of the type I error, different tests were applied to randomised fMRI data (simulating the null hypothesis with real data). Seven subsets of conditions are considered: 1) all 24 images 2) 12 images belonging to the animate category 3) 12 images belonging to the inanimate category 4) 6 face exemplars 5) 6 images of bodies 6) 6 man-made objects 7) 6 natural objects.
+5
Different exemplar sets (A) and ROIs (B) considered for the analysis of real fMRI data In order to evaluate the different tests, we apply the tests to real fMRI data. Having data from different regions of interest (V1, V2, V3, LOC, FFA and PPA, displayed for an example subject on the right) and testing the discriminability for different exemplar sets (i.e. different sets of experimental conditions, displayed on the left) confronts the tests with a wide range of situations. To obtain an estimate of the type I error, different tests were applied to randomised fMRI data (simulating the null hypothesis with real data). Seven subsets of conditions are considered: 1) all 24 images 2) 12 images belonging to the animate category 3) 12 images belonging to the inanimate category 4) 6 face exemplars 5) 6 images of bodies 6) 6 man-made objects 7) 6 natural objects.
… 
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CORRECTION
Correction: Inferring exemplar discriminability
in brain representations
Hamed Nili, Alexander Walther, Arjen Alink, Nikolaus Kriegeskorte
The captions for Figs 1–10 are missing. Please see all complete, correct figure captions here.
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OPEN ACCESS
Citation: Nili H, Walther A, Alink A, Kriegeskorte N
(2021) Correction: Inferring exemplar
discriminability in brain representations. PLoS ONE
16(4): e0250474. https://doi.org/10.1371/journal.
pone.0250474
Published: April 19, 2021
Copyright: ©2021 Nili et al. This is an open access
article distributed under the terms of the Creative
Commons Attribution License, which permits
unrestricted use, distribution, and reproduction in
any medium, provided the original author and
source are credited.
Fig 1. Definition of the exemplar discriminability index (EDI). (A) The exemplar discriminability index (EDI) is defined as the difference of
the average between-exemplar and average within-exemplar distance. Split-data RDMs (sdRDMs) are shown for two scenarios. In the scenario
on the left, the face exemplars are not discriminable and the EDI is insignificant (H
0
simulation). In the scenario on the right, the EDI is large
and exemplars are discriminable (H1 simulation). (B) Applying multi-dimensional scaling (MDS; Kruskal and Wish, 1978; Torgerson, 1958)
allows simultaneous visualisation of the representational geometries from both data sets (H
0
on the left, H1 on the right). Exemplars are colour-
coded and there are two pattern estimates for each exemplar. MDS gives a low dimensional (2D here) arrangement, wherein the distances
between projections approximate the original distances in the high-dimensional pattern space. Here, to obtain MDS plots, we consider a
distance matrix composed of comparisons between any two patterns (aggregated from both datasets A and B). When the EDI is significantly
positive (right), repetitions of the same exemplar yield patterns that are more similar to each other than presentations of two different
exemplars.
https://doi.org/10.1371/journal.pone.0250474.g001
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PLOS ONE | https://doi.org/10.1371/journal.pone.0250474 April 19, 2021 2 / 11
Fig 2. The EDI is not strictly symmetrically distributed under H
0
.EDI is the difference between the average
diagonal entries (blue) and the average off-diagonal entries (red). The distributions of these averages can be skewed or
non-skewed and have the same or different shapes. Note that under H
0
, the two distributions have the same expected
value (gray vertical line). When the two distributions are both symmetric (right panels) or when they are both of the
same shape (lower panels), their difference will be symmetrically distributed about 0. To see this, consider the fact that
for any two points (one from each distribution, e.g. blue X and red X), another equally probable two points exist (blue
O and red O) such that their difference has the same absolute value and the opposite sign. However, the distribution
can be asymmetric (and thus non-Gaussian) when the two distributions are both skewed and have different shapes
(upper left panel).
https://doi.org/10.1371/journal.pone.0250474.g002
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Fig 3. ttest and Wilcoxon signed-rank tests examine different null hypotheses about the data distribution. The
graph depicts four different sets of simulated EDIs (A to D) sampled from Gaussian distributions with different means
and variances. Gray points indicate the sample data. The superimposed black point depicts the mean of each sample.
The sample variance around its mean is depicted by black lines emanating from the mean (i.e., the black dot). Each set
consists of 12 values, representing 12 subjects, which are submitted to both a ttest and a Wilcoxon signed-ranked test.
Both tests are right-tailed (meaning the tail extending into the positive direction) and test the null hypothesis that the
data come from a distribution with zero mean (meaning no exemplar information across participants). p-values were
pronounced significant (Y) if they passed the conventional threshold of p<0.05. Otherwise, they were unsignificant
(N). (A) The sample mean is positive and the sample variance is small, hence both tests pass the significance threshold.
