Glossa Psycholinguistics

1. Introduction

Bilinguals regularly encounter word forms that exist in both of their languages. These word forms often have the same meaning, like the Dutch–English cognate wolf. Alternatively, they can have two entirely unrelated meanings, like the interlingual homograph angel, which means ‘insect’s sting’ in Dutch. A wealth of research indicates that bilinguals process these two types of words differently from translation equivalents, pairs of words like wortel and carrot that share their meaning but do not have similar orthographic or phonological forms. For example, in lexical decision experiments, participants respond more quickly to cognates than to translation equivalents, but tend to respond more slowly to interlingual homographs (for a review, see e.g. Dijkstra, 2005; Dijkstra & Van Heuven, 2012; Poort & Rodd, 2017a; Dijkstra & Van Heuven, 2018; Poort & Rodd, 2019b). Such findings support the view that words from a multilingual’s different languages are stored in a shared mental lexicon, and can become active in parallel during lexical access.

Additional evidence for cross-lingual interaction in the lexicon comes from a study showing that processing of cognates and interlingual homographs in a bilingual’s second language (L2) is influenced by recent experience with these words in their native language (L1; Poort et al., 2016). In this experiment, Dutch–English bilinguals first read sentences in Dutch that each contained either a cognate, an interlingual homograph or the Dutch translation of an English control word. After a delay of approximately 15 minutes, they then made English lexical decisions to these words. For the cognates, priming was beneficial: participants responded 28 ms more quickly to primed cognates than unprimed cognates. In contrast, for interlingual homographs, where there is a meaning change between the two languages, an interference effect was observed: priming delayed participants’ responses by 49 ms relative to unprimed homographs, suggesting that the increased availability of the Dutch meanings interfered with the participants’ ability to process the English reading of these ambiguous word forms. Given that many bilinguals regularly switch between their languages in their daily lives, this cross-lingual priming could have consequences for the ease with which they access and process the meanings of individual words in those languages after a switch.

In two follow-up experiments, we explored whether such cross-lingual priming also occurs for non-identical cognates (e.g. kat in Dutch and cat in English; Poort & Rodd, 2017b). These experiments, however, did not replicate the original findings: there was no statistical or numerical evidence for cross-lingual priming for the non-identical cognates, nor for the identical cognates or identical interlingual homographs. This inconsistency across experiments may result from the inconsistent need for participants to fully access semantic representations during lexical decision tasks (for details, see Poort & Rodd, 2017b, 2019b). This view was supported by our subsequent replication of Poort et al.’s (2016) original finding of facilitative priming for cognates but disruptive priming for interlingual homographs using a semantic relatedness task, which does consistently require participants to access the words’ meanings (Poort & Rodd, 2019b, Exp. 2). Still, the priming effects in this second follow-up study were smaller than expected based on the earlier findings: only 5 ms for the cognates and 10 ms for the interlingual homographs.

It remains uncertain, therefore, how much interaction occurs on a daily basis between word forms that exist in both of a bilingual’s languages, or what factors might modulate the size of this cross-lingual priming effect. Importantly, all the aforementioned experiments measured the extent to which experience in the participants’ L1 (Dutch) influenced subsequent processing in their L2 (English). The primary aim of the current experiment is therefore to determine whether the cross-lingual priming effect can also be observed when we reverse the direction of transfer, by priming items in the participants’ L2 (English) and then assessing the processing of these primed words in their L1 (Dutch). In addition, by directly comparing the results of the current experiment to our earlier data (Poort & Rodd, 2019b), we can investigate whether this switch in direction of priming will increase or decrease the magnitude of the observed effect.

Intuitively, most bilinguals would assume that experience in their L2 would be unlikely to influence subsequent processing of words in their L1 more than their L1 can influence their L2. Many bilinguals feel that their second language is more malleable than their first language, because it is the one they learnt later in life and so will always feel like a work in progress. In contrast, they have a lifetime of experience with their first language, which lends it a sense of relative stability. Current experimental evidence, while not decisive, appears to be in line with this subjective experience. If there exists any asymmetry at all in the extent to which each of a bilingual’s languages can exert influence on the other language, L1 indeed seems more likely to influence L2 than the other way around.

Support for this comes from a large body of research on (masked) translation priming that has shown that priming effects are consistently obtained when an L2 target item is primed by its L1 translation (i.e. L1-to-L2 priming; De Groot & Nas, 1991; Williams, 1994; Gollan et al., 1997; Jiang, 1999; Jiang & Forster, 2001; Kim & Davis, 2003; Basnight-Brown & Altarriba, 2007; Voga & Grainger, 2007; Duyck & Warlop, 2009; Dimitropoulou et al., 2011a, 2011b; Lupker et al., 2015; Wen & Van Heuven, 2017). When the direction of priming is reversed and an L1 target is primed by its L2 translation (i.e. L2-to-L1 priming), the priming effect is not as consistently observed and is usually weaker (Gollan et al., 1997; Grainger & Frenck-Mestre, 1998; Jiang, 1999; Jiang & Forster, 2001; Finkbeiner et al., 2004; Duyck & Warlop, 2009; Schoonbaert et al., 2009; Davis et al., 2010; Dimitropoulou et al., 2011a, cf. 2011b; Witzel & Forster, 2012; Nakayama et al., 2013; Wang, 2013; Chen et al., 2014; Wang & Forster, 2014).

