Supervised methods with genomic data: a review and cautionary view

Ramón Díaz-Uriarte
Bioinformatics Unit
Spanish National Cancer Center (CNIO)
Melchor Fernández Almagro 3
Madrid, 28029


Keywords: differential expression, prediction, prognostic, microarrays, multiple testing, molecular signatures, software, statistics, machine learning, observation study

Contributed chapter to Data analysis and visualisation in genomics and proteomics, by F. Azuaje, and J. Dopazo, (eds.).


We review well accepted methods to address questions about differential expression of genes and class prediction from gene expression data. We highlight some new topics that deserve more attention: testing of differential expression of specific groups of genes, intra-group heterogeneity and class prediction, gene interaction in predictors, visualisation, difficulties in the biological interpretation of predictor genes and molecular signatures, and the use of ROC[Receiver Operating Characteristic curve]-based statistics for evaluating predictors and differential expression. We end with a review of some serious problems that can limit the potential of these methods; we focus specially on inadequate assessment of the performance of new methods (due to inadequate estimation of error rates and to the use of few and ``easy'' data sets) and failure to recognise observational studies and include needed covariates. A final comment is made about the need for freely available source code.

Chapter objectives

Reviews of the analysis of gene expression data (e.g. Slonim, 2002; Parmigiani et al., 2003; Speed, 2003; Simon et al., 2003; Tumor Analysis Best Practices Working Group, 2004; Draghici, 2002) often mention three objectives: a) class comparison, or finding/ranking of differentially expressed genes; b) class prediction or prognostic prediction; c) class discovery, also know as clustering or unsupervised analyses. We will not discuss class discovery or clustering here (it is discussed elsewhere on this book) and will concentrate on class comparison and class prediction. For the remaining two broad type of problems, this chapter has three main objectives: a) To bring a statistician, computer scientist, or computational biologist quickly up to speed by providing pointers to the literature on well accepted and standard methods 1. b) To emphasise some topics that deserve more attention and are open to additional theoretical, empirical, and computational contributions. c) To alert editors, reviewers, and general practitioners to several serious problems that can undermine the full potential of these techniques.

Class prediction and class comparison

Class comparison asks if different classes of subjects (e.g., lung cancer and prostate cancer patients) differ in their gene expression; the result is often a list of genes ranked by their degree of differential expression between classes; this objective can alternatively be to examine whether other non-categorical variables (such as expression of certain proteins or survival) are associated to gene expression. Class prediction or prognostic prediction tries to predict the class membership (or survival or protein expression or any prognostic variable) of a set of subjects given their gene expression data. Although related, these are different objectives that answer different biological questions and require different methods (unfortunately, this difference is not always recognised in empirical work). Ranking genes often precedes trying to use genes for class prediction (see also Sackett and Haynes, 2002), but genes that show large expression differences are not necessarily good predictors (e.g., p. 299 of Whitfield et al., 2003).

Class comparison: finding/ranking differentially expressed genes

The most common procedures analyse each and all of the genes of the array, ``asking the same question'' (e.g., ``is this gene differentially expressed between prostate and lung cancer patients?'') for each gene of the array. In contrast, when there are prespecified groups of genes, one can ask whether that subset of genes, as a whole, shows evidence of differential expression (e.g., ``are genes X, Y, Z, which are involved in cell cycle, differentially expressed between prostate and lung cancer patients?'').

Specially when asking the same question for each gene of the array, there are often two different objectives: to obtain a list of genes for which ``their differential expression is statistically significantly different'' and to rank genes based on some measure of how distant is the expression level between conditions (and this measure can be the p-value computed before) or how likely they are to differ. These objectives are related, but measuring the likelihood of differential expression requires additional assumptions, and obtaining p-values is more delicate than simply ranking. Even when p-values are obtained, however, they are used as informal rules of inference and to guide future experiments, rather than to provide ``black or white'' answers.

