- Methodology article
- Open Access
A Bayesian calibration model for combining different pre-processing methods in Affymetrix chips
- Marta Blangiardo^{1}Email author and
- Sylvia Richardson^{1}
https://doi.org/10.1186/1471-2105-9-512
© Blangiardo and Richardson; licensee BioMed Central Ltd. 2008
- Received: 12 May 2008
- Accepted: 01 December 2008
- Published: 01 December 2008
Abstract
Background
In gene expression studies a key role is played by the so called "pre-processing", a series of steps designed to extract the signal and account for the sources of variability due to the technology used rather than to biological differences between the RNA samples. At the moment there is no commonly agreed gold standard pre-processing method and each researcher has the responsibility to choose one method, incurring the risk of false positive and false negative features arising from the particular method chosen.
Results
We propose a Bayesian calibration model that makes use of the information provided by several pre-processing methods and we show that this model gives a better assessment of the 'true' unknown differential expression between two conditions. We demonstrate how to estimate the posterior distribution of the differential expression values of interest from the combined information.
Conclusion
On simulated data and on the spike-in Latin Square dataset from Affymetrix the Bayesian calibration model proves to have more power than each pre-processing method. Its biological interest is demonstrated through an experimental example on publicly available data.
Keywords
- Combine Model
- Combine Method
- Deviance Information Criterion
- Relative Bias
- Exponential Component
Introduction
In gene expression studies, one of the first steps of the statistical analysis is to quantify the signal and correct the systematic noise through pre-processing, a series of actions designed to extract the signal and the sources of variability due to the technology used rather than to biological differences between the RNA samples. Many studies in the literature present the importance of pre-processing and show how this can influence the results in terms of differential expression (see, for example, [1] and [2]).
However, an agreed gold standard method does not exist and as Allison and colleagues [3] discuss in a recent paper, researchers are torn between the different pre-processing methods and usually end up restricting their analysis to using only one (often the most commonly used or the most user-friendly). A simple alternative strategy is to perform the analysis using two different pre-processing methods and then compare the results in terms of differential expression, focussing attention on the genes in the intersection.
The former strategy is reductive while the latter relies on the arbitrary choice of two methods and on that of considering only their intersection. Neither of these approaches makes optimal use of all the information provided by the pre-processing methods. As an alternative Hein and colleagues [4] and Turro and colleagues [5] proposed a method that, starting at the probe level, integrates all the pre-processing steps and their associated uncertainty in one single framework, and they demonstrated an increase in power compared to the methods most used in the literature.
A different strategy is proposed in this paper, as we aim at providing a method of analysis of a gene-expression experiment that combines and synthesises the information from several pre-processing methods, to obtain a better calibrated estimate of differential expression. In the context of microarray, many researchers have proposed different strategies to pool together several independent studies. In particular, they have focused attention on the integration of each gene effect across studies [6] or on the evaluation of the consistency of differential expression across platforms [7]. Conlon et al. [8, 9] have proposed a different approach: they do not estimate a combined gene effect across studies, but rather evaluate its differential expression through a pooled binary indicator for independent studies performed on the same platform. Their model is formulated in a Bayesian perspective and the posterior probability of differential expression is the quantity of interest. Later, Scharpf et al. [10] have extended Conlon's work to include the comparison of several platforms and to focus attention also on genes discordant across the experiments through the estimate of the sign of differential expression for each experiment. These papers share a meta-analytical framework, devoting their interest to synthesise several independent studies. In a different perspective Yang et al. [11] proposed a model for combining several measures of differential expression on the same experiment into an index; then they ranked the genes accordingly, using a permutation based test to select the differentially expressed genes. Similarly, we work on a single experiment, but concentrate on combining several pre-processing techniques; thus we adopt a measurement error perspective assuming for each gene a latent (unmeasured) 'true' differential expression and for each pre-processing method: (i) a measured value that departs from it (bias) and (ii) a variance component.
Modelling measurement error is common practice in epidemiology, where errors in the recording of explanatory variables are a frequent problem that has to be taken into account during the statistical analysis (see for example [12–15]); the formulation in a Bayesian framework has been discussed in the early 1990s by Thomas et al. [16], Richardson and Gilks [17, 18] and Richardson [19] among others, placing particular emphasis on the way their approach propagates coherently all sources of uncertainty in the data onto the estimation of the parameters of interest.
We follow Richardson and Gilks and specify a Bayesian calibration model for assessing differential expression in Affymetrix microarray, which combines the information from several pre-processing techniques and it is implemented using the freely available software WinBUGS [20]. The performance of the combined model for assessing differential expression is compared with the performance of the equivalent Bayesian model on each single pre-processing method using simulated data and the Latin square data set provided by Affymetrix [21]. The combined method is shown to have better operating characteristics; its biological interest is discussed on an experiment publicly available to evaluate the effect of High Fat Diet versus Normal Fat Diet in mice adipose tissue.
Results
Bayesian calibration model
A combined model for different pre-processing methods is characterised by two measurement error components: (i) a measure of the bias from the 'true' differential expression, that can assume additive or multiplicative form and (ii) a measure of variability around the mean gene expression.
The parameter α_{ gj }represents the global gene expression that is specific to the gene g and the pre-processing method j, whereas δ_{ g }is the 'true' (unknown) differential expression that we would like to capture for gene g. The method specific coefficient φ_{ j }quantifies the multiplicative bias of the method with respect to the latent quantity δ_{ g }.
where ϕ_{ j }is replaced by ξ_{ j }.