The situation is not as simple as this, however, and proficiency appears to modulate priming significantly: the asymmetry in priming appears most robustly in unbalanced bilinguals (Gollan et al., 1997; Jiang, 1999; Finkbeiner et al., 2004; Nakayama et al., 2013), but decreases or is even entirely absent in balanced bilinguals (Perea et al., 2008; Duñabeitia, Dimitropoulou, et al., 2010; Duñabeitia, Perea, et al., 2010). For example, Davis et al. (2010) found that a group of beginning bilingual participants only showed significant masked priming effects for pairs of Spanish–English cognates (e.g. rich–rico) when the primes were in Spanish (L1) and the targets in English (L2). However, for more highly proficient bilinguals, Davis et al. (2010) found no effect of language dominance on the magnitude of this cognate priming effect. Similarly, using a picture naming paradigm with highly fluent Spanish–English bilinguals, Francis et al. (2003, Exp. 1) found that priming from L1-to-L2 was equally effective as priming from L2-to-L1.

Going even further, Lee et al. (2018) observed L2-to-L1 priming in unbalanced Korean–English bilinguals, although only when the stimulus onset asynchrony (SOA) was 150 ms. When the SOA was only 60 ms, no L2-to-L1 priming was observed, which they argue is due to the fact that 60 ms is not sufficient time for the L2 translation to prime the L1 target. This could also explain the lack of L2-to-L1 priming effects in unbalanced bilinguals in previous research. In other words, there may not be an asymmetry and, under the right circumstances, L2-to-L1 priming may be observed as consistently and be equally effective as L1-to-L2 priming.

Further support for this idea comes from previous research on monolingual word-meaning priming (Rodd et al., 2013; Rodd et al., 2016; Betts et al., 2018; Gilbert et al., 2018; Gaskell et al., 2019; Rodd, 2020; Gilbert et al., 2021), which as a paradigm is much closer in design to our bilingual studies (Poort et al., 2016; Poort & Rodd, 2017b, 2019b; Exp. 2) than (masked) translation priming. This research has shown that an encounter with one meaning of a homonym (e.g. the ‘tree covering’ meaning of bark) can boost the availability of this primed meaning relative to its unprimed meaning (i.e. the ‘dog noise’ meaning of bark). Importantly, larger priming effects are observed when the prime encounter is with the strongly subordinate meaning (Rodd et al., 2013; Betts, 2018). This effect of meaning dominance most likely arises because, for (relatively rare) subordinate meanings, even a single encounter is highly informative, indicating that this meaning is more likely to occur in the near future than the participant would otherwise have expected. In contrast, a single encounter with the highly frequent dominant meaning, which is more regularly encountered by participants in their everyday lives, is unlikely to produce a meaningful change to subsequent behaviour (Rodd et al., 2013; Betts, 2018). By analogy, we might therefore expect that for bilingual participants, an encounter with the relatively less familiar L2 interpretation of the word form would produce a more substantial change in how this word form is subsequently processed in L1, compared with the influence of an encounter with its L1 interpretation on L2 processing.

The current experiment addresses these questions about cross-lingual priming from L2 to L1 by replicating our previous experiment (Poort & Rodd, 2019b, Exp. 2), but switching the direction of priming: participants first encountered cognates and interlingual homographs (and translation equivalents) embedded in English (L2) sentences; the impact of this experience on processing of these words in the participants’ L1 was then tested using a Dutch semantic relatedness task. Setting the current experiment up as a replication of the previous experiment allowed for a direct, statistical comparison of the magnitude of the priming effect between these two experiments. We predicted that this comparison would show that the priming effect is greater when cognates and interlingual homographs are primed in the participants’ L2 (English, as in the current experiment) than when they were primed in the participants’ L1 (Dutch, as in the previous experiment).

Furthermore, in addition to these primary aims regarding the presence of cross-lingual priming, the data from the unprimed conditions in the current study can also contribute to the ongoing discussion surrounding the cognate facilitation and interlingual homograph inhibition effects. A number of studies have shown that (regardless of any priming), in lexical decision experiments, participants tend to respond more quickly to cognates than single-language control words, but more slowly or equally quickly to interlingual homographs compared with control words (for a review, see e.g. Dijkstra, 2005; Dijkstra & Van Heuven, 2012; Poort & Rodd, 2017a; Dijkstra & Van Heuven, 2018; Poort & Rodd, 2019b). Using a semantic relatedness task, we previously found a different pattern of results, however: we observed a strong interlingual homograph inhibition effect, but found no evidence for a cognate facilitation effect (Poort & Rodd, 2019b).

This pattern of results across these two tasks is most consistent with a ‘semantic settling’ account of lexical access (Rodd et al., 2004; Rodd, 2020) in which familiar word meanings correspond to stable states within a high-dimensional lexical-semantic space. Under this view, word-meaning access is characterised as a settling process during which the system takes time to resolve on a single familiar meaning (Armstrong & Plaut, 2016). We previously hypothesised that task differences in the processing of cognates and interlingual homographs arise because these tasks tap into different stages of this settling process (Poort & Rodd, 2019b). In particular, this hypothesis was informed by drawing analogies with research with monolinguals showing that the processing of different types of semantically ambiguous words also varies across tasks (Rodd et al., 2002, 2004; Beretta et al., 2005; Hino et al., 2006; Klepousniotou & Baum, 2007; Armstrong & Plaut, 2008, 2016).