(B) The sample mean is positive. However, an extreme value in the sample inflates the variance. Therefore the ttest
does not return a significant p-value anymore. (C) The mean is positive, but the sample contains a considerable
number of negative EDIs. Accounting for the weight of the ranks of the datapoints, the Wilcoxon-sigend rank test
penalizes this presence of negative data points, therefore its p-value does not pass the significance threshold. (D) The
sample mean is close to zero, hence neither test indicates significant exemplar information.
https://doi.org/10.1371/journal.pone.0250474.g003
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Fig 4. H
0
simulation settings. We use simulations to test the distribution of EDIs under the null hypothesis, H
0
(here
illustrated for only two response channels). For both datasets 1 and 2, patterns of exemplars were simulated in a multi-
dimensional response space. The number of dimensions of the response space is the same as the number of response
channels (e.g. voxels in a region of interest in fMRI data). Response patternswere simulated by first moving along the
all-ones vector to reach the average activation, whose strength was determined by the activation component. A
Gaussian pattern with a pattern component variance was added to the activation component, resulting in the response
pattern centroid. Individual exemplars of each data split were then generated by adding random Gaussian noise with a
noise component variance to the centroid. At each iteration of the simulation, the same centroid underlay all exemplars
in both data splits (as exemplars are indiscriminable under H
0
). Exemplar labels were randomly assigned to
experimental conditions and their repetitions. For details of the simulations see section 2.4.1.
https://doi.org/10.1371/journal.pone.0250474.g004
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Fig 5. Different exemplar sets (A) and ROIs (B) considered for the analysis of real fMRI data. In order to evaluate the different tests, we apply the tests to real fMRI
data. Having data from different regions of interest (V1, V2, V3, LOC, FFA and PPA, displayed for an example subject on the right) and testing the discriminability for
different exemplar sets (i.e. different sets of experimental conditions, displayed on the left) confronts the tests with a wide range of situations. To obtain an estimate of
the type I error, different tests were applied to randomised fMRI data (simulating the null hypothesis with real data). Seven subsets of conditions are considered: 1) all
24 images 2) 12 images belonging to the animate category 3) 12 images belonging to the inanimate category 4) 6 face exemplars 5) 6 images of bodies 6) 6 man-made
objects 7) 6 natural objects.
https://doi.org/10.1371/journal.pone.0250474.g005
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Fig 6. The EDI can be non-Gaussian under the null hypothesis (H
0
). Top panel: The distribution of the diagonal (blue)
and off-diagonal (red) dissimilarities. Distributions are obtained from the dissimilarities aggregated across pairs and
simulated subjects. There are fewer diagonal entries but they are sampled from the same distribution as off-diagonal
dissimilarities under the null hypothesis. Therefore the two distributions differ only by a scaling factor. Middle panel: The
distribution of the diagonal and off-diagonal average dissimilarities. Data were simulated for each subject (1,000 in total)
and the distributions were obtained by pooling the averages across all subjects. Bottom panel: The EDI distribution in this
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case is significantly non-Gaussian. Furthermore, a one-sided ttest or Wilcoxon signed rank test both yield significant p-
values. This means that EDI is not strictly zero-mean and Gaussian under H
0
. This could potentially inflate both types of
error and weaken inference based on the ttest. For each of the 1,000 subjects, data were simulated for 64 exemplars and
five response channels.
https://doi.org/10.1371/journal.pone.0250474.g006
Fig 7. Marginal histograms of deviations from Gaussianity of the simulated EDInull distribution. In order to richly characterise the properties
of the EDI distribution under H
0
, we simulate data for a large number of subjects for a range of values of any parameter. At every point of the cross-
product parameter space, sdRDMs (based on Pearson correlation distance) were obtained from simulated patterns for 10,000 subjects (patterns were
simulated independently for each subject). Lilliefors tests (testing whether the distribution of EDIs is significantly non-Gaussian) were applied to the
EDI distributions. These bars correspond to results from the Lilliefors Gaussianity test. A significance threshold of 5% was applied to the tests. Each
column also corresponds to one dimension of the parameter space. The first two (number of exemplars and number of response channels) mainly
depend on the experiment and analysis and the other three characterise the multivariate response space. For example, the red bar on the leftmost
bar-graph would correspond to the proportion of all Lilliefors tests that were significant (p <0.05, Bonferroni correction) when the number of
conditions were fixed at 2, 4, 8, etc. and the other parameters had any possible value. Two other tests were performed in a similar way to test EDIs for
the simulated data under H
0
. One was a one-sided Wilcoxon signed rank test and the other a one-sided ttest. The signed rank test was applied to
assess whether the EDIs were zero centered and the ttest was applied to obtain an estimate of the false-positive rates. Interestingly, for both tests less
than 1% violations occurred (p <0.05) and none of those survived correction for multiple comparisons (FDR and Bonferroni, p <0.05).
https://doi.org/10.1371/journal.pone.0250474.g007
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Fig 8. False-positives rates for different tests of exemplar discriminability. We estimated the false positive rate
(type-I error rate) by applying tests to randomised fMRI data (simulating H
0
, see section 2.4.2 for details). The null
data were simulated for a large number of iterations. At each iteration an ROI was randomly selected and data for each
subject from that ROI were randomised by permuting the condition labels for a full set containing experimental
conditions from both data splits. Randomisation was independently carried out for the two data splits. Under the null
hypothesis, exchangeability implies that there should not be a pattern difference between replications of the same
exemplar and a different exemplar. At each iteration (1,000 in total), we simulated group level data under H
0
and
applied all tests to it. We then computed the average percentage of false positives (proportion of significant cases from
the total number of 1,000 simulated cases, i.e. height of the bars) and their standard deviation (error bars). All tests had
a false positive rate that was not significantly different from 5% after a threshold of 0.05 was applied to the tests
(p >0.5, two-sided Wilcoxon test). This means that all tests protect against false positives at an expected rate and are
therefore valid.