We empirically investigated the performance of additive and multiplicative bias on two datasets we use in the rest of the paper (see Materials and Methods for the description of the datasets). A pairwise comparison of the differential expression for the J methods is presented in Figure 1 and 2 of Additional file 1 and shows that the data are not lying on the diagonal, but are characterised by a slope different from 1. This suggests the presence of a multiplicative bias, while in general there is no evidence of a shift in the differential expression that would suggest the presence of an additive bias. For this reason, in the rest of the paper we consider the multiplicative bias model (1) and call ϕ_{ j }the relative bias as it indicates the inflation or deflation factor of the 'true' differential expression characteristic of a particular pre-processing method. Other details about the investigation we carried out on the two datasets are presented in the Discussion. In (1), for each gene, pre-processing and condition, the variance ${\sigma}_{gjk}^{2}$ is the result of a gene and condition specific component ${\sigma}_{gk}^{2}$ and an exponential error term specific to the pre-processing method and to the condition:
To complete the model we specify weakly informative prior distributions for all the remaining parameters as we do not have specific prior information available on these, so they will be informed by the data. To be precise, we specify a centred Normal distribution with large variance (10^{5}) on α_{ gj }and δ_{ g }. The relative bias coefficients are modelled as ϕ_{ j }~logN (0, 10^{5}) independently for j = 1, ..., J, imposing the identifiability constraint that ∏ ϕ_{ j }= 1.
The coefficients for the exponential component in (3) are assumed independent and to follow Normal distributions λ_{1jk}~N (0, 10^{5}), λ_{2jk}~N (0, 10^{5}), λ_{3jk}~N (0, 10^{5}), so that the function can model a wide variety of trends. As identifiability constraint, we impose that the sum of the experimental parameters in (3) over the three pre-processing methods is equal to 0 (Σ_{ j }λ_{1jk}= 0, Σ_{ j }λ_{2jk}= 0, Σ_{ j }λ_{3jk}= 0). We use the exponential parametrization to ensure positivity of (3). Finally, a_{ k }~Ga(0.01, 0.01) and we model $1/\sqrt{{b}_{k}}~U[0,mi{n}_{g}{({s}_{gk}^{2})}^{-0.5}]$, where ${s}_{gk}^{2}$ is the sample variance for gene g and condition k; this choice ensures that the posterior distributions of a_{ k }and b_{ k }are proper and well adapted to the scale of the data, as justified in [24].
Models (1) and (3) allow the borrowing of information across genes to estimate ϕ_{ j }and λ_{ jk }, and across methods to estimate δ_{ g }and ${\sigma}_{gk}^{2}$. The hierarchical model specified by (1), (3), (4) and the prior distributions are estimated using an MCMC algorithm coded in WinBUGS [20] to simulate the prior/posterior distribution of all unknown parameters. More details can be found in the Materials and Methods section and in Additional file 1.
Tail posterior probability
Here R is the number of replicates for each condition and J is the number of pre-processing methods considered. A value of z_{ g }is obtained at each iteration of the MCMC simulation after convergence is reached and the tail posterior probability statistic is then defined as follows:p(z_{ g }; z_{ α }) = P(|z_{ g }| > z_{ α }| y_{g})
where y_{g} denotes all the data available for gene g and z_{ α }is the α quantile of the standard normal distribution (usually α = 0.05 and consequently z_{ α }= 1.96). As discussed in [24] the histogram of p(z_{ g }; 1.96) is characterised by a local peak on the right tail in the presence of differentially expressed genes; this peak can be used to define a reasonable cut off for the differential expression (see the section that describes the results on the experimental data for an illustrative example).
The tail posterior probability statistic can be loosely interpreted as a Bayesian analogy of the t-test. It makes full use of the Bayesian output being a function of the differential expression (δ_{ g }) and of the variability $({\sigma}_{gj1}^{2}\text{and}{\sigma}_{gj2}^{2})$, is easy to use and was shown to have good statistical properties (see [24] for more details).
Assessing the fit of the model
Before using the calibration model as specified in (1)–(4) to infer differentially expressed genes, it is important to assess its ability to capture the sources of variability of the pre-processing methods, and in particular whether the parametrisation used in (3) is appropriate. One of the added benefits of working in a Bayesian framework is the ability to perform model checks by means of the predictive distribution of the parameters of interest. We use Mixed Posterior Predictive checks [25, 26], applied on gene expression data by Lewin et al. [22, 27] and focus attention on checking the gene specific variance, characterised by a hierarchical structure as described in equations (3) and (4). For each method, we compare the observed sample variance, calculated for the expression values, and the variance of the predicted expression values of each gene under the model using an empirical p-value.
Under the null hypothesis of the model being true, the distribution of the p-values should be approximatively uniform, while a poor model fit is indicated by departure from uniformity in the plot, suggesting a systematic difference between the observed values and the predicted ones (see Additional file 1 and [22, 27] for more details).
The Mixed Posterior Predictive check is characterised by a certain degree of subjectivity as the researcher has to visually assess the fit of the model through a plot. Complementary to this, we also provide a quantitative measure of model fit, enabling a more direct comparison between different models by means of the Deviance Information Criterion (DIC) [28]. This index has been proposed as an extension of the Akaike Information Criterion when dealing with Bayesian hierarchical models. It is defined as a function of the deviance of the model and of the effective number of parameters included:DIC = E_{ θ }[D(θ)] + p_{ D }
where E_{ θ }[D(θ)] is the posterior mean of the deviance of the model and p_{ D }is the effective number of parameters. When comparing two or more models, the one characterised by the smallest DIC shows the best fit to the data in hand.