Specifically, we drew a comparison between cognates and polysemous words, which are ambiguous within a given language (e.g. run evokes a different sense in each of the phrases the athlete/politician/film/river runs; Rodd, 2018). We noted that, like a polysemous word, the two (L1 and L2) interpretations of a cognate are highly similar, but often not entirely identical, as in the case of the cognate alarm, which in both Dutch and English is used to refer to a warning signal and the state of apprehension or fear immediately following the perception of such a signal. In English, however, the word alarm is also used to refer to the device that wakes a person up in the morning, which in Dutch is called a wekker. Like cognates, polysemous words are known to show a processing benefit (relative to unambiguous words) in monolingual lexical decision tasks in which participants do not need to settle on one specific sense in order to make a decision, which can be made on the basis of a general assessment of the familiarity of such words. Indeed, a benefit can arise from their relatively rich semantic representations (Rodd et al., 2002). Crucially, such ambiguity between related senses can delay processing in more explicitly semantic tasks that require participants to settle on a specific semantic representation (i.e. to resolve the ambiguity between the different senses; Rodd et al., 2002).

We further suggested that interlingual homographs, in contrast, are more akin to homonyms, which have multiple unrelated meanings within a language (e.g. bark meaning ‘the covering of a tree’ or ‘the sound a dog makes’). Monolinguals typically show a strong processing cost for these words in tasks that involve semantic processing (Hino et al., 2006) that is driven by strong competition between their unrelated and mutually inconsistent meanings (Rodd, 2020). This is consistent with the findings of slower responses by bilinguals to interlingual homographs in a semantic relatedness task (Poort & Rodd, 2019b), which is most likely also due to competition between the words’ unrelated, and mutually inconsistent, meanings. When a decision can be made based on a general assessment of the familiarity of the homonym or interlingual homograph (e.g. in lexical decision tasks), both monolinguals and bilinguals do not consistently exhibit such a large processing disadvantage (or at all).

The participants in the current experiment completed the semantic relatedness task in their native language instead of their second language as previously (Poort & Rodd, 2019b). The current experiment, therefore, allowed us to provide convergent evidence on this issue, by testing the prediction from this semantic settling account that our previous findings should also replicate in the participants’ native language. We predicted we would observe slower responses to interlingual homographs compared with translation equivalents, but that responses to cognates should not differ from these unambiguous word forms. We also tested an additional prediction, informed by research using lexical decision and picture naming tasks and eye-tracking during reading of a novel, that suggests that the cognate facilitation effect is greater in L2 than L1 (Kroll et al., 1999; Van Hell & Dijkstra, 2002; Cop et al., 2016). Although we did not expect to find a cognate effect, we predicted that the same logic would apply to interlingual homographs and, therefore, that when participants were tested in their native language (as in the current experiment), the interlingual homograph inhibition effect would be smaller than when participants were tested in their second language (as in the previous experiment).

In summary, the primary aim of this experiment was to obtain additional evidence regarding the high level of interaction between the words in a bilingual’s two languages by testing whether recent experience with ambiguous word forms in their L2 can influence subsequent processing of these items in their L1 and, secondary to that, whether L2-to-L1 priming effects will be greater than L1-to-L2 priming effects. In addition, data from the unprimed conditions of this experiment will speak to the debate concerning whether semantic settling accounts that were derived to explain processing of ambiguous word forms in monolinguals can also account for how word form ambiguity affects bilingual speakers (Rodd, 2020), by testing whether we can replicate the presence of an interlingual homograph inhibition effect and lack of a cognate facilitation effect in a semantic relatedness judgements task and, secondary to that, whether the size of the interlingual homograph inhibition effect will be smaller in L1 than L2.

3. Results

All analyses were carried out in R (version 3.4.4; R Core Team, 2018) using the lme4 package (version 1.1-17; Bates et al., 2015), following guidelines proposed by Barr et al. (2013) for model fitting (with the same amendments as previously; Poort & Rodd, 2019b) and using Type III Sums of Squares likelihood ratio tests to determine significance. Unless specified, the significance level was set at .05. Reaction times were analysed using the lmer() function with the default Nelder-Mead optimiser; accuracy data were analysed using the glmer() function with the bobyqa optimiser. The output for all analyses can be found in the experimentOverview.xlsx document in our OSF project. This section reports the results of our planned confirmatory analyses. No exploratory analyses are reported.

Three items (the cognates motto–motto and nest–nest and the translation equivalent bot–rude) were excluded from the analyses, as accuracy on the Dutch semantic relatedness task (43.4%, 50.0%, and 53.7%, respectively) was more than three standard deviations below the items’ word type mean. After excluding these items, we re-checked the matching of the word types and found that the difference in log-transformed frequency between the cognates and interlingual homographs was now significant (p = .047). This did not change our analysis, as we had already planned to include log-transformed frequency in the 2×3 analysis (see below).