Asking the same question for each gene of the array

Widely accepted methods, with available software, involve the use of standard statistical tests (e.g., $ t-$test for two-class comparisons, ANOVA for multi-class comparisons, Cox models for survival data, etc), where analyses are carried out gene-by-gene (reviews in Dudoit et al., 2002 b; Cui and Churchill, 2003; Simon et al., 2003; Reiner et al., 2003, ch. 7). These analyses, although conducted gene-by-gene, need to take into account that thousands of null hypotheses are being tested (one for each gene): if we were to consider any of the genes with a ``rejected null'' as differentially expressed, we would end up with many false rejections. Appropriate correction for multiple testing is often conducted using either control of the Family Wise Error Rate or the False Discovery Rate. Controlling the Family Wise Error Rate refers to controlling the probability of making one or more false discoveries, or falsely rejecting the null, over the whole family of tests; this approached was detailed in Westfall and Young (1993) and its application to microarrays was pioneered by Dudoit et al. (2002 b). In contrast, the False Discovery Rate approach controls the expected proportion of erroneously rejected nulls among the rejected hypotheses; FDR controlled has been worked on mainly by Yoav Benjamini, Daniel Yekutieli, and their collaborators (see roee/index.htm) for lists of references and links; a recent review and applications to microarrays is Reiner et al. (2003); other approaches related to, or variations of, FDR are Storey (2002); Storey and Tibshirani (2003) and references therein; Ge et al. (2003) compare and discuss most of these different approaches. Detailed discussion of whether control of FWER or FDR is the most appropriate for a given situation is beyond the scope of this chapter; however, in many exploratory studies control of FDR is probably what most researchers need. In addition, methods for control of FDR do not require the subset pivotality assumption (Westfall and Young, 1993) to hold, and therefore are applicable to a wider range of tests; in addition, although control of FDR, as originally proposed by Benjamini and Hochberg (Benjamini and Hochberg, 1995), works only for independent (or positively regression dependent) tests statistics, the results in Reiner et al. (2003) show that violation of this assumption is generally inconsequential and there are also resampling-based FDR approaches that account for the dependence of the tests statistics.

Most gene-by-gene approaches, when computing the statistic for each gene, do not use the information contained in the rest of the genes, which could be wasteful; hierarchical Bayes or empirical Bayes methods allow to ``borrow information'' from all of the genes in the array when making inferences about each of the genes (see Smyth, 2004)2. Although not as well known as the above methods, Parmigiani and colleagues (Parmigiani et al., 2002; Garrett and Parmigiani, 2003) model gene expression using latent categories that are interpreted as a gene being over-expressed, under-expressed, or at baseline expression3; these models allows for denoising of the expression data, can enhance interpretability and help with visualisation, and ease comparisons among platforms. Finally, Bickel (2004) has argued for testing customised null hypothesis that redefine differential expression in a biologically meaningful way (e.g., any non-zero difference is not necessarily biologically relevant), and use ROC-based statistics4 (see below, section 5).

Asking questions about prespecified groups of genes

Among the tens of thousands of genes in an array, there might be prespecified sets of genes (e.g., those involved in cell cycle, or those found as relevant in a previous study) about which we might want to ask whether, as a whole, these subset of genes shows evidence of differential expression between groups of patients (or whether the expression of the whole set of genes is related to some other clinical variable, such as survival). Goeman et al. (2004) have proposed a method to test whether the expression pattern of a group of genes is related to some outcome of interest (be it class membership, survival, or a non-censored continuous variable). Their approach exploits the connection between differential expression among groups and predictability of clinical outcome, and the problem of number of genes being much larger than the number of samples is overcome using penalised regression models5. This method constitutes a very promising way of conducting tests of differential expression of subsets of genes6.

A different approach has been suggested by Mootha et al. (2003), who examine if the members of a set of genes are enriched (i.e., a proportion larger than expected) among the most differentially expressed genes between two classes. This method should be applicable to any other type of comparison, such as multiclass comparisons (via ANOVA) or survival data. The main differences between the approaches of Mootha et al. (2003) and Goeman et al. (2004) are listed in Table 1. Although with a different objective, a method similar to that of Mootha et al. (2003) was proposed in Díaz-Uriarte et al. (2003) (see also Al-Shahrour et al., 2004); as in Mootha et al. (2003), the approach in Díaz-Uriarte et al. (2003) only works if genes with similar ranking or order belong to the the same set but, in contrast to Mootha et al. (2003), the approach of Díaz-Uriarte et al. (2003) will detect sets of genes that are not extreme in their statistic of differential expression; however, it is a method targeted towards exploratory purposes rather than for statistical testing of prespecified hypotheses.

Table 1: Comparison of methods in Goeman et al., 2004 and Mootha et al., 2003 for testing hypotheses about pre-specified sets of genes.
  Goeman et al., 2004 Mootha et al., 2003
Testing If the set of genes that belongs to set $ S$ shows differential expression between classes $ A$ and $ B$. If the ``most differentially expressed'' genes are mainly of one of the sets.


Multivariate: all genes in the set fitted simultaneously using a generalised linear model1. Univariate (gene-by-gene).
Ease of application Requires development of math for different cases (already done for two-class, multiclass, and censored data). Only needs ordering of genes with criteria of our choice.
Assumes equal behaviour of genes in set No. Genes in the set(s) of interest must have a similar ranking of the statistic2.
Application to different sets Need to carry out different tests for each of different sets of genes. Can be applied at once over different sets, and a permutation test carried out to test the single null hypothesis that no gene set is associated with the class distinction.