Performance on simulated data
To first evaluate the benefits of using a model that combines several pre-processing methods we simulated log expression values for 1000 genes, 5 pre-processing and 2 conditions, following the approach described in the section Materials and Methods. We set 200 genes as differentially expressed and the remaining 800 genes as not differentially expressed. We considered a simplification of equation (3) assuming (i) the same variance for the two conditions $({\sigma}_{gj1}^{2}\equiv {\sigma}_{gj2}^{2})$ and (ii) the exponential component as independent from the global gene expression (λ_{2jk}= 0, λ_{3jk}= 0). This does not detract from evaluating the comparative performance of the calibration model versus each method. To evaluate the consistency of our results we repeated the simulation 10 times and averaged the results.
Ranking the pre-processing methods based on the ROC curve, Method 3, characterised by large relative bias and small variability (in the simulation set up ϕ_{3} = 2 and exp(λ_{3}) = 0.5), shows the best performance, while on the other end, Method 4, characterised by small relative bias and high variability (in the simulation set up ϕ_{4} = 0.5 and exp(λ_{4}) = 2), shows the worst ROC curve. Note that the 'reference' method, characterised by exp(λ_{1}) = 1 and ϕ_{1} = 1 shows an average specificity and sensitivity.
The influence of the relative bias coefficient ϕ_{ j }can be evaluated when doing pairwise comparisons of methods with the same variability (Method 2 vs Method 3 and Method 4 vs Method 5). If it is greater than 1 (Method 3 and Method 5), the corresponding pre-processing methods are assumed to magnify the 'true' differential expression. This results in a stronger signal and consequently in a greater ability to discern true positives and true negatives.
Operating characteristics for simulated data
DE | Non DE | FP (%) | TN (%) | TP (%) | FN (%) | sd | |
---|---|---|---|---|---|---|---|
Combined | 200 | 800 | 21 (2.6) | 779 (97.4) | 179 (89.5) | 21 (10.5) | 5.0 |
Method 1 (exp(λ) = 1, ϕ = 1) | 200 | 800 | 61 (7.6) | 739 (92.4) | 139 (69.5) | 61 (30.5) | 5.5 |
Method 2 (exp(λ) = 0.5, ϕ = 0.5) | 200 | 800 | 80 (10.0) | 720 (90.0) | 120 (60.0) | 80 (40.0) | 4.8 |
Method 3 (exp(λ) = 0.5, ϕ = 2) | 200 | 800 | 26 (3.2) | 774 (96.8) | 174 (87.0) | 26 (13.0) | 4.1 |
Method 4 (exp(λ) = 2, ϕ = 0.5) | 200 | 800 | 122 (15.2) | 678 (84.8) | 78 (49.0) | 122 (61.0) | 3.3 |
Method 5 (exp(λ) = 2, ϕ = 2) | 200 | 800 | 45 (5.6) | 755 (94.4) | 155 (77.5) | 45 (22.5) | 5.8 |
Latin square data set
Operating characteristics for Latin Square dataset
First 64 ranked probesets | ||||
---|---|---|---|---|
FP (%) | TN (%) | TP (%) | FN (%) | |
Combined | 12 (0.1) | 10545 (99.9) | 52 (81.3) | 12 (18.7) |
MAS5 | 23 (0.2) | 10534 (99.8) | 41 (64.1) | 23 (35.9) |
RMA | 14 (0.1) | 10543 (99.9) | 50 (78.1) | 14 (21.9) |
dChip | 31 (0.4) | 10526 (99.6) | 33 (51.6) | 31 (48.4) |
Posterior mean and credibility interval for ϕ_{ j }
Latin Square data set | Real Experiment: HFD vs NFD | |||
---|---|---|---|---|
E(ϕ_{ j }| y) | CI 95% | E(ϕ_{ j }| y) | CI 95% | |
MAS5 | 1.382 | [1.345–1.417] | 1.449 | [1.438–1.462] |
RMA | 1.144 | [1.124–1.165] | 1.064 | [1.056–1.074] |
dChip | 0.632 | [0.623–0.644] | 0.648 | [0.643–0.653] |
Similar results in terms of parameter estimates and performance are obtained for additional pairwise comparisons on this dataset, as presented in Tables 1, 2, 3 of Additional file 1. These comparisons have been selected to represent a wide spectrum of differential expression and to show that the results are consistent across different experiments.
Biological example: High Fat Diet versus Normal Fat Diet in mice adipose tissue
We use the histogram of the tail posterior probability to identify a reasonable cut off for calling a gene differentially expressed. Contrary to what happens for each single pre-processing method, the histogram of the tail posterior probability for the combined model shows a local peak on the right tail of the distribution (see Figure 3 of Additional file 1), indicating more evidence of differential expression. As suggested in [24], we select a high cut off value, in our case equal to 0.98, corresponding to the local peak on the combined distribution, and obtained a list of 292 'top' probesets classified as differentially expressed by the combined method. If we fixed the same cut off on the tail posterior probability for each pre-processing method, we would obtain substantially smaller lists with only 20 probesets classified as differentially expressed by MAS5, 32 by RMA and 41 by dChip. Selecting a standard threshold such as 0.95 gives similar results in term of the size of the lists: the combined one is the largest, while MAS5 is characterised by the smallest. This highlights the typical gain of confidence provided by the combined analysis: many more probesets have high posterior probability of being differentially expressed in the combined model than if we proceeded for each method separately. In order to perform a further comparison between the methods we also consider the first 292 probesets ranked according to the tail posterior probability for each single method. Note that in doing so we lower the cut off on the tail posterior probability scale for each method (0.78 for MAS5, 0.79 for RMA and 0.83 for dChip), introducing more noise in the list.