Finally, note that the figures display the (harmonic) participant means, while the effects (and marginal means, MM) reported in the text were derived from the estimates of the fixed effects provided by the model summary (calculations are provided in the Excel document experimentOverview.xlsx in our OSF project). In figures that show harmonic participant means, we first calculated condition means for each participant (the dots) of the inverse-transformed reaction times and then transformed these back to the original scale before calculating an overall average per condition (the horizontal bars). For example, if a condition mean of inverse-transformed RTs for a particular participant was 1.63, their harmonic condition mean was 1000/1.63 = 613 ms and this is plotted as a dot in the figures. This can result in small discrepancies between the numbers reported in the text and the figures.

3.3 Test task: Dutch semantic relatedness task

3.3.1 Analysis procedure

The same analysis procedure was employed for the reaction times and accuracy data. In all cases, positive effects of priming indicate a facilitative effect of priming (i.e. faster reaction times and higher accuracy for primed items), while negative effects indicate a disruptive effect of priming (i.e. slower reaction times and lower accuracy for primed items). Positive (negative) effects of word type indicate an advantage (disadvantage) for the first-named word type over the second-named word type. The significance level (α-level) for all analyses was set at .05 unless otherwise noted.

Two fixed factors were included in the main 2×3 analysis: priming (2 within-participant/within-items levels: unprimed, primed) and word type (3 within-participant/between-items levels: cognate, interlingual homograph, translation equivalent). The maximal random effects structure of this model included a correlated random intercept and random slope for word type, priming and their interaction by participants and a correlated random intercept and random slope for priming by items. This maximal model did not converge for the reaction time analysis, nor did any other model with a random slope, so we used a random intercepts-only model. For the accuracy analysis, the maximal model also did not converge. As per our pre-registration, we removed the correlations between the random effects, but the model still did not converge, so next we removed the random slope for priming by items, which was the random effect with the smallest variance in the maximal model. The random effects in this model were again allowed to correlate and this model converged. Furthermore, as per our pre-registration, the 2×3 analysis was conducted both with and without the covariates item length, log-transformed frequency and OLD20, as there were significant differences between the three word types on these variables. These covariates were centred to have a mean of zero prior to including them in the model. Random slopes were not included for the covariates. The results of these analyses are only reported when the significance level of the effects of priming or word type (or their interaction) are different compared to the analysis without covariates. For transparency, the output is included in the experimentOverview.xslx document on the OSF.

In addition, three 2×2 analyses were conducted comparing the effect of priming for the cognates and interlingual homographs, the cognates and translation equivalents and the interlingual homographs and translation equivalents. The maximal random effects structure for these models included a correlated random intercept and random slope for word type, priming and their interaction by participants and a correlated random intercept and random slope for priming by items. For the reaction times, the maximal model again did not converge, so as per our pre-registration we removed the correlations between the random effects, after which the model did converge. For the accuracy analyses the maximal model converged. All p-values from these three analyses were compared against a Bonferroni-corrected α of .0167.

To examine the effect of priming for each of the three word types separately, three simple effects analyses were conducted. The maximal model converged for both the reaction times and accuracy analysis and included a correlated random intercept and random slope for priming by both participants and items. The p-values for these three analyses were compared against a Bonferroni-corrected α of .0167.

Finally, three pairwise comparisons were conducted on the unprimed data only, comparing the cognates, interlingual homographs and translation equivalents to each other. Only the unprimed trials were analysed as priming was expected to essentially increase both the cognate facilitation effect and the interlingual homograph inhibition effect. The maximal model converged for both the reaction times and accuracy analysis and included a correlated random intercept and random slope for word type by participants and a random intercept by items. The p-values for these three analyses were compared against a Bonferroni-corrected α of .0167.

Table 4 presents an overview of which effects were significant in which of the analyses we describe below. Naturally, given our hypotheses, our pre-registration also included analyses that compare the effects of priming and word type between the current experiment and our previous experiment (Poort & Rodd, 2019b, Exp. 2). These analyses revealed no significant differences between the two experiments. For brevity, we report these analyses in Appendix B of the Supplementary materials instead of in the main text. A summary of these results is presented in Table 5.

Table 4

Overview of the significant effects per analysis and dependent variable (reaction times and accuracy). Significant effects are significant at the α-level specified for that analysis in the Analysis procedure. Effect sizes (unstandardised) are provided only for significant effects and only if the factor had no more than two levels. As before, positive (negative) effects of priming indicate a facilitative (disruptive) effect of priming (i.e. faster (slower) reaction times and higher accuracy for primed items); positive (negative) effects of word type indicate an advantage (disadvantage) for the first-named word type over the second-named word type. ‘%pts’ means percentage points.