$ $In general, for multivariate hypotheses (``are the genes of set $ S$ differentially expressed between groups $ A$ and $ B$?'') we should prefer procedures that are fully multivariate (Krzanowski, 1988, pp. 235 and ff.).
$ $Requiring the set of genes to have a similar ranking of the statistic does not by itself guarantee that the set of genes will be made of genes that are co-expressed.

Class prediction and prognostic prediction


As explained above, the goal here is to predict the clinically relevant characteristic of a subject (be it class membership, survival, prognosis, or any other variable of interest) given the genetic profile of this subject. This is also an area of extremely active research, where the disciplines of statistics and machine learning have contributed much; Table 2 shows widely accepted methods and references.

Available reviews (see Table 2) show that relatively simple and well known methods such as k-Nearest Neighbour (KNN) and Diagonal Linear Discriminant Analysis (DLDA), together with Support Vector Machines (SVM), perform very well in most classification tasks in microarray data. Because of their performance and free availability3 in quality implementations, DLDA, KNN, and SVM should probably be used routinely as benchmarks when proposing new methods.

Table 2: Well known and good-performing class prediction methods. Because classification has been much more studied than prediction of survival, the methods listed for survival data are not as well known.
Method References
Diagonal Linear Discriminant Analysis (DLDA) Dudoit et al. (2002 a), Simon et al. (2003), Romualdi et al. (2003), Huang & Pan (2003), Duda et al. (2001) and Hastie et al. (2001)1
K-Nearest Neighbour Dudoit et al. (2002 a), Simon et al. (2003), Romualdi et al. (2003), Duda et al. (2001) and Hastie et al. (2001)
Support Vector Machines (SVM) Guyon et al. (2002), Lee & Lee (2003), Simon et al. (2003), Romualdi et al. (2003), Duda et al. (2001) and Hastie et al. (2001)
Partial Least Squares Stone & Brooks (1990), Garthwaite (1994), Ghosh (2003), Gusnanto et al. (2003), Huang & Pan (2003), Nguyen & Rocke (2002)
Random forests Breiman (2001), Liaw & Wiener (2002), Bureau et al. (2003), Gunther et al. (2003)
Survival data  
Partial Least Squares Park et al. (2002)
Penalised Cox regression Pawitan et al. (2004)

$ $Dudoit et al. (2002 a), Simon et al. (2003), Romualdi et al. (2003) are general reviews that include reviews and results from different data sets. Huang & Pan (2003) show the relationships between several of these (and other) methods. Duda et al. (2001) and Hastie et al. (2001) are general overviews, with additional background material in statistics and machine learning.

Five specific issues

We will discuss five issues that probably deserve more attention. First, for the user it quickly becomes evident that many methods yield non-unique solutions (see also section 6.3) or, in other words, can return different solutions of very similar quality (e.g., prediction error rate), which itself leads to the question of how to choose among solutions. A direct way of approaching this problem is via model combination and model averaging. Model averaging is well known among Bayesians (e.g., Wasserman, 2000; Hoeting et al., 1999), and theory shows that a (weighted) average of predictions from several models should perform better (at least no worse) than predictions from any single model. Bayesian Model Averaging approach is not without problems, however, specially selection of priors and computation, and model definition. Model averaging is also available outside the Bayesian camp; stacking was initially proposed by Wolpert (1992) in the machine learning community, and later developed by Breiman (1996) and Ting and Witten (1999) (see also Ripley, 1996; Hastie et al., 2001, for short accounts). AIC-based model averaging has been developed by Buckland et al. (1997) and Burnham and Anderson (2002). Somorjai et al. (2002) show succesfull examples of stacking applied to MR and IR spectra2. Finally, random forests do a kind of model averaging by using an ensemble of trees.

Regardless of which model(s) are used, two general problems can affect all models/algorithms. First, most of the available methods assume additive effects of genes. Non-additive relationships or interactions, also called synergistic (or antagonistic) effects, are present when the outcome (e.g., being of class A) depends no just on the sum of the independent contributions of X and Y, but on their combined effects. Non-additive relationships are likely both between genes (e.g., the snail [NM_005985] gene) and between genes and other factors (section 6.4). Random forests (Breiman, 2001a; Liaw and Wiener, 2002) implicitly incorporate interactions as they are an ensemble of classification trees, but the actual interactions are not easy to see. Boulesteix and Tutz (2004); Boulesteix et al. (2003) have attempted to explicitly search for patterns of interactions and use them in predictive models. Second, the predictive capacity of many models can be hampered by unrecognised heterogeneity within classes that are regarded as homogeneous. Not much work has been done in this area. This problem, for instance, was recognised in the past (e.g. Rosenwald et al., 2003) and is dealt with by Munagala et al. (2004)3.