The contribution of MAS5 to the combined output has been further investigated by implementing a version of our model that combines only RMA and dChip and comparing the results. In general, combining only these two methods results in 205 probesets classified as differentially expressed for a cut off of 0.98, again chosen on the basis of the local peaks on the tail posterior probability histogram (see Figure 5 of Additional file 1). Out of these, 163 are in common with the model combining the 3 methods and 42 are specific to the two-methods calibration model. In the three-methods combined model, these 42 probesets are either (i) borderline or (ii) far from the top of the list, characterised by high tail posterior probability for only one method. This shows that for some probesets there remains a non negligible source of uncertainty when only two methods are considered. When a third method is included, evidence in favor of the null hypothesis of no differential expression is added and these probesets are not classified as significant anymore. Moreover, including an additional method in the synthesis results in 129 new probesets called differentially expressed: these probesets are borderline in the model with only two methods, characterised by a posterior probability ranging between 0.81 and 0.96; for them the evidence against H_{0} is thus strengthened by the introduction of MAS5 in the model.
Annotation for differentially expressed genes
First 292 genes with the largest posterior probability | ||||
---|---|---|---|---|
Combined Model | MAS5 | RMA | dChip | |
Biological Processes | 181 | 146 | 171 | 168 |
Molecular Functions | 193 | 154 | 180 | 173 |
Cellular Components | 179 | 145 | 173 | 169 |
KEGG pathways | 223 | 182 | 215 | 205 |
We compared the annotations of each method and of the combined one using the Fisher exact test [30] for the list of differentially expressed genes against the remaining genes in the array. We found that the combined method enriches 7 additional biological processes compared to those found by RMA, 5 additional biological processes and 3 additional cellular components compared to those found by MAS5 and 3 additional biological processes compared to those found by dChip.
The most represented KEGG pathways are related to immune response and oxidation (Antigen processing and presentation, MAPK signalling pathway, PPAR signaling pathway), biological regulators of physiological functions as energy metabolism, insulin action, immunity and inflammation and known from the literature to be associated with high fat diet (see [31] and [32]). Out of the KEGG pathways only the Insulin signaling pathway is enriched for the combined method with respect to the annotations of each single method.
Discussion
The Bayesian calibration model that we propose makes use of the dependence between the results of the pre-processing methods and in doing so it gives a better assessment of the 'true' unknown differential expression between the conditions. Note that our calibration model relies on having generally applicable pre-processing methods for single arrays, and it is not aimed at replacing any of these as its applicability is limited to differential expression assessment, but it demonstrates that when dealing with several pre-processing methods, there is a good alternative to choosing just one for assessing differential expression. By using natural assumptions on the structure of the measurement errors it enables (i) to borrow information across the genes to estimate the method specific operational characteristics (λ_{ j }, ϕ_{ j }) and (ii) to borrow information across the methods to estimate a measure of differential expression (δ_{ g }) and a component of variability $({\sigma}_{g}^{2})$ specific to each probeset.
The reader should bear in mind that for the sake of clarity, the simulation has been set to represent a wide range of pre-processing methods, thus large differences between the parameters λ_{ j }and ϕ_{ j }for the 5 methods included were set. Note that when the bias is larger than 1, the signal is inflated, hence conditions are more favorable to detect differential expression. This does not mean that, paradoxically, the best pre-processing method is always characterised by the largest bias as the variability plays also a key role. The method that shows the best performance in terms of sensitivity and specificity achieves a balance between degree of inflation and variability. A realistic scenario is presented for the Latin Square dataset in Table 2 and 3 where RMA shows the best performance and is characterised by a relative bias ϕ_{ j }= 1.14 larger than dChip, but smaller than MAS5.
The inflation/deflation of differential expression found by the different methods could be related to the different steps that build each pre-processing method. MAS5 is the only method that normalises each array separately, while the internal normalisation of RMA and dChip forces the distribution to be the same between arrays and between conditions. This tends to reduce the width of the distribution, and can impact on the differential expression, leading to some shrinkage. Moreover, it has been shown in the literature that methods that subtract the mismatch are subject to higher variability in the expression ([33] and [34]). This finds confirmation in our examples as MAS5 is the only method included in the model that considers the mismatch correction and it exhibits the largest variability within biological replicates.
We want to stress the importance of having a way to assess whether the formulation of the synthetic model is an adequate representation of what is common and specific to the different methods. Bayesian model checks based on prediction allow the comparison of the observed data and the 'predicted' data under the model with respect to any feature of interest. We believe that the Mixed Posterior Predictive checks that we considered are effective for model criticism. They have the advantage of calculating a measure of discrepancy (the empirical p-value) for each gene that can be easily displayed through a histogram. On the Latin square and the experimental data we chose a flexible and realistic model, that allows a relation between the expression $({\overline{y}}_{g})$ and the variance ${\sigma}_{gk}^{2}$. Comparing this formulation to a simpler version which assumes no relation between variability and level of expression by means of the Mixed Posterior Predictive checks, we clearly see a better fit for the more flexible model, which indicates that a variance parametrisation linked to the level of expression is better suited to the complexity of experimental data.