		Reaction times			Accuracy
		priming	word type	interaction	priming	word type	interaction
2×3		sig. (–5 ms)	sig.	n.s.	n.s.	sig.	n.s.
2×2s
	cognates vs. interlingual homographs	n.s.	sig. (59 ms)	n.s.	n.s.	sig. (8.7 %pts)	n.s.
	cognates vs. translation equivalents	n.s.	n.s.	n.s.	n.s.	n.s.	n.s.
	interlingual homographs vs. translation equivalents	sig. (–7 ms)	sig. (–65 ms)	n.s.	n.s.	sig. (9.1 %pts)	n.s.
simple effects
	cognates	n.s.	—	—	n.s.	—	—
	interlingual homographs	sig. (–16 ms)	—	—	n.s.	—	—
	translation equivalents	n.s.	—	—	n.s.	—	—
pairwise comparisons
	cognates vs. interlingual homographs	—	sig. (50 ms)	—	—	sig. (8.4 %pts)	—
	cognates vs. translation equivalents	—	n.s.	—	—	n.s.	—
	interlingual homographs vs. translation equivalents	—	sig. (–56 ms)	—	—	sig. (9.7 %pts)	—

Table 5

Overview of the significant effects per analysis and dependent variable (reaction times and accuracy) for the between-experiment analyses (see Supplementary materials, Appendix B). Significant effects are significant at the α-level specified for that analysis in the Analysis procedure outlined in the Supplementary materials, Appendix B. Effect sizes (unstandardised) are provided only for significant effects. Positive (negative) effects of priming indicate a facilitative (disruptive) effect of priming (i.e. faster (slower) reaction times and higher accuracy for primed items); positive (negative) effects of word type indicate an advantage (disadvantage) for the first-named word type over the second-named word type. Effects of experiment were never significant. ‘%pts’ means percentage points.

		Reaction times			Accuracy
		priming/word type	experiment	interaction	priming/word type	experiment	interaction
priming effects
	cognates	n.s.	n.s.	n.s.	n.s.	n.s.	n.s.
	interlingual homographs	sig. (–14 ms)	n.s.	n.s.	n.s.	n.s.	n.s.
word type effects
	cognates vs. interlingual homographs	sig. (46 ms)	n.s.	n.s.	sig. (7.5 %pts)	n.s.	n.s.
	cognates vs. translation equivalents	n.s.	n.s.	n.s.	n.s.	n.s.	n.s.
	interlingual homographs vs. translation equivalents	sig. (–50 ms)	n.s.	n.s.	sig. (–8.5 %pts)	n.s.	n.s.

3.3.2 Reaction times

Reaction times (RTs) faster than 300 ms or slower than 1500 ms were discarded (1.1% of the data), as were RTs for incorrect trials and trials that participants had not responded to (11.0% of the remaining data). The RTs were inverse-transformed (inverse-transformed RT = 1000/raw RT) to remedy violations of the assumption of normality and to be consistent with our previous approach (Poort & Rodd, 2019b, Exp. 2). After inverse-transforming the RTs, any inverse-transformed RTs were removed that were more than three standard deviations above or below a participant’s mean inverse-transformed RT (0.1% of the remaining data).

Figure 2 shows the reaction times per word type and priming condition. The violin is a symmetrical density plot showing the distribution of reaction times. Each dot in the figure represents a participant’s condition mean. The grey lines connect primed and unprimed means from the same participant for each word type. The horizontal bar and the number above it provide the mean across all participants for that condition. The figure suggests that priming was not effective for the cognates or translation equivalents, but hints at a disruptive effect of priming for the interlingual homographs. The figure further shows that there was considerable between-participant variation in reaction times and in the size of the priming effect for the three word types.

Figure 2

Harmonic participant means of the current experiment’s inverse-transformed Dutch semantic relatedness reaction times (in milliseconds) by word type (cognates, translation equivalents, interlingual homographs; x-axis) and priming (unprimed, dark grey; primed, light grey). Each point represents a condition mean for a participant with lines connecting means from the same participant. Each bar provides the mean across all participants in that condition. The violin is a symmetrical density plot.

3.3.2.1 2×3

The main effect of priming was significant [χ²(1) = 4.339, p = .037], with participants responding on average 5 ms more slowly to primed items than to unprimed items. The main effect of word type was significant [χ²(2) = 37.01, p < .001], indicating that there was a difference in reaction times between the three word types. The interaction between word type and priming was significant [χ²(2) = 7.057, p = .029].

3.3.2.2 2×2s

The 2×2 analysis that included the cognates and interlingual homographs showed no significant main effect of priming at the Bonferroni-corrected α-level of .0167 [χ²(1) = 5.642, p = .018, Δ = –7 ms]. The main effect of word type was significant [χ²(1) = 20.74, p < .001], with participants responding on average 59 ms more quickly to the cognates than to the interlingual homographs. The interaction between word type and priming was also not significant [χ²(1) = 4.929, p = .026; α = .0167].

The 2×2 analysis that included the cognates and translation equivalents showed no significant effects of priming [χ²(1) = 0.017, p = .897, Δ = –0.3 ms], word type [χ²(1) = 0.460, p = .498, Δ = –6 ms] or the interaction between priming and word type [χ²(1) = 0.001, p = .998].

In the 2×2 analysis that included the interlingual homographs and translation equivalents, the main effect of priming was significant [χ²(1) = 5.782, p = .016], with participants responding on average 7 ms more slowly to primed items than to unprimed items. The main effect of word type was also significant [χ²(1) = 27.02, p < .001], with participants responding on average 65 ms more slowly to the interlingual homographs than to the translation equivalents. The interaction between word type and priming was not signficant [χ²(1) = 5.143, p = .023].

3.3.2.3 Simple effects of priming

The simple effect of priming was not significant for the cognates [χ²(1) = 0.064, p = .800, Δ = –1 ms] or for the translation equivalents [χ²(1) = 0.009, p = .926, Δ = –0.3 ms]. However, the effect of priming was significant for the interlingual homographs [χ²(1) = 10.21, p = .001], with participants responding on average 16 ms more slowly to primed interlingual homographs than to unprimed interlingual homographs.