A final set of problems involves the biological interpretation of class prediction models (together with making sense of information for potentially tens of thousands of coefficients). Most methods for building predictors tend not to return models that allow for easy biological interpretation of why and how those predictors are used, and how the genes in the predictors affect and relate to the class prediction. These problems are detailed in Díaz-Uriarte (2004) and an example are methods that use dimension reduction via PCA or PLS, where all genes have loadings on all the components, making it virtually impossible to interpret the biological meaning, if any, of the components4.

Visualisation methods can help with biological interpretation in this task. For microarray data the biplot, as extended by Pittelkow and Wislon (2003)5, is particularly useful, specially use of the GE-biplot both before and after selecting genes according to different criteria of relevance.

In addition, ``molecular signatures'' or ``gene expression signatures'' are key features in many studies in cancer research (Rosenwald et al., 2002; Shaffer et al., 2001; Golub et al., 1999; Alizadeh et al., 2000; Pomeroy et al., 2002; Shipp et al., 2002) and seem to imply the idea of coordinate expression of subsets of genes, so that some of these sets of coordinate expression would be related to some criterion of interest (e.g., cancer type, or survival) (for an almost definition of a signature see p. 375 in Shaffer et al., 2001). Recently Stegmaier et al. (2004) provide a very interesting example of a high-throughput, generic, method for screening of compounds that induce differentiation of leukaemia cells, based on gene expression signature of five genes; so gene expression signatures work as a surrogate for a biological state. In spite of their apparent relevance, however, there seems to be no approach for identifying molecular signatures. Recently, we proposed a method that is explicitly designed to try to identify molecular signatures: it finds sets of genes that are tightly coexpressed and that can be used as successful predictors (Díaz-Uriarte, 2004). This method could also help uncover situations that are inconsistent with the assumptions underlying the existence of a few, easily interpretable, signature components of coexpressed genes. However, there are several unsolved issues. On the one hand, the implicit model underlying Díaz-Uriarte (2004) is one where most of the genes are not relevant for prediction, relevant genes are involved in one and only one ``signature component'' (i.e., non-overlapping signature components), and the signature components are common, and behave similarly, in different groups; there are, however, richer biological models for biological signatures. In addition, there are related issues regarding differences in patterns of gene coexpression within and among-groups and potential instability concerns (see also section 6.3) about some results (see sections 3.2 and 3.3 in Díaz-Uriarte, 2004). Some of these issues might be solved with extensions to the method, and some might require completely different approaches. For example, modifications of the Plaid model of Lazzeroni and Owen (Lazzeroni and Owen, 2002) (see also Turner et al., 2004), which might allow a more principled, model-based, approach to the problem, within a richer class of models; or an extension of the simultaneous clustering and classification approach in Jörnsten and Yu (2003), where we could add normal mixture models with restrictions on the covariance matrix for clustering; or an approach based on the latent class methods of Parmigiani and colleagues (Parmigiani et al., 2002; Garrett and Parmigiani, 2003), where signature components are based on under-, over- or baseline expression (instead of expression levels), and potentially non-overlapping sets of genes for different classes. Work along these lines is currently in progress in our group. In any case, regardless of the exact method used, it is also relevant that the search for molecular signatures highlights that finding a few sets of genes with biological interpretability can be worth even if it leads to small loses in predictive performance (see also Somorjai et al., 2003) because good classification performance, per se, does not shed any light into the underlying biological or clinical phenomena.

ROC curves for evaluating predictors and differential expression

Specially for the two-class setting, common measures of performance (e.g. Pepe, 2003; Hastie et al., 2001; Baker et al., 2002) are Sensitivity, or True Positive Rate, the probability of predicting a positive outcome when the true state is positive (i.e., $ \frac{TP}{TP + FN}$ in Table 3) and Specificity, the probability of predicting a negative outcome when the true state of a case is negative (i.e., $ \frac{TN}{TN
+ FP}$)6.