Performance of simulated data on ϕ and λ
E(λ_{ j }| y) [CI 95%] | E(ϕ_{ j }| y) [CI 95%] | sd (λ_{ j }) | sd (ϕ_{ j }) | |
---|---|---|---|---|
Method 1 (λ = 1, ϕ = 1) | 1.0 [1.0-1.0] | 1.0 [1.0-1.0] | n.a. | n.a. |
Method 2 (λ = 0.5, ϕ = 0.5) | 0.5 [0.49–0.51] | 0.5 [0.49–0.51] | 0.03 | 0.04 |
Method 3 (λ = 0.5, ϕ = 2) | 0.50 [0.49–0.51] | 2.0 [1.97–2.04] | 0.04 | 0.05 |
Method 4 (λ = 2, ϕ = 0.5) | 2.0 [1.96–2.07] | 0.5 [0.48–0.51] | 0.06 | 0.04 |
Method 5 (λ = 2, ϕ = 2) | 2.0 [1.97–2.04] | 2 [1.99–2.03] | 0.05 | 0.03 |
As already pointed out in the Results section, we have carried out an empirical investigation of the relative performance of the multiplicative bias model presented in (1) and of the additive bias model presented in (2). The pairwise comparison of the differential expression for the J methods considered supports the multiplicative bias formulation (Figure 1 and 2 of Additional file 1). In addition we reported the Deviance Information Criteria (DIC) for the Latin Square dataset in Table 4 and for the High Fat Diet dataset in Table 5 of Additional file 1. For both experiments, the DIC confirms the worst fit of the additive bias model.
We showed that combining several pre-processing methods is a way of including more information about the differential expression and that this improves the performance against each single method. We illustrated this using three commonly used pre-processing methods for Affymetrix chips, but we want to point out that our approach is generic in nature and that other pre-processing could be added or substituted instead of the three considered, as may be deemed appropriate by the analyst. We have carried out a limited investigation where we found that (i) including gcRMA [33] instead of RMA improves slightly the performance of the combining method, and (ii) including the vsn method [35] as expected alters the estimated parameters for the variability as vsn corrects for the dependence between the expression level and the variability, so the posterior estimates of λ_{2jk}and λ_{3jk}are very close to 0.
Further extensions
The model proposed is designed for studies where the primary interest is the estimate of differential expression. Even if this is not the focus of the present paper, we want to point out that our methodology could be adapted to deal with studies where the interest is the quantification of the signal. Depending on the pre-processing methods, calibration models for quantifying the gene expression level would include multiplicative or additive bias, which might be a certain functions of the expression level. Our approach is also applicable to experiments where more than two conditions are included (see Bochkina and Richardson [24] for the methodology related to Bayesian hierarchical models for multiclass studies). Other extensions could be of interest: instead of returning a measure of differential expression for each probeset, the model could be modified to obtain a measure for each gene, adding a new component for estimating the variability between probesets mapping on the same gene. Moreover, the approach could easily be adapted to deal with data coming from different types of chips (e.g. Agilent, Illumina).
Conclusion
This paper presents a Bayesian calibration model to synthesise information from several pre-processing methods. It is framed in a measurement error perspective, where each method is assumed to produce a biased measure of the latent true variable of interest. We specified the model for Affymetrix chips and focussed attention on differential expression studies, but we pointed out that the model is a methodological contribution and that our approach is applicable to a wide range of data types and experimental contrasts.
Methods
Simulated data
To test the performance of our method we simulated log expression values for 1000 genes, two conditions and five pre-processing methods from the model previously described, and extracted 5 replicates for each combination of condition and pre-processing (r = 1, ..., 5). We specified 200 differentially expressed genes characterised by a log expression ${y}_{gjkr}~N({\alpha}_{gj}+{(-1)}^{k}\frac{1}{2}{\delta}_{g}\times {\varphi}_{j},{\sigma}_{gj}^{2})$, with k = 1, 2, while for the remaining 800 genes the log expressions for both conditions came from the same distribution: ${y}_{gjkr}~N({\alpha}_{gj},{\sigma}_{gj}^{2})$. The parameters for simulating the distributions of α_{ gj }~N (6.79, 4.77) and δ_{ g }~N (0, 0.25) were obtained from experimental data we have analysed [36]; the model on the variance ${\sigma}_{gjk}^{2}$ is presented in equation (3) with (λ_{2jk}= 0, λ_{3jk}= 0) and we assume the same variance for the two conditions $({\sigma}_{gj1}^{2}\equiv {\sigma}_{gj2}^{2})$, obtained from experimental data, where the mean is 0.03 and the distribution can be summarised by the following quartiles: 0.02, 0.04, 0.09 and 0.15.
Values of the parameters for the 5 pre-processing methods in the simulation scenario
Method 1 | Method 2 | Method 3 | Method 4 | Method 5 | |
---|---|---|---|---|---|
exp(λ_{ j }) | 1 | 0.5 | 0.5 | 2 | 2 |
ϕ _{ j } | 1 | 0.5 | 2 | 0.5 | 2 |
We assume that the first method is the 'reference', being characterised by exp(λ_{1}) = 1 and ϕ_{1} = 1. Method 2 and Method 4 have a smaller relative bias (ϕ_{2} = ϕ_{4} = 0.5), while that of Method 3 and Method 5 is larger (ϕ_{3} = ϕ_{5} = 2); Method 2 and Method 3 have a variability smaller than 1 (exp(λ_{2}) = exp(λ_{3}) = 0.5) while that of Method 4 and Method 5 is larger (exp(λ_{4}) = exp(λ_{5}) = 2).
We used an MCMC algorithm with two chains to estimate the parameters of interest (we checked convergence for 10000 iterations and then extracted a sample of 1000 iterations so that the MC error was smaller that the 5% of the sample standard deviation, as recommended).