3.3.2.4 Pairwise comparisons of unprimed trials

The pairwise comparisons on the unprimed trials revealed a significant difference between the cognates and the interlingual homographs [χ²(1) = 16.93, p < .001], with participants responding on average 50 ms more quickly to the cognates. There was also a significant difference between the interlingual homographs and the English controls [χ²(1) = 21.58, p < .001], with participants responding on average 56 ms more slowly to the interlingual homographs. There was no significant difference between the cognates and the English controls [χ²(1) = 0.416, p = .519, Δ = –5 ms].

3.3.3 Accuracy

In line with the trimming procedure for the reaction times, any trials with RTs faster than 300 ms or slower than 1500 ms were removed.

Figure 3 shows the percentages correct per word type and priming condition. As in Figure 2, the violin is a symmetrical density plot showing the distribution of percentages correct. Each dot in the figure represents a participant’s condition mean. The grey lines connect primed and unprimed means from the same participant for each word type. The horizontal bar and the number above it provide the mean across all participants for that condition. The figure suggests that priming was not effective for the translation equivalents, but hints at a small facilitative effect of priming for the cognates and a small disruptive effect for the interlingual homographs. The figure further shows that there was a ceiling effect, especially for the cognates and translation equivalents and that, again, there was considerable between-participant variation in accuracy and in the size of the priming effect for the three word types. Note that the analyses below modelled the accuracy of individual responses (i.e. 1 if correct or 0 if incorrect) using logistic mixed effects models, not the participant means displayed in the figure.

Figure 3

Participant means of the current experiment’s Dutch semantic relatedness accuracy (in percentages correct) by word type (cognates, translation equivalents, interlingual homographs; x-axis) and priming (unprimed, dark grey; primed, light grey). Each point represents a condition mean for a participant with lines connecting means from the same participant. Each bar provides the mean across all participants in that condition. The violin is a symmetrical density plot.

3.3.3.1 2×3

The main effect of priming was not significant [χ²(1) = 0.091, p = .764, Δ = 0.1 percentage points]. The main effect of word type was significant [χ²(2) = 34.07, p < .001], indicating that there was a difference in accuracy between the three word types. The interaction between word type and priming was not significant [χ²(2) = 3.712, p = .156].

3.3.3.2 2×2s

In the 2×2 analysis that included the cognates and interlingual homographs, the main effect of priming was not significant [χ²(1) = 1.802, p = .180, Δ = 1.0 percentage points]. The main effect of word type was significant [χ²(1) = 19.53, p < .001], with participants’ accuracy being on average 8.7 percentage points higher for the cognates than the interlingual homographs. The interaction between word type and priming was also not significant [χ²(1) = 3.403, p = .065].

In the 2×2 analysis that included the cognates and translation equivalents, the main effect of priming was again not significant [χ²(1) = 0.076, p = .783, Δ = 0.2 percentage points], nor was the main effect of word type [χ²(1) = 0.407, p = .523, Δ = –0.5 percentage points] or the interaction between word type and priming [χ²(1) = 1.572, p = .210].

In the 2×2 analysis that included the interlingual homographs and translation equivalents, the main effect of priming was also not significant [χ²(1) = 0.369, p = .543, Δ = –0.4 percentage points]. The main effect of word type was significant [χ²(1) = 25.21, p < .001], with participants’ accuracy being on average 9.1 percentage points lower for the interlingual homographs than the English controls. The interaction between word type and priming was not significant [χ²(1) = 0.011, p = .916].

3.3.3.3 Simple effects of priming

None of the simple effects of priming were significant [for the cognates: χ²(1) = 2.030, p = .154, Δ = 1.1 percentage points; for the interlingual homographs: χ²(1) = 0.002, p = .970, Δ = –0.04 percentage points; for the translation equivalents: χ²(1) = 1.185, p = .276, Δ = –0.8 percentage points].

3.3.3.4 Pairwise comparisons of unprimed trials

The pairwise comparisons on the unprimed trials revealed a significant difference between the cognates and the interlingual homographs [χ²(1) = 15.74, p < .001], with participants’ accuracy being on average 8.4 percentage points higher for the cognates. There was also a significant difference between the interlingual homographs and the English controls [χ²(1) = 26.80, p < .001], with participants’ accuracy being on average 9.7 percentage points lower for the interlingual homographs. There was no significant difference between the cognates and the English controls [χ²(1) = 2.000, p = .157, Δ = –1.4 percentage points].