Table 3: Confusion matrix for a two-class classification problem, with indication of the usual labels for the four types of outcome.
True Diseased Healthy
Diseased True Positive (TP) False Negative (FN)
Healthy False Positive (FP) True Negative (TN)

Sensitivity and Specificity are often used to construct a Receiver Operating Characteristic (ROC) curve7. A ROC curve (see, e.g., figure 1) (e.g. van Belle, 2002; Pepe et al., 2001; Pepe, 2003, ch. 4) is a plot of Sensitivity in the ordinate against one minus Specificity or the False Positive Rate (i.e., = $ \frac{FP}{TN +
FP}$) in the abscissa. In other words, a plot of the probability of a hit against the probability of false alarm (Duda et al., 2001). This shows us how the sensitivity and the false positive rate change as we modify the threshold that classifies a subject as a member of one class or the other. In addition, we can use as a statistic the ``Area under the curve'' for a ROC curve, which is ``(...) an overall measure of classification accuracy over all possible decision thresholds'' (Pepe, 2003; Bickel, 2004).

Figure 1: Two ROC curves from real microarray data; on top of each we indicate the Area Under the ROC Curve.

ROC curves and ROC-based statistics are widely (and successfully) used to evaluate the diagnostic utility of medical tests (e.g., X-rays, ultrasounds, biochemical tests, etc, as reviewed in the excellent book by Pepe, 2003). It seems reasonable that similar approaches could be used with microarray data, specially since ROC-based statistics are very flexible devices that allow us, for example, to model covariate effects on the ROC curves, and to combine multiple test results (see Pepe, 2003, for review). As mentioned above (section 3), Bickel (2004) and Pepe et al. (2003) have argued for the use of ROC-based statistics to rank genes. These authors (see also Xu and Li, 2003) argue that ranking genes using ROC-based statistics is more meaningful than using t- and F-based statistics or p-values. Using the area under the ROC curve for two groups is a measure of differential expression that also provides information on the discriminatory capacities of genes: the empirical area under the ROC curve is equal to the probability that a randomly selected patient from one of the groups will have a larger expression value than a randomly selected patient from the other group (Pepe, 2003; Bickel, 2004), and this summary, from the clinical or biological perspective, is often much more meaningful than a t-statistic or a p-value. In addition, the area under the ROC curve is equivalent to the Wilcoxon rank sum statistic ($ \equiv$ Mann-Whitney U statistic), and thus it is a distribution-free rank statistic (Pepe, 2003; Pepe et al., 2003). Besides the area under the whole curve, Pepe et al. (2003) suggest using the empirical estimates of the ROC at a given False Positive Rate, $ t_0$, $ ROC(t_0)$, and the partial area under ROC at $ t_0$, $ pAUC(t_0)$, as measures of differential expression. These statistics do depend on $ t_0$, and a reasonable $ t_0$ could be the False Positive Rate that is acceptable in practice: when screening asymptomatic people, where prevalence of cancer is very low in average risk populations, it is important to keep the False Positive Rate extremely low because otherwise there would be large numbers of people undergoing expensive and invasive procedures (Pepe et al., 2003; Baker et al., 2002).

Caveats and admonitions

Estimating the error rate of the predictor

To evaluate the performance of a predictor, it is common to provide the error rate of the predictions. However, many papers, including ``high-profile'' ones, report error rates that are severely biased, leading to overoptimistic claims about the performance of different methods. This is a most unfortunate situation because lack of appropriate rigour in the application and adherence to appropriate rules of evidence undermines trust in the promises of these technologies. These severe problems were addressed in the bioinformatics literature in Ambroise and McLachlan (2002) and Simon et al. (2003). In spite of the seriousness of the problem, the practice of reporting severely biased error rates is still common, and this has prompted a recent review (Ransohoff, 2004) that tries, once again, to alert users, reviewers, and editors against computing, reporting, and accepting overly optimistic error rates. We will review here the two most common problems, remembering that our objective when providing an estimate of the error rate is to provide an estimate of the likely error rate we will make when we apply our classifier to new data sets from the same population.

On possible problem is reporting the ``resubstitution rate'', the error rate computed from the very same observations that were used to build the classifier, because the resubstitution error rate is severely biased-down due to overfitting: if we fit a classifier to a data set, we can expect it to ``adapt to'' some peculiarities of the data, which will make it work well with those data, but might lead it to work poorly with data not yet seen by the classifier or learner. This problem is even more serious with microarray data, where there are thousands of genes that can be part of a predictor. With so many variables, and so few samples, it is very easy to find a predictor that works perfectly in a completely random data set (see, for example, Fig. 8.4 in Simon et al., 2003). To solve this problem either cross-validation or bootstrap have been used; both methods build the predictor using a subset of the data, and then predict the values for the remaining data, thus insuring that the predictions are from data not used for the training.