To evaluate the consistency of our results we repeated the simulation process 10 times and performed our Bayesian analysis for each run. The model estimates well the values of the parameters λ and ϕ (see Table 5 for their posterior mean and 95% credibility intervals; see Figures 7 and 8 of Additional file 1 for the associated posterior density plots).
As a point of comparison we also ran the model separately for each pre-processing method and compared the performance of both combined and single pre-processing methods in terms of sensitivity and specificity (see the results on simulated data).
Spike-in example: Latin Square Affymetrix data
We tested our method by means of the Affymetrix Latin Square data set [21]. The array used is human Hgu 133a and contains 22300 probesets. There are 14 experimental conditions and each has 3 replicates. We considered the experimental condition 2 versus 1 and the 42 spike-in probesets indicated by Affymetrix plus the 22 new spike-in probesets proposed by McGee et al. [37]. The ratio of the concentrations is 2:1 for 60 spike-in probesets and 0 for the remaining 4. Among the remaining 22236 probesets we extracted only the ones present in at least one condition, evaluated using the present/absent call included in the Affy R package http://www.r-project.org and obtained 10621 probesets.
Characteristics of MAS5, RMA and dChip
Background correction | Perfect Match correction | Normalisation | Summary | |
---|---|---|---|---|
MAS5 | Divide the chip in 16 regions. The lowest 2% is the background. Weighted average over all the regions. | Ideal Mismatch | Scaling | 1 step Tukey Biweight |
RMA | Global model for the distribution of the probe intensities | No correction | Quantile | Median Polish |
dChip | No correction | No correction | Invariant set using one array as default | Multi-chip linear model |
We ran the combined model, but also treated separately each pre-processing method. Again we performed the MCMC estimation with two chains (we checked convergence for 10000 iterations and then extracted a sample of 1000 iterations, characterised by a MC error smaller than 5% of the sample standard deviation, as recommended). The MCMC simulation of the combined model takes 9 hours and 52 minutes to run 11000 iterations on the 10621 probesets.
To see if the results are consistent we ran the model on additional pairwise comparisons, selected to represent a wide range of real differential expression (real fold change ranging from 0.002 to 128). The behavior is similar for the variability parameters λ across the experiments and the relative bias ϕ shows the same ranking for all the comparisons (MAS5>RMA>dChip). Also the performance of the methods is consistent in all the comparisons with the combined method showing the highest specificity and sensitivity. The results are presented in Tables 1, 2, 3 of Additional file 1.
Biological example: High Fat Diet versus Normal Fat Diet in mice adipose tissue
There are many studies in the literature describing the effect of high fat diet on gene expression of several tissues in mice (see for example [42] and [43] for adipose tissue and [44] for liver). They are particularly interesting since the effect of diet can trigger obesity, hypertension and be related to major pathologies as diabetes. In order to assess if our model leads to a more powerful analysis that improves the biological interpretability of the results we ran it on a publicly available experiment to study the effect of high fat diet (HFD) versus normal fat diet (NFD) on mice adipose tissue. The array used is mouse MG _U 74Av 2 and contains 12488 probesets. The experiment analyzes the strain 129 and for each condition there are 4 replicates. The .CEL files and the description of the experiments are available at the DGAP project website [29].
Again we pre-processed the data using MAS5, RMA and dChip and ran the combined model, but also treated separately each pre-processing method. The MCMC estimation was performed with two chains (we checked convergence for 10000 iterations and then extracted a sample of 1000 iterations. The MC error is smaller than the 5% of the sample standard deviation as recommended). The MCMC simulation for the combined model takes 12 hours and 10 minutes to run 11000 iterations on the 12488 probesets.
Implementation
The standard model built by equations (1), (3), (4) and by the prior distributions has been implemented in the free software WinBUGS and the code is provided in Additional file 1. All the analyses were performed on a DELL Precision workstation with 3.20 GHz and 2 GB of RAM. Note that it is relatively quick to run with a small number of genes (it takes around 5 minutes to perform 1000 iterations for 1000 genes, 2 conditions, 3 pre-processing and 5 replicates), but the time increases linearly with the number of genes. To increase the speed on the Latin Square data set (that includes 22300 probesets), we filtered the present probesets for at least one condition, using the present/absent call implemented in the Affy R package; this halved the computational time (from 20 to around 10 hours for 11000 iterations). If needed it is possible to apply different or more stringent criteria (e.g. selecting the most variable genes between the two conditions), as long as the non differentially expressed genes are well represented in the subset, otherwise the null distribution of δ_{ g }, and in particular its variability, is not properly identified and it can affect the correct estimate of the tail posterior probability.
Declarations
Acknowledgements
The authors would like to thank Alex Lewin and Natalia Bochkina for valuable statistical discussion, Xinzhong Li, Peter Thomason and James Scott for useful discussions that motivated the development of this work. We also would like to thank the three anonym referees that helped us strengthen our work. Marta Blangiardo and Sylvia Richardson's work was supported by BBSRC 'Exploiting Genomics' grant 28EGM16093.