4. Discussion

This experiment addressed several questions regarding how bilinguals process cognates and interlingual homographs in their L1 and L2. Chief among these was whether processing of these words in their L1 would be affected by having encountered them approximately 15 minutes earlier in their L2. We predicted that this cross-lingual priming manipulation would result in faster and more accurate semantic relatedness judgements for cognates but slower and less accurate ones for interlingual homographs. In contrast, priming was not predicted to affect processing of the translation equivalents (also known as noncognates). We found that priming indeed did not affect the translation equivalents, but neither did it result in a significant benefit for the cognates. For the interlingual homographs, our hypothesis was confirmed in part: participants responded 16 ms more slowly to pairs including a primed interlingual homograph (compared with pairs including an unprimed homograph), but this significant priming effect did not carry over into the accuracy data nor was it significantly different from the null effects for the translation equivalents and the cognates.⁶

That we observed L2-to-L1 priming might be surprising, given that for many bilinguals their native language feels more stable and less susceptible to short-term changes than their second language. This finding supports the notion that the bilingual mental lexicon is indeed highly interactive, and that cross-lingual priming is bidirectional. On a daily basis, many bilinguals will regularly switch from speaking in one language to the other. This can result in language interference effects in many different experimental paradigms in both language production and comprehension (for a review, see Declerck & Philipp, 2015). In particular, as we have shown previously, it may lead to situations of cross-lingual priming that can have consequences for the ease with which bilinguals can access and process the meanings of individual words in those languages after a switch. We have now shown that this is not only the case when bilinguals switch from their native language to their second language, but also when they switch from their second to their native language. In addition, the fact that cross-lingual priming has been observed again using a semantic relatedness task as the test task adds support to the idea that such priming effects are more evident in semantic relatedness tasks than lexical decision tasks. Evidence for priming using lexical decision tasks has been inconsistent, perhaps due to the inconsistent need to fully access semantic representations during this task (Poort & Rodd, 2019b). This experiment highlights again the need for researchers in the field of bilingualism to move beyond using lexical decision tasks to study lexical access.

Based on findings from the monolingual word-meaning priming literature (Rodd et al., 2013; Betts, 2018), we had predicted that the priming effects for the cognates and interlingual homographs in this experiment would be larger than the priming effects observed in our previous experiment (Poort & Rodd, 2019b), in which these words were primed in the participants’ L1 instead. Although numerically larger, the priming effect for the interlingual homographs that we observed here (16 ms) was not statistically larger than the effect previously found (10 ms). The priming effect for the cognates in the current experiment was numerically smaller and in the opposite direction (–1 ms) compared to the previous experiment (5 ms), but this difference was also not significant. This means that, in contrast to our hypothesis, we did not find any evidence for an asymmetry of cross-lingual priming.

While this finding is, therefore, not entirely in line with the observation that, in monolingual participants, the subordinate meaning of a homonym is just as likely or even more likely to interfere with processing of the dominant meaning as the dominant meaning is to interfere with processing of the subordinate meaning (Rodd et al., 2013; Betts, 2018), it does fit in with (but unfortunately does not clarify) previous research on (masked) translation and repetition priming that produced mixed findings regarding the strength of L1-to-L2 and L2-to-L1 priming (Gollan et al., 1997; Jiang, 1999; Jiang & Forster, 2001; Francis et al., 2003; Finkbeiner et al., 2004; Perea et al., 2008; Duyck & Warlop, 2009; Schoonbaert et al., 2009; Davis et al., 2010; Duñabeitia, Dimitropoulou, et al., 2010; Duñabeitia, Perea, et al., 2010; Dimitropoulou et al., 2011a, 2011b; Witzel & Forster, 2012; Nakayama et al., 2013; Wang, 2013; Chen et al., 2014; Wang & Forster, 2014; Lee et al., 2018). As discussed in Section 1, this may be due to the fact that our participants were relatively balanced bilinguals, but were still more proficient in their L1 than L2.

Since this research suggests that priming is more effective from L1 to L2 in bilinguals who are more proficient in their L1 than their L2 (Gollan et al., 1997; Jiang, 1999; Finkbeiner et al., 2004; Nakayama et al., 2013), while there is no evidence for a directional asymmetry in more balanced bilinguals (Perea et al., 2008; Davis et al., 2010; Duñabeitia, Dimitropoulou, et al., 2010; Duñabeitia, Perea, et al., 2010), it could be that priming will be greater from L2 to L1 in (simultaneous) bilinguals who are equally proficient in both languages. Given this, and the fact that this absence of evidence should also not be taken as evidence for the absence of such an asymmetry, we suggest that future experiments should investigate our hypothesis further by studying bilinguals who are at least equally proficient in their L2 as their L1.

Although this finding of cross-lingual priming adds to a growing literature concerning interaction between the lexical representations of words in a bilingual’s different languages, the precise mechanism by which this priming occurs remains somewhat unclear. Early reports of word-meaning priming in monolingual speakers were taken as evidence that an encounter with one meaning of an ambiguous word (e.g. the ‘tree covering’ meaning of bark) resulted in a direct strengthening of this particular form-to-meaning mapping in the lexicon, making this meaning more available in the near future, relative to its alternative ‘dog noise’ meaning (Rodd et al., 2013). However, more recently it has been suggested that word-meaning priming does not reflect a direct change in the mental lexicon itself. In Gaskell et al.’s (2019) contextual binding account, they suggest that when homonyms (and other words) are encountered in sentence contexts, a temporary representation is formed in which the word is bound to the context in which it was experienced, in order to facilitate comprehension. They propose that this contextually bound representation of a word provides an additional source of information alongside long-term lexical knowledge that could influence the subsequent interpretation of that same word in the minutes (or hours) that follow exposure (Gaskell et al., 2019). Additional work is needed to discriminate between the predictions of these different mechanistic accounts of how recent lexical experience can influence how ambiguous word forms are processed for both within-language and cross-lingual priming.