A second common problem is to carry out the cross-validation after the gene selection: all samples are used for gene selection, and the cross-validation process does not include gene selection. This leads to very optimistic estimates of the error rate, as shown in Ambroise and McLachlan (2002) and Simon et al. (2003) because we incur in a problem similar to overfitting when the gene selection is carried out. The solution is to perform cross-validation or bootstrap so that all steps of the analysis (including gene selection, but also other potential steps such as imputation) are included in the cross-validation8. Whether cross-validation (and what size of folds) or bootstrap (and what type of bootstrap) should be used is beyond the scope of this review (see Efron and Gong, 1983; Efron and Tibshirani, 1997; Braga-Neto and Dougherty, 2004; Davison and Hinkley, 1997; Simon et al., 2003; Ambroise and McLachlan, 2002; Efron and Tibshirani, 1993).

Reinventions of the wheel and comparisons among methods

There are two related problems that slow the development of the field just simply by overwhelming researchers with new publications and algorithms. On the one hand, there is a fair amount of ``repeated reinventions of the wheel'', or ignorance of previously dealt with problems (many of them, with solutions by now). In addition, many new methods that are published are not evaluated against ``standard'' competing methods (see also section 4), or are evaluated using only data sets regarded as ``easy'' (e.g., the Leukaemia data set of Golub et al., 1999), making it hard to asses how new methods really perform (in sharp contrast, for example, Dettling and Bühlmann, 2004, use six different data sets and three competing predictors). Hopefully, more strict standards for evaluation of proposed methods (together with the requirements of a freely available ``reference implementation'' --section 7) will decrease the amount of new proposed methods, will shorten the ``to-read'' pile, and will allow researchers to carry out wider and more exhaustive searches for more mature solutions to similar problems from other fields.

Stability of results or which set of candidate genes is biologically relevant?

Suppose a predictor has been built that includes 20 genes. How far can we take biological interpretation on the relevance of these genes? A paper by Somorjai et al. (2003) suggests that often not very far; the problem is the instability or non-uniqueness of results, a phenomenon called the ``Rashomon effect'' by L. Breiman (Breiman, 2001b). It is very common that, if we re-run a given procedure with only minor changes or using bootstrap samples, we end up with very different sets of models, suggesting that there are many different ``optimal'' subsets of genes (because there are many different descriptions that give approximately the same minimum error rate Breiman, 2001b). Somorjai et al. (2003) show how this can arise because of small sample sizes and an extremely small sample per feature ratio (i.e., very small number of arrays relative to the number of genes). Somorjai et al. (2003) suggest using a variety of classifiers or predictors and finding whether the same features are selected; if the same set of genes is repeatedly selected, we would be more confident that the set is reasonably robust. Of course, this way of examining robustness to selection methods cannot be used if feature selection is carried out using the same filter method for different classifiers (e.g., finding the 200 genes with largest $ F$-ratio, and then using those 200 genes with DLDA, KNN, and SVM). Additionally, the bootstrap can be used to examine variation in solutions achieved. The multiplicity problem deserves much more careful attention and prompts for cautious interpretation of results.

Recognising observational studies and the need of including covariates

Although microarray studies are often referred to as ``experiments'' they are frequently observational studies. The differences between observational and experimental studies are well known in statistics and epidemiology, and affect both analyses and interpretation of results. Observational studies present several potential problems, specially:

These issues are well known in epidemiology, which studies patterns of disease and possible factors that affect these patterns of disease by using mainly observational data (Collett, 2003; Potter, 2003). However, as indicated by Potter (2003) concerns related to microarrays being often observational studies are mostly absent from standard papers and textbooks on microarray design and analysis (Yang and Speed, 2002; Simon and Dobbin, 2003; Speed, 2003; Simon et al., 2003; Churchill, 2002; Draghici, 2002; Tumor Analysis Best Practices Working Group, 2004). In particular, it is surprising that confounding and interaction have not been given more consideration (see also Ntzani and Ioannidis, 2003, who show that an alarmingly large number of predictive studies with DNA arrays do not include adjustments for other known, and potentially competing, predictors). Confounding and interaction can be addressed, at least partially, by appropriately using relevant covariates in the statistical models9.