Authors’ Affiliations
References
- Bolstad B, Irizarry R, Astrand M, Speed T: A comparison of normalization methods for high density oligonucleotide array data based on variance and bias. Bioinformatics 2003, 19(2):185–193. 10.1093/bioinformatics/19.2.185View ArticlePubMedGoogle Scholar
- Cope L, Irizarry R, Jaffee H, Wu Z, Speed T: A benchmark for Affymetrix genechip expression measures. Bioinformatics 2004, 20(3):55–65. 10.1093/bioinformatics/btg410View ArticleGoogle Scholar
- Allison D, Cui X, Page G, Sabripour M: Microarray data analysis: from disarray to consolidation and consensus. Nature Reviews 2006, 7: 55–65. 10.1038/nrg1749View ArticlePubMedGoogle Scholar
- Hein A, Richardson S, Causton H, Ambler G, Green P: BGX: a fully Bayesian integrated approach to the analysis of Affymetrix GeneChip data. Biostatistics 2005., 6(3):Google Scholar
- Turro E, Bochkina N, Hein A, Richardson S: BGX: a Bioconductor package for the Bayesian integrated analysis of Affymetrix GeneChips. BMC Bioinformatics 2007, 8: 439. 10.1186/1471-2105-8-439PubMed CentralView ArticlePubMedGoogle Scholar
- Choi J, Yu U, Kim S, Yoo O: Combining multiple microarray studies and modelling inter-study variation. Bioinformatics 2003, i84-i90. 10.1093/bioinformatics/btg1010Google Scholar
- Parmigiani G, Garrett-Mayer E, Anbazhagan R, Gabrielson E: A cross-study comparison of gene expression studies for the molecular classification of lung cancer. Clinical Cancer Research 2004, 5(81):2922–2927. 10.1158/1078-0432.CCR-03-0490View ArticleGoogle Scholar
- Conlon E, Song J, Liu J: Bayesian models for pooling microarray studies with multiple sources of replications. BMC Bioinformatics 2006, 7: 247. 10.1186/1471-2105-7-247PubMed CentralView ArticlePubMedGoogle Scholar
- Conlon E, Song J, Liu J: Bayesian meta-analysis models for microarray data: a comparative study. BMC Bioinformatics 2007, 8: 80. 10.1186/1471-2105-8-80PubMed CentralView ArticlePubMedGoogle Scholar
- Scharpf R, Tjelmeland H, Parmigiani G, Nobel A: A Bayesian model for cross-study differential gene expression. Johns Hopkins University, Dept of Biostatistics Working Papers 2008, 158: 1–42. [http://www.bepress.com/jhubiostat/paper158]Google Scholar
- Yang Y, Xiao Y, Segal M: Identifying differentially expressed genes from microarray experiments via statistic synthesis. Bioinformatics 2005, 1084–1093.Google Scholar
- Carroll R: Covariance analysis in generalized linear measurement error models. Statistics in Medicine 1989, 9: 1075–1093. 10.1002/sim.4780080907View ArticleGoogle Scholar
- Armstrong B: The effects of measurement error on relative risks regression. American Journal of Epidemiology 1991, 132: 1176–1184.Google Scholar
- Carroll R, Ruppert D, Stefanski L: Measurement error in nonlinear models. Chapman & Hall, London; 1995.View ArticleGoogle Scholar
- Gustafson P: Measurement error and misclassification in statistics and epidemiology. Chapman & Hall/CRC, New York; 2004.Google Scholar
- Thomas D, Gauderman W, Kerber R: A non-parametric Monte Carlo approach to adjustment for covariate measurement errors in regression problems. Technical report, Department of Preventive Medicine, University of Southern California 1991.Google Scholar
- Richardson S, Gilks W: A Bayesian approach to measurement error problems in epidemiology using conditional independence models. American Journal of Epidemiology 1993, 138: 430–442.PubMedGoogle Scholar
- Richardson S, Gilks W: Conditional independence models for epidemiological studies with covariate measurement error. Statistics in Medicine 1993, 12: 1703–1722. 10.1002/sim.4780121806View ArticlePubMedGoogle Scholar
- Richardson S: Markov Chain Monte Carlo in practice. Chapman & Hall, London chap. Measurement error; 1996:401–418.Google Scholar
- Spiegelhalter D, Thomas A, Best N, Lunn D: WinBUGS User Manual. Version 1.4.3, MRC Biostatistics Unit, Cambridge; 2007.Google Scholar
- Affymetrix: Latin Square Data Set.[http://www.affymetrix.com/support/technical/sample_data/datasets.affx]
- Lewin A, Richardson S, Marshall C, Glazier A, Aitman T: Bayesian Modelling of Differential Gene Expression. Biometrics 2006, 62: 1–9. 10.1111/j.1541-0420.2005.00394.xView ArticlePubMedGoogle Scholar
- Baldi P, Long A: A Bayesian framework for the analysis of microarray expression data: regularized t-test and statistical inferences of gene changes. Bioinformatics 2001, 17: 509–519. 10.1093/bioinformatics/17.6.509View ArticlePubMedGoogle Scholar
- Bochkina N, Richardson S: Tail Posterior Probability for Inference in Pairwise and Multiclass Gene Expression Data. Biometrics 2007, 63(4):1117–25.View ArticlePubMedGoogle Scholar
- Gelman A, Meng X, Stern H: Posterior predictive assessment of model fitness via realized discrepancies. Statistica Sinica 1996, 6: 733–807.Google Scholar
- Marshall C, Spiegelhalter D: Identifying outliers in Bayesian hierarchical models: a simulation-based approach. Bayesian Analysis 2007, 2(2):409–444.View ArticleGoogle Scholar