Our final question of interest concerned the unprimed trials only. Previously, we found that the cognate facilitation effect disappears when a semantic relatedness task is used rather than the more usual lexical decision task and, simultaneously, that the interlingual homograph inhibition effect is enhanced (Poort & Rodd, 2019b, Exp. 1). In line with our current hypothesis, our data also showed that there was a strong interlingual homograph inhibition effect (of 56 ms and 9.7 percentage points) but no cognate facilitation effect. We based the explanation of our findings on research in the monolingual domain on semantic ambiguity resolution (Rodd et al., 2002, 2004; Armstrong & Plaut, 2008, 2016; Rodd, 2018). We posited that the representation of a cognate in the mental lexicon consists of two or more related senses (since many cognates do not have identical meanings, but often differ in their nuances). Like the senses of a polysemous word, they facilitate processing when only a superficial impression of the word’s meaning or word-likeness suffices (e.g. in lexical decision), but may compete to some extent for selection when a specific sense needs to be retrieved (e.g. in semantic relatedness). The representation of an interlingual homograph, in contrast, is thought to be linked to two distinct meanings, as for a homonym. These meanings may or may not compete when the situation does not require settling on a particular meaning (e.g. in lexical decision), but certainly will when one of these two meanings needs to be selected (e.g. in semantic relatedness). In other words, our replication of our previous finding (Poort & Rodd, 2019b, Exp. 1) provides further confirmation for the theory we outlined then.

We also predicted that, because the semantic relatedness task was performed in the participants’ first language (Dutch), the interlingual homograph inhibition effect would be smaller in this experiment than in our previous experiments (Poort & Rodd, 2019b), where the effect was tested in the bilinguals’ second language (English). This was not the case, and surprisingly, the interlingual homograph inhibition effect was numerically larger in the current experiment than the previous by 19 ms and 3.2 percentage points (56 ms and 9.7 percentage points compared with 37 ms and 6.5 percentage points). In line with this but also against our prediction, the difference between the cognates and interlingual homographs was not smaller in this experiment than in our previous experiment (50 ms and 8.4 percentage points versus 41 ms and 6.3 percentage points previously). However, these (between-participants) differences were not significant and so should not be taken as strong evidence for an asymmetry in the opposite direction than predicted.

These findings diverge from previous literature (Kroll et al., 1999; Van Hell & Dijkstra, 2002; Cop et al., 2016) showing that cognate effects tend to be larger in L2 than L1. Van Hell and Dijkstra (2002), for example, found that the cognate facilitation effect in their lexical decision experiment was larger when participants were tested in their second language than when they were tested in their native language. This finding likely matches most researchers’ (and lay bilinguals’) intuition, as the representation of a cognate is thought to be less stable in a second language. This means there would be more scope for the L1 representation to aid processing in the L2. If we extend this reasoning to interlingual homographs, we would expect that the L1 representation would be more likely to interfere with processing in the L2 than vice versa. Certainly anecdotally, interlingual homographs tend to be especially tricky in one’s second language, and we imagine this is even more true in a semantic relatedness task than a lexical decision task. It is, therefore, unclear to us why we failed to observe a similar effect for interlingual homographs. We suggest that future research using a larger sample should explore in more detail how the characteristics of the participants may influence the size of the cognate facilitation effect and interlingual homograph inhibition effect. For example, in this case as well, language proficiency and differences in recent and longer-term experiences (i.e. language dominance) might be modulating the relative magnitude of these effects in the bilingual’s different languages. Furthermore, a similar study that employed a within-participants design instead of the between-participants design that we used would be able to establish with greater certainty whether priming from L1 to L2 or L2 to L1 is stronger.

In summary, this experiment replicated the finding that a bilingual’s prior experience with an interlingual homograph can influence its subsequent processing in their other language (Poort et al., 2016; Poort & Rodd, 2019b). For the first time, we have also shown that this interference effect is not restricted to the processing of words in the participants’ (potentially more malleable) L2: in the current experiment participants’ ability to process these words in their (more stable) L1 was hindered by their prior encounter with the word form in their L2. In addition, we showed that the interlingual homograph inhibition effect is present not only when participants make semantic relatedness judgements in their L2, but also in their L1. These findings add to a growing literature that emphasises the high level of interaction between words in a bilingual’s different languages within their mental lexicon (for a review, see e.g. Dijkstra, 2005; Dijkstra & Van Heuven, 2012; Poort & Rodd, 2017a; Dijkstra & Van Heuven, 2018; Poort & Rodd, 2019b).

Data Accessibility Statement

All data and supplementary materials are available on the Open Science Framework via https://osf.io/b9az4/ (doi: 10.17605/OSF.IO/2SWYG). The experiment itself can be previewed and cloned via Gorilla Open Materials (app.gorilla.sc/openmaterials/245694).

Ethics and Consent

The UCL Experimental Psychology Ethics Committee provided approval of the study protocols (Project ID: EP/2017/009).

Competing Interests

The authors have no competing interests to declare.

Author contributions

• Eva D. Poort: conceptualisation, data curation, formal analysis, funding acquisition, investigation, methodology, project administration, resources, software, validation, visualisation, writing – original draft, writing – review & editing.

• Jennifer M. Rodd: conceptualisation, funding acquisition, methodology, project administration, resources, supervision, writing – original draft, writing – review & editing.