How is this relevant for microarray data? As Potter (2003) illustrates, many of the differences seen in expression profiles between different types of cancers can be the result of confounding by age and sex. Another example is provided by Patel et al. (2004), who have reviewed evidence that clearly indicates that there are sex-specific differences in susceptibility to, and biology and progression of, lung cancer. Some of these sex-specific differences could be related to differential expression of certain genes, decreased DNA repair capacity in women, increased incidence of certain mutations, and estrogen signalling. All of these factors and differences make it extremely likely that both confounding and interaction will occur related to sex in studies of the relationship between gene expression and cancer10, and in the development of predictive models. However, the good news is that sex and age of patients are often known for each microarray sample; these two variables, thus, should routinely be included in the analysis as covariates and to examine possible interactions. (Interestingly, Patel et al., 2004, call for undertaking sex-specific research in lung cancer). Of course, comments regarding sex and age are extensive to other potential confounders (e.g., diet, exercise, region of origin, etc), for which information might be available. Controlling for the effect of confounders with strong effects (and, from the biology we know, sex and age are likely to be confounders with strong effects in many cases), can lead to increases in statistical power, because a source of variation is being taken into account rather than being thrown into the error term11. Thus, by controlling the effects of covariates we can be more likely to detect differential expression between conditions. On the other hand, if differences between groups are mainly due to confounders (e.g., because of a disproportionate presence of one sex in one of the groups), only after controlling for the confounder can we trust that differential expression of certain genes or the predictive ability of our model is not due to confounding. With respect to interactions (e.g., that the effects of changes in the expression of certain genes depend strongly on, say, sex), their presence can be an important finding in itself, as is the case of sex-differences and lung cancer biology (Patel et al., 2004). Finally, if there are interactions with, say, sex, we will obtain lower error rates if we develop different predictive models for men and women than if we use a model that makes predictions independently of sex.

Collaboration between statisticians and biologists and the use of software ``magic bullets''

Successful use of microarrays to answer biologically relevant questions will require close collaboration between biologists and statisticians during the complete process of the study. The need for statisticians' advice during the experimental design has been discussed before (Simon and Dobbin, 2003; Yang and Speed, 2002; Churchill, 2002) and is not the subject of this chapter; however, it should be remembered that full details of the experimental set up are necessary for the use of appropriate statistical methods. In the context of this chapter, statisticians need to realize that there are often many subtleties in the interpretation of microarray results that preclude simple mappings from RNA expression data to phenotypes (O'Neill et al., 2003). At the same time, statistical help is needed to insure that the statistical model and test being used is addressing the biological questions of interest. What in any case is unrealistic is to expect that if the biologist sends a file with 15000 rows by 200 columns (genes by subject) to the statistician, the statistician will return to the biologist the list of, say, 30 genes that are the answer to the biological question. But that is, in fact, what some users often expect from software tools or statistical consulting, and what some statisticians might believe is possible/desirable. And this also means that the questions asked are sometimes reformulated to accommodate the available software.

The problem of those expectations and procedures is that they lack key ingredients often needed to provide an answer to the underlying biological question. Table 4 lists some typical questions that a statistician might ask12. Only after these (and other) questions have been answered, it is time to search for the appropriate tool, which might be a web tool, a GUI-based stats program, or might require the competent use of command-driven programs or the development of new programs to carry out the customised required analyses.

Table 4: Some relevant questions statisticians and biologists should engage in a dialog about.
Are genes grouped in families, and are we interested in the overall responses of groups of genes, or should we look at individual genes?
Are certain genes or spots in the array more relevant biologically, maybe because they are easier to measure reliably with other assays?
Is there additional information on which genes are likely to be differentially expressed?
Do you really need the best possible predictor that statistical computing will get you, or do you want a small list of genes very likely to be differentially expressed?
In what stage of the scientific discovery process is this study, and how tight control do you require over the Type I error rate?
What other information and variables about the patients, besides the microarray data, do you have available?
What population do you expect the results of these studies to be relevant for?
Are these the original, complete data, and are these the original biological questions, or have the data and questions gone through an already long run of analyses which has already filtered data and reoriented hypotheses?
What is the next stage of this study, or what do you want to do with these results?
What additional studies could be done to confirm the results from these analyses?

Final note: source code should be available

Many new methods papers are published every month, and biologists and applied statisticians do not have the time to implement each and every idea that is published, nor to deal with the complications associated with patented algorithms. Sometimes, however, when researchers ask for software from authors of methods paper they face answers such as `` method is straightforward to implement from the explanations in my paper'', ``...the method will soon be available as part of program XYZ (which is proprietary)'', or `` ...I am not in the business of providing software to anyone''.

In the opening lecture of the Royal Statistical Society meeting of 2002, titled ``Statistical methods need software'', Brian Ripley (Ripley, 2002) proposed ``(...) a reference implementation, some code which is warranted to give the authors intended answers in a moderately-sized problem. It need not be efficient, but it should be available to anyone and everyone.'' Calls for availability of software, including source code, in bioinformatics research have also been made in other settings (e.g. Dudoit et al., 2003; Marshall, 2003), and the Open Bioinformatics Foundation ( is ``focused on supporting open source programming in bioinformatics.'' The Free Software Foundation ( and the Open Source Initiative ( explain free and open source software. The reasons for making source code available in bioinformatics and microarray research are summarised by Dudoit et al. (2003, p. 46) and are reproduced in Table 5.