- Lewin A, Bochkina N, Richardson S: Fully Bayesian mixture model for differential gene expression: simulations and model checks. Stat Appl Genet Mol Biol 2007, 6: Article36.PubMedGoogle Scholar
- Spiegelhalter D, Best N, Carlin B, Linde A: Bayesian measures of model complexity and fit (with discussion). Journal of the Royal Statistical Society, Series B 2002, 64(4):583–639. 10.1111/1467-9868.00353View ArticleGoogle Scholar
- Diabetes Genome Anatomy Project website[http://www.diabetesgenome.org/arraydata.cgi]
- Al-Shahrour F, Minguez P, T'arraga J, Montaner D, Alloza E, Vaquerizas J, Conde L, Blaschke C, Vera J, Dopazo J: BABELOMICS: a systems biology perspective in the functional annotation of genome-scale experiments. Nucleic Acids Research 2006, 34: W472-W476. 10.1093/nar/gkl172PubMed CentralView ArticlePubMedGoogle Scholar
- Verwaerde C, Delanoye A, Macia L, Tailleux A, Wolowczuk I: Influence of high-fat feeding on both naive and antigen-experienced T-cell immune response in DO10.11 mice. Scandinavian Journal of Immunology 2006, 64(5):457–466. 10.1111/j.1365-3083.2006.01791.xView ArticlePubMedGoogle Scholar
- Masternak M, Bartke A: PPARs in Calorie Restricted and Genetically Long-Lived Mice. PPAR Research 2007, ID28436: 1–7. 10.1155/2007/28436View ArticleGoogle Scholar
- Wu Z, Irizarry R, Gentleman R, Martinez-Murillo F, Spencer F: A Model-Based Background Adjustment for Oligonucleotide Expression Arrays. Journal of the American Statistical Society 2004, 99(468):909–917.View ArticleGoogle Scholar
- Qin L, Beyer R, Hudson F, Linford N, Morris D, Kerr K: Evaluation of methods for oligonucleotide array data via quantitative real-time PCR. BMC Bioinformatics 2006, 7: 23. 10.1186/1471-2105-7-23PubMed CentralView ArticlePubMedGoogle Scholar
- Huber W, von Heydebreck A, Sultmann H, Poustka A, Vingron M: Variance stabilization applied to microarray data calibration and to quantification of differential expression. Bioinformatics 2002, 18 Suppl 1: S96-S104.View ArticlePubMedGoogle Scholar
- Biological Atlas of Insulin Resistance[http://www.bair.org.uk]
- McGee M, Chen Z: New Spiked-In Probe Sets for the Affymetrix HGU – 133A Latin Square Experiment. COBRA preprint Series 2006.Google Scholar
- Affymetrix: Statistical algorithms description document, Technical Report. 2002.Google Scholar
- Irizarry R, Hobbs B, Collin F, Beazer-Barclay Y, Antonellis K, Scherf U, Speed T: Exploration, Normalization, and Summaries of High Density Oligonucleotide Array Probe Level Data. Biostatistics 2003, 4(2):249–264. 10.1093/biostatistics/4.2.249View ArticlePubMedGoogle Scholar
- Li C, Hung Wong W: Model-based analysis of oligonucleotide arrays: model validation, design issues and standard error application. Genome Biol 2001, 2(8):RESEARCH0032.PubMed CentralPubMedGoogle Scholar
- Gentleman R, Carey V, Bates D, Bolstad B, Dettling M, Dudoit S, Ellis B, Gautier L, Ge Y, Gentry Jea: Bioconductor: open software development for computational biology and bioinformatics. Genome Biology 2004, 5: R:80. 10.1186/gb-2004-5-10-r80View ArticleGoogle Scholar
- Koza R, Nikonova L, Hogan J, Rim J, Mendoza T, Faulk C, Skaf J, Kozak L: Changes in gene expression foreshadow diet-induced obesity in genetically identical mice. PLoS Genetics 2006, 2(5):e81. 10.1371/journal.pgen.0020081PubMed CentralView ArticlePubMedGoogle Scholar
- Yang X, Schadt E, Wang S, Wang H, Arnold A, Ingram-Drake L, Drake T, Lusis A: Tissue-specific expression and regulation of sexually dimorphic genes in mice. Genome Research 2006, 16(8):995–1004. 10.1101/gr.5217506PubMed CentralView ArticlePubMedGoogle Scholar
- Recinos A, Carr B, Bartos D, Boldogh I, Carmical J, Belalcazar L, Brasier A: Liver gene expression associated with diet and lesion development in atherosclerosis-prone mice: induction of components of alternative complement pathway. Physiological Genomics 2004, 19: 131–142. 10.1152/physiolgenomics.00146.2003View ArticlePubMedGoogle Scholar
- Wang L, Balas B, Christ-Roberts C, Yeo Kim R, Ramos F, Kikani C, Li C, Deng C, Reyna S, Musi N, Dong L, DeFronzo R, Liu F: Peripheral Disruption of Grb10 Gene Enhances Insulin Signaling and Sensitivity in vivo. Mol Cell Biol 2007, 27(18):6497–6505. 10.1128/MCB.00679-07PubMed CentralView ArticlePubMedGoogle Scholar
- Ohtsubo K, Takamatsu S, Minowa M, Yoshida A, Takeuchi M, Marth J: Dietary and Genetic Control of Glucose Transporter 2 Glycosylation Promotes Insulin Secretion in Suppressing Diabetes. Cell 2005, 123(7):1307–1321. 10.1016/j.cell.2005.09.041View ArticlePubMedGoogle Scholar
Copyright
This article is published under license to BioMed Central Ltd. This is an Open Access article distributed under the terms of the Creative Commons Attribution License (http://creativecommons.org/licenses/by/2.0), which permits unrestricted use, distribution, and reproduction in any medium, provided the original work is properly cited.