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Fig 9. EDI and linear decoding accuracy are sensitive to differences in variance between conditions, but crossvalidated distance estimators are not. Consider
two distributions of pattern estimates for two conditions (A, dark gray; B, light gray), which have identical means (central point), but different variances (small for
A, large for B). (A) EDI: The mean distance between conditions (red) is larger than the mean distance within conditions (blue), because the distances within
condition A are very small. The EDI, therefore, is sensitive to the difference between the conditions. (B) Linear decoder: A linear decoder may maximise its
accuracy by placing the decision boundary to one side of the condition-A distribution. This way the decoder gets 100% accuracy for condition A. For condition B,
accuracy is below chance (50%), but above 0. On average across conditions, the accuracy will be above chance (50%). Linear decoders, therefore, are sensitive to
differences in variance (like the EDI). (C) Crossvalidated distance estimator: A crossvalidated distance estimator (e.g. crossnobis/LDC or LD-t) is a mean of inner
products of pattern-difference vectors (arrows: condition-B pattern estimate minus condition-A pattern estimate). On each fold of crossvalidation, an inner
product is computed between a training-set vector (blue) and a test-set vector (red). When the distributions are point-symmetric about the mean (e.g. Gaussian),
each test-set vector (solid-line red arrow) has an equally probable opposite twin (dashed-line red arrow). The inner products of the training-set vector with the twin
test-set vectors (red) will have opposite sign and the same absolute value. These pairs of values will cancel in the expectation. Crossvalidated distance estimates are
therefore insensitive to variance differences (as long as the distributions of the pattern estimates are point-symmetric, e.g. non-isotropic Gaussian, for both
conditions).
https://doi.org/10.1371/journal.pone.0250474.g009
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Reference
1. Nili H, Walther A, Alink A, Kriegeskorte N (2020) Inferring exemplar discriminability in brain representa-
tions. PLoS ONE 15(6): e0232551. https://doi.org/10.1371/journal.pone.0232551 PMID: 32520962
Fig 10. Comparing the sensitivity of different tests. Each test was applied to fMRI data from six different ROIs. In
each ROI, discriminabilities of seven category subsets were assessed. A test was pronounced significantif p <0.001.
The height of the bars shows the percentage of significant cases out of the 42 tested scenarios (e.g. 50% corresponds to
21 significant tests). Tests on EDIs computed from multivariately noise-normalized response patterns were
significantly more powerful (the last five triple of bars). Inference was done using subject bootstrap. All pairwise
comparisons between the two groups of tests (i.e. the first three triple of bars and the next five triples of bars
corresponding to unnormalized and multivariate noise-normalized data) were significant after controlling the
expected false discovery rate at 0.01.
https://doi.org/10.1371/journal.pone.0250474.g010
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Inferring exemplar discriminability in brain representations
Article
Full-text available
Representational distinctions within categories are important in all perceptual modalities and also in cognitive and motor representations. Recent pattern-information studies of brain activity have used condition-rich designs to sample the stimulus space more densely. To test whether brain response patterns discriminate among a set of stimuli (e.g. exemplars within a category) with good sensitivity, we can pool statistical evidence over all pairwise comparisons. Here we describe a wide range of statistical tests of exemplar discriminability and assess the validity (specificity) and power (sensitivity) of each test. The tests include previously used and novel, parametric and nonparametric tests, which treat subject as a random or fixed effect, and are based on different dissimilarity measures, different test statistics, and different inference procedures. We use simulated and real data to determine which tests are valid and which are most sensitive. A popular test statistic reflecting exemplar information is the exemplar discriminability index (EDI), which is defined as the average of the pattern dissimilarity estimates between different exemplars minus the average of the pattern dissimilarity estimates between repetitions of identical exemplars. The popular across-subject t test of the EDI (typically using correlation distance as the pattern dissimilarity measure) requires the assumption that the EDI is 0-mean normal under H0. Although this assumption is not strictly true, our simulations suggest that the test controls the false-positives rate at the nominal level, and is thus valid, in practice. However, test statistics based on average Mahalanobis distances or average linear-discriminant t values (both accounting for the multivariate error covariance among responses) are substantially more powerful for both random- and fixed-effects inference. Unlike average cross-validated distances, the EDI is sensitive to differences between the distributions associated with different exemplars (e.g. greater variability for some exemplars than for others), which complicates its interpretation. We suggest preferred procedures for safely and sensitively detecting subtle pattern differences between exemplars.
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