In this review, and following the above spirit, we have been highly biased towards methods for which software, including source code, is available; besides the philosophical issues involved, this is also a pragmatic decision.

Table 5: Reasons why source code should be available in bioinformatics, from p. 46 of Dudoit et al. (2003).
...nt verification)


J. Goeman and V. Mootha for answers about the workings of their procedures. C. Lázaro-Perea provided detailed and careful comments on the ms. that forced me to rewrite it and greatly improved it. The editors and reviewers for very helpful comments on the ms. The author was partially supported by the Ramón y Cajal program of the Spanish MCyT (Ministry of Science and Technology); funding partially provided by project TIC2003-09331-C02-02 of the Spanish MCyT.


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About this document ...

Supervised methods with genomic data: a review and cautionary view

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... methods1
Lack of space precludes a full review; other lists of references can be found in weip/course/ge/syl1.html and gparmigi/688/readings.html, from two well-known statisticians
Another review of ``moderated'' or ``modified'' t and F statistics is Cui and Churchill (2003). The approach developed by Gordon Smyth (Smyth, 2004) is applicable to a wide range of linear models (in contrast to some earlier approaches, that were only suited for specific comparisons), and an R ( package, limma, is available from Bioconductor (, and also incorporates accounting for multiple testing. However, although applicable to linear models, borrowing strength from all other genes is not as yet implemented in an easy to use tool for problems such as censored data, often analysed with Cox models.
... expression3
They use a bayesian hierarchical mixture model --with uniform distributions for abnormally high and abnormally low expression and normal distribution for baseline expression--, and the model returns, for each gene and sample, the probability that it is over-, under-, or baseline-expressed. Software --R code-- is available from See also Newton et al. (2004) who use a semiparametric hierarchical mixture model for a somewhat similar problem.
... statistics4
R code is available from
... models5
Penalised regression models are related to shrinkage methods, such as ridge regression, and models with random effects, and will drive many coefficients towards zero; they allow the fitting of models even when the number of samples (i.e., arrays) is smaller than the number of variables (i.e., genes).
... genes6
Code is available as package ``globaltest'' from Bioconductor.
... availability3
For instance, in R, DLDA is available in package ``sma'', KNN in package ``class'' (part of the VR bundle), and SVM in package ``e1071'', the latter from the libsvm library of Chang and Lin (2003).
... spectra2
However, the author has attempted, without success, both stacking and AIC-based model combination of logistic and multiresponse linear regression with genomic data.
Unfortunately, their code depends on non-free software.
... components4
Naively interpreting components using loadings or eliminating genes with small loadings is often not justified and can lead to unexpectedly suboptimal solutions (Cadima and Jolliffe, 2001; Jolliffe, 2002)
R code is available from Y. Pittelkow on request (see
Lemon et al. (2003) have argued that the Positive-predictive value (PPV), ``(...) the likelihood that a positive test result indicates a true positive'' (i.e., $ \frac{TP}{TP + FP}$) can be more relevant than sensitivity and specificity; however, this needs to be done carefully. In fact, for cancer screening the Predictive Value Positve (PVP) (similar in spirit to the PPV) and the Predictive Value Negative (PVN) are probably more important than the sensitivity and specificity, but they must be computed taking into account the prevalence, and not just the entries from the Table 3, as explained in van Belle (2002); Pepe (2003); Baker et al. (2002). This caveat is particularly important for very low prevalence diseases.
... curve7
The package ROC in Bioconductor offers several utilities for building and using ROC curves.
... cross-validation8
Of course, all these comments apply to other approaches, such as stepwise, forward, and backward selection methods in linear or logistic regression; in addition, these selection methods are well known for their instability and their leading to biased p-values (e.g., section 4.3 in Harrell, 2001). Anyway, these variable selection methods ought to be subject, too, to cross-validation or bootstrap.
... models9
Harrell (2001, pp. 3 and 390) emphasises the importance of multivariable modelling in observational studies because they allow us to control (hold constant mathematically the effect of) variables that might differ between groups because the study is observational
... cancer10
Interactions are very likely given the complex mappings between transcript levels and protein levels (O'Neill et al., 2003)
... term11
This is the idea behind blocking in experimental design: controlling a know source of variation.
... ask12
van Belle (2002) provides a very accessible account for the reasons behind those, and many other, questions statisticians